H25 | Macro Paper Warehouse

International Trade Responses to Labor Market Regulations

Mon, 01 Jan 0001 00:00:00 +0000

Overview

Research Question. This paper asks whether differences in labor market regulations — specifically payroll taxes and minimum wages — shape countries’ comparative advantage in the cross-border provision of labor-intensive services. The question has broad policy relevance: if lower labor standards confer a systematic trade advantage, countries may face pressure to race to the bottom in labor protections, and political support for economic integration may erode.

Setting and Identification. The paper exploits the EU “posting policy,” a large trade program established in 1959 that allows firms in one EU member state to temporarily send their employees to perform service contracts in another member state. In 2017, posting accounted for roughly one-third of all within-EU trade in services (approximately 2% of EU GDP), involving about 2 million workers (in full-time equivalents) in 2019. The setting is analytically attractive because competing foreign and domestic firms serve the same customers at the same physical location using shared capital, holding most determinants of comparative advantage constant while labor market regulations vary by the firm’s country of origin.

Under posting rules, payroll taxes are generally origin-based (exporting firms pay their home country’s tax rate) but become destination-based when contracts exceed a regulatory duration threshold (12 months pre-2010, 24 months from 2010–2020, 18 months from 2020 onward). Minimum wages are destination-based: foreign firms must match the importing country’s statutory minimum wage floor when it exceeds the workers’ home-country wage level. This generates the paper’s key identifying variation — payroll taxes and minimum wages vary across countries, over time, and within countries across sectors.

Data. The author uses administrative A1 social security forms filed for every EU posting contract from 2007–2018, collected from 25 EU member states, supplemented by micro-level national posting registries in Belgium (LIMOSA), France (SIPSI), and Luxembourg (matched employer-employee data). Labor cost data (wages, payroll tax rates, minimum wages) come from Eurostat and the OECD Taxing Wages Dataset.

Methodology. The paper proceeds in three steps. First, it documents steady-state cross-sectional correlations between bilateral posting flows and labor cost differentials. Second, it estimates difference-in-differences (DiD) elasticities from four quasi-natural experiments. Third, it estimates a theory-consistent gravity model using all sources of variation across 25 EU countries from 2009–2018.

Main Findings.

Steady-state correlation: A strong negative relationship exists between bilateral posting flows and labor cost differentials, with a cross-sectional elasticity of approximately –0.58 (SE 0.08). In sharp contrast, the relationship between bilateral goods trade and labor cost differentials is weak and if anything marginally positive (point estimate +0.13), confirming that labor cost differences are a distinctive driver of trade specifically in labor-intensive services rather than goods.
Belgian tax shift (2016–2018): When Belgium cut employers’ social security contributions from 33% to 25%, imports of posting services into Belgium slowed relative to France (a neighboring control country on parallel pre-reform trends). The reduced-form elasticity of posting imports with respect to the payroll tax rate is 1.45 (SE 0.3).
Luxembourg EU regulation reform (2010): A new EU regulation required temporary employment agencies in border regions to pay destination-based payroll taxes, raising statutory rates faced by Luxembourgish exporters from 15% to 44%. Posting exports from Luxembourg’s temporary employment sector fell by 40% relative to the pre-reform level and relative to the domestic (control) sector, while the sheltered road transportation sector showed no response. The reduced-form elasticity with respect to the statutory payroll tax rate is –1.55 (SE 0.24), and the triple-difference estimate is –1.37 (SE 0.08).
Bunching at duration thresholds: The distribution of posting contract lengths in France (which has the EU’s highest payroll taxes) shows a sharp spike just below the 24-month payroll tax threshold. When the threshold was moved to 18 months in 2020, excess mass migrated to the new threshold, confirming that bunching reflects behavioral responses to the tax notch rather than reference-point effects. This documents that payroll tax differentials shape not only the quantity (extensive margin) but also the length (intensive margin) of posting contracts.
German minimum wage reform (2015): Germany’s introduction of a national minimum wage of €8.50 per hour — which was already binding on construction workers through a sectoral minimum, but not on foreign firms providing non-construction services — caused postings to Germany in manufacturing to fall by approximately 60% relative to the construction (control) sector. The reduced-form elasticity is –1.34 (SE 0.43). Heterogeneity analysis shows that export declines were monotonically larger for low-wage origin countries where the new minimum wage was binding, and placebo estimates using Germany’s high-wage neighboring countries (where minimum wage requirements did not change) are statistically indistinguishable from zero.
Gravity estimates: The preferred specification (PPML with origin-year, destination-year, and pair fixed effects, exploiting bilateral variation in minimum wage bindingness across origin countries) yields a model-implied trade elasticity θ of –1.2 (SE 0.2). The range across specifications is –1.2 to –2.4. These estimates are smaller than the goods trade elasticity (typically estimated around 5) and below the medium-run reduced-form elasticities from the DiD case studies, consistent with short-run gravity estimates capturing only partial adjustment while DiD designs measure longer-run equilibrium responses.

Policy Counterfactual. The paper’s estimates imply that the Bolkestein Directive — which proposed exempting foreign firms from all destination-country labor regulations — would have doubled exports of physical services from Eastern European countries (upper bound), as their cost advantage would have been dramatically amplified by removal of minimum wage requirements. Counterpart to this export boom, average posted workers’ wages would have fallen by approximately 16%, since workers would lose their entitlement to destination-country minimum wages. The paper documents that the Bolkestein controversy — sparked by the “Polish plumber” debate in early 2005 — coincided with a sharp and persistent drop in French voter support for the EU constitutional treaty, which was subsequently rejected.

Scope Conditions. Results apply specifically to trade in physical (labor-intensive) services traded via temporary worker posting within the EU, where productivity differences across countries for these tasks are plausibly small (Balassa-Samuelson), making institutional factors a primary driver of wage differences. The paper estimates intent-to-treat effects, assuming perfect compliance by exporting firms. The paper does not perform a comprehensive welfare analysis covering consumer price effects or general equilibrium wage and trade-balance responses.

In depth

Q1. What is the EU posting policy and why does it provide an unusually clean setting for identifying the causal effect of labor regulations on trade?

The EU posting policy, established in 1959, allows firms in one EU member state to temporarily send employees to perform service contracts in another member state. The policy keeps most determinants of comparative advantage constant — competing foreign and domestic firms serve the same customers at the same physical location using shared capital — while labor market regulations vary by the firm’s country of origin. Productivity differences for physical services across countries are also plausibly limited (Balassa-Samuelson), making institutional wage differences the primary cost driver. Enforcement is facilitated by the on-site nature of the service, and administrative A1 forms create a direct measure of the number of workers involved in cross-border transactions without a minimum reporting threshold.

Q2. What are the three sources of labor cost differences the paper identifies and quantifies?

Foreign firms competing for posting contracts face different costs through three channels: (i) equilibrium gross wages differ across origin countries, reflecting both productivity differences and institutional/information frictions that allow wage discrimination between posted and domestic workers; (ii) payroll tax rates are origin-based and differ substantially across countries (for example, France’s employer payroll tax is approximately 40% versus approximately 15% for Luxembourg before the 2010 reform); and (iii) destination-specific minimum wages impose a “posting allowance” on firms from countries with lower wages, equal to the shortfall between the firm’s home-country wage and the importing country’s minimum wage floor. Micro-level wage data from France confirm that most posted workers from low-wage countries are paid exactly at the French minimum wage, demonstrating the bindingness of the third channel, while French workers performing the same tasks receive wages near the French average (approximately €21.1 per hour versus a minimum wage of approximately €10 per hour in 2018).

Q3. What does the cross-sectional evidence show about the relationship between labor cost differentials and posting flows, and how does this compare to goods trade?

Bilateral posting flows and bilateral labor cost differentials have a tight negative cross-sectional relationship with an estimated elasticity of –0.58 (SE 0.08), indicating that countries export more posting services when their labor costs are substantially below those of the destination country. The same exercise applied to bilateral goods trade yields a coefficient of +0.13 (SE 0.07) — weak and marginally positive — consistent with goods trade being driven by capital, technology, and scale rather than labor cost differentials. The gap confirms that labor cost differences are a distinctive comparative advantage mechanism for labor-intensive services but not for less labor-intensive goods.

Q4. What does the Belgian tax shift reform demonstrate, and how is identification established?

Belgium cut employer social security contributions from 33% to 25% between 2016 and 2018 in a revenue-neutral reform (financed by VAT, excise duties, and dividend taxes). The DiD compares posting imports into Belgium with those into France (a neighboring, similarly sized importer on parallel pre-reform trends). Belgium and France imported posting services at similar rates before 2015; Belgian imports slowed immediately after the reform while French imports continued growing. The reduced-form elasticity of posting flows with respect to the destination payroll tax rate is 1.45 (SE 0.3). The elasticity with respect to total labor cost is 3.7 (SE 0.7). No discernible response is detected for trade in manufacturing goods, providing a within-reform placebo. A synthetic control using all available importing countries yields a smaller elasticity of 0.6 (SE 0.22).

Q5. How does the Luxembourg EU regulation reform (2010) improve on the Belgian case for identification?

The 2010 EU regulation required temporary employment agencies in border regions to pay destination-based (rather than origin-based) payroll taxes, raising statutory rates for Luxembourgish exporters from 15% to 44%. Unlike the Belgian reform, this created within-country variation: the same Luxembourgish firms were exposed in the temporary employment sector but not in road transportation (which received a 10-year exemption). This within-exporter, cross-sector design controls for all Luxembourg-wide demand or supply shocks. Posting exports by the temporary employment sector fell 40% relative to pre-reform levels and relative to the domestic (control) sector, while road transportation posting showed zero response. The monthly data confirm the drop occurred in the exact month following the regulation with no anticipation. The triple-difference elasticity (with respect to the payroll tax rate) is –1.37 (SE 0.08).

Q6. What does the bunching evidence at payroll tax duration thresholds add to the DiD findings?

When posting contracts exceed a regulatory duration threshold (24 months during 2010–2020, then 18 months from July 2020), payroll taxes become destination-based. Because France has the highest payroll tax in the EU, all exporting firms face strong incentives to avoid crossing the threshold. The distribution of posting contract lengths in France shows sharp excess mass just below 24 months in 2017. When the threshold moved to 18 months in 2020, the excess mass migrated to the new threshold while diminishing at the old one, confirming that bunching is tax-motivated rather than driven by a reference-point at 24 months. This establishes that labor tax differentials shape not only the quantity of posting contracts (extensive margin) but also their length (intensive margin).

Q7. What are the main findings from the German minimum wage reform, and how do the heterogeneity tests strengthen identification?

Germany’s January 2015 introduction of a national minimum wage of €8.50 per hour (preceded by a sectoral minimum in meat processing in August 2014) raised wage costs for foreign firms providing non-construction services, but not for construction firms already covered by a higher sectoral minimum. Postings to Germany in manufacturing fell by approximately 60% relative to the construction (control) sector, implying a reduced-form elasticity of –1.34 (SE 0.43). Two heterogeneity tests reinforce identification: (i) within the treated German sector, posting declines are monotonically increasing in the degree to which the new minimum wage is binding in the origin country, with Luxembourg (where the minimum is non-binding) showing no statistically significant effect; (ii) the same industry-by-country comparison in Germany’s high-wage neighboring countries (which did not change minimum wage rules) yields placebo estimates statistically indistinguishable from zero. The reform raised wages for German workers by an average of 6% (and up to 10% for most affected workers) but automatically raised wages for posted workers by an average of 40%, doubling them for workers from the poorest sending countries.

Q8. How do the gravity model estimates compare to the reduced-form DiD estimates, and what explains the difference?

Across gravity specifications, model-implied elasticities range from –0.75 to –2.4. The preferred specification — PPML with pair fixed effects, destination-year fixed effects, and origin-year fixed effects — yields θ = –1.2 (SE 0.2). These estimates are systematically below the medium-run reduced-form DiD estimates because: (a) the gravity model uses nationwide average tax and minimum wage measures that introduce measurement error relative to the sector-specific reforms in the case studies; and (b) the gravity model captures year-to-year (short-run) adjustments, while the DiD designs compare outcomes several years before and after the reform, picking up longer-run equilibrium reallocation. The finding that responses grow over time mirrors evidence on dynamic adjustment in goods trade (Boehm, Levchenko and Pandalai-Nayar, 2023), and contradicts the conventional belief that fiscal devaluations boost exports only in the short run.

Q9. What does the gravity model reveal about trade in goods as a function of posting-specific wage costs?

When the same gravity specification is applied to bilateral goods trade rather than posting flows, posting-specific wage costs have a positive — not negative — coefficient on goods trade. This is inconsistent with a model where unobserved shocks affect all exports symmetrically, and instead suggests a small substitution effect: as the cost to import labor services rises (due to tighter posting regulations), countries substitute toward importing goods. For some activities (such as meat processing), importing finished goods is a partial substitute for importing labor services to produce on-site.

Q10. What are the Bolkestein Directive counterfactual implications, and how do they connect to the political economy evidence?

The Bolkestein Directive (proposed 2005) would have enforced a “country of origin principle,” exempting foreign posting firms from destination-country minimum wages. Using the preferred lower-bound elasticity from the gravity model (column 5, θ = –1.2) and an upper bound averaging gravity and DiD estimates, the paper predicts this would have at least doubled exports of labor services from Eastern European countries. Tax revenues collected on posted workers in origin countries would also double. However, average posted workers’ wages would fall by approximately 16%, as workers would lose their entitlement to destination-country minimum wages. The paper documents that the Bolkestein controversy — introduced to the EU Parliament in March 2005 and popularized via the “Polish plumber” trope — coincided with a sharp and permanent drop in French voter support for the EU constitutional treaty, which was subsequently rejected in referendum. This is consistent with Rodrik’s (1998) hypothesis that voters withdraw support for economic integration when comparative advantage appears to be based on institutional choices that conflict with importing countries’ social norms.

Q11. How does the paper handle the incidence of payroll taxes — does the canonical result that payroll taxes are fully passed through to workers hold in this context?

The canonical competitive labor market model predicts full pass-through of payroll taxes to workers’ net wages, leaving firms’ labor costs unchanged. The paper finds substantial trade responses to payroll tax reforms, inconsistent with full pass-through. Nominal rigidities — including binding minimum wages that constrain downward wage adjustment — help rationalize incomplete pass-through in the EU context. The paper estimates elasticities both with respect to statutory tax rates (the reduced-form, making no incidence assumption) and with respect to total wage costs (instrumented with the reform, allowing for gross wage responses). Wage data from Belgium show no distinguishable wage response to the Belgian tax cut, suggesting the incidence fell largely on firms’ costs rather than workers’ wages in that episode.

Q12. What do the destination-based taxation counterfactual (tax cooperation proposal) calculations show?

A proposal to shift all posting payroll taxation to destination-based rates would decrease posting exports from Eastern European countries by between 10% and 25%. Despite the volume reduction, total taxes collected on posted workers would still increase under this reform even when the upper-bound elasticity (approximately –3.7 with respect to total wage cost) is used, because a 1% increase in the payroll tax rate translates to a much smaller proportional increase in total wage cost.

Key Concepts

Posted workers / posting policy: Employees temporarily sent by their employer (the “exporting firm”) to perform a service contract in another EU member state. Posted workers maintain their employment contract with the firm in the origin country but physically work in the destination country. This creates a setting where competing domestic and foreign firms serve the same customers at the same location under different labor regulations.

Posting allowance: The additional wage component that exporting firms must pay to posted workers to satisfy the destination country’s minimum legal wage when that minimum exceeds the firm’s home-country wage level. The posting allowance is zero when the exporting country’s average wage already exceeds the destination minimum wage; it can be large for low-wage origin countries. The allowance enters directly into firms’ labor costs and is the minimum-wage channel of the paper’s labor cost formula.

Origin-based vs. destination-based payroll taxation: Under posting, payroll taxes are normally assessed in the country where the exporting firm is registered (origin-based), creating tax rate differentials between competing firms in the same job site. EU regulations convert payroll taxes to destination-based when posting contracts exceed a duration threshold, eliminating the tax advantage of lower-tax origin countries for those contracts. The 2010 EU regulation additionally imposed destination-based taxation on border-region temporary employment agencies.

Trade elasticity for physical services (θ): The structural parameter from the Eaton-Kortum (2002) gravity model that governs the elasticity of bilateral posting flows with respect to changes in firms’ total wage costs when exporting services from country i to country j. The paper’s preferred estimate is –1.2 (from gravity estimation) to approximately –1.3 to –1.5 (from reduced-form DiD designs), substantially smaller in absolute value than the goods trade elasticity (typically estimated around 5).

Social standards as comparative advantage: The paper uses “standards” to refer to countries’ domestic policy choices about payroll taxes (which finance social insurance programs) and minimum wages (which set worker protection floors). The paper demonstrates that these regulatory choices — distinct from productivity differences, factor abundance, or technology — create measurable cost advantages that shape specialization in labor-intensive service sectors. This is in contrast to “benign” sources of comparative advantage.

Bolkestein Directive / country of origin principle: A 2005 EU legislative proposal that would have required posting firms to operate under the laws of their home country when supplying services in other EU member states, eliminating the hard core of destination-country regulations (including minimum wages) that the 1996 Posted Workers Directive had imposed on foreign firms. The proposal was withdrawn after a wave of protests and its association with a sharp fall in French support for the EU constitutional treaty.

Bunching / notch at duration threshold: A behavioral response in which exporting firms strategically keep posting contract lengths below the duration threshold that triggers destination-based payroll taxation, generating an excess mass in the distribution of contract lengths just below the threshold. The paper uses this bunching, together with the movement of the threshold from 24 to 18 months in 2020, as additional evidence that payroll tax differentials affect the intensive margin of posting.

On the Nature of Entrepreneurship

Mon, 01 Jan 0001 00:00:00 +0000

This paper uses a novel longitudinal administrative dataset drawn from U.S. Internal Revenue Service (IRS) and Social Security Administration (SSA) records to characterize income dynamics and the determinants of entrepreneurial entry for pass-through business owners — sole proprietors, partners, and S corporation owners — who collectively account for over 50 percent of all U.S. business net income. The sample covers 2000–2015 and includes up to 1.3 billion person-year observations for individuals aged 25–65. The authors construct balanced panels using birth cohorts 1950–1975, impute education (college attainment) and skill (cognitive, interpersonal, manual) via machine-learning classifiers trained on CPS and O*NET data, and estimate life-cycle income profiles using a three-component model that separates individual fixed effects, group-specific time effects, and group-cohort-specific age effects.

The paper’s central departure from prior work is coverage of the full income distribution, including the high-earning right tail that household surveys such as the CPS misrepresent due to top-coding and small samples. When the IRS and CPS samples are compared on a consistent classification basis, median self-employment income is lower in the IRS data at all ages, consistent with the survey literature’s emphasis on the “typical” self-employed individual. However, mean incomes diverge sharply: the IRS shows mean self-employment income rising from $23 thousand at age 25 to $93 thousand at age 55, whereas the CPS (with incorporated owners reclassified) shows a rise from only $41 thousand to $73 thousand. Roughly 80 percent of self-employment income in the IRS data accrues to individuals above the $100 thousand threshold, compared to 42–53 percent in the CPS. The IRS-CPS gap is dominated by the right tail and concentrated in professional services and health care. For paid-employed individuals, the IRS and CPS medians and means are close at all ages, confirming the discrepancy is specific to self-employment.

The life-cycle estimation finds that individuals who have “tried self-employment” — a group earning virtually all self-employment income — start at similar average incomes to primarily paid-employed peers at age 25 but reach $134 thousand by age 55, compared with $79 thousand for paid-employed peers with the same observable characteristics. Age effects for the self-employed are 63 percent higher than for the paid-employed at age 26 and remain elevated until age 55. Time effects show dramatically greater cyclical volatility for the self-employed: income growth declined by $9,655 (2008) and $8,785 (2009) for the self-employed versus $373 and $1,583 for paid-employed in the same years, concentrated in real estate and construction.

On the determinants of entry, the paper finds: (i) no evidence that house-price appreciation raises entry rates, contra collateral-constraint hypotheses; (ii) most entrants have lower asset incomes than future entrants with the same characteristics, arguing against a liquid-wealth precondition; (iii) most entrants have higher prior labor income than future entrants, consistent with entry being driven by on-the-job experience rather than fallback from low-paid work; (iv) almost all founders report positive individual tax income in their first year of operation despite negative business net income and no external debt financing. Self-employed income growth exhibits greater dispersion — a 10th-to-90th percentile range roughly 2.5 times wider than for the paid-employed — and a Kelly skewness about 0.1 higher. A standard consumption-risk model calibrated with household-finance estimates of risk aversion rationalizes the patterns if individuals are insured against the most adverse downside shocks. Entry and exit rates are stable across the sample period, including the Great Recession, and the entrepreneurship share does not decline.

The subgroup congruent with non-pecuniary motivation — primarily self-employed individuals earning less than paid-employed peers with matching characteristics — comprises roughly 57 percent of primarily self-employed by count but earns only 16 percent of total self-employment income.

In depth

Q1. Why do IRS and CPS data give such different pictures of self-employment income?

The CPS suffers from top-coding of high incomes and small samples that underrepresent high earners in key industries. The IRS-CPS mean income gap for the self-employed is dominated by the right tail: in the main IRS sample, individuals above the $100 thousand threshold earn roughly 80 percent of all self-employment income, versus 42 percent in the comparable CPS sample. The average income of top earners above $100 thousand is $355 thousand in the IRS versus $218 thousand in the CPS. The gap is concentrated in professional services and health care and persists across all income thresholds and sample definitions tested. No analogous discrepancy exists for paid-employed individuals, where IRS and CPS medians and means are close at all ages.

Q2. What does the comparison look like at the median versus the mean?

At the median, IRS self-employment income is lower than both CPS samples at all ages, with the gap largest for younger owners and those with incorporated businesses — a pattern consistent with the survey-based “self-employment discount” narrative. At the mean, the IRS shows much higher income at older ages: by age 55, IRS mean self-employment income is $93 thousand versus $73 thousand in the CPS sample that includes reclassified incorporated-owner wages. The divergence arises because the mean is sensitive to the right tail, which the CPS systematically underrepresents.

Q3. How does the paper estimate life-cycle income profiles while separating age, time, and cohort effects?

Individual income is decomposed into an individual fixed effect (permanent latent ability and preferences), a group-specific time effect (business-cycle fluctuations common to a group), and a group-cohort-specific age effect (life-cycle income growth). Identification exploits the overlapping cohort structure of the 16-year panel: age effects are assumed equal across cohort bins of size at least two, allowing time and age effects to be separately identified. The model is estimated in levels rather than logs to accommodate business losses. Groups are defined as a Cartesian product of 32,256 subgroups based on education, three skill dimensions, industry (21 two-digit NAICS codes), demographics (gender, cohort, marital status, children), and employment-status history.

Q4. What are the headline life-cycle income profile findings for self- versus paid-employed?

Among the “primarily employed” group, those who have tried self-employment and those who are primarily paid-employed have similar average incomes at age 25. By age 55 the self-employed reach an estimated $134 thousand (2012 dollars) versus $79 thousand for paid-employed peers with identical observable characteristics. The estimated age effect for the self-employed is 63 percent higher than for the paid-employed at age 26 and remains higher through age 55. These gaps would widen further if incomes were adjusted upward for the BEA-estimated net misreporting rates of 46 percent for unincorporated owners and 14 percent for S corporation owners.

Q5. How large is the group consistent with non-pecuniary motivation, and how much income does it earn?

The non-pecuniary subgroup — primarily self-employed individuals (at least 12 years in self-employment) who earn less on average than primarily paid-employed peers matched on gender, education, skills, and other characteristics — is numerically larger, comprising approximately 57 percent of primarily self-employed by count. However, this group earns only 16 percent of total self-employment income. Adjusting for paid-employed fringe benefits and self-employed income misreporting can change the group’s size but does not alter the finding that it accounts for a small income share. The paper concludes that non-pecuniary motives may guide occupational choice for many individuals but are not the driver of the typical dollar earned in self-employment.

Q6. How does idiosyncratic income risk compare between self- and paid-employed?

Self-employed income changes are substantially more dispersed: the 10th-to-90th percentile range of income growth is roughly 2.5 times wider for the self-employed than for the paid-employed. Income changes for the self-employed are also more right-skewed, with a Kelly skewness difference of approximately 0.1. When a standard consumption-risk model — augmented with a lower bound on consumption growth to allow for external insurance — is parameterized with risk-aversion estimates from the household finance literature, the observed patterns are rationalized if individuals are insured against the most adverse downside shocks, i.e., the attractive aspect of self-employment is large potential upside with insured downside.

Q7. What happened to self-employed income and exit rates during the Great Recession?

Time effects show steep income growth declines for the self-employed of -$9,655 in 2008 and -$8,785 in 2009, compared with much more modest declines of -$373 and -$1,583 for paid-employed peers. The aggregate income declines are concentrated in cyclically sensitive self-employed subgroups in real estate and construction, with their paid-employed counterparts experiencing only modest declines. Despite these large income shocks, exit rates from self-employment showed little change during the Great Recession, either in aggregate or in the cyclically sensitive sectors. Entry rates were likewise stable, and the share of entrepreneurs in the population did not decline over the full sample period.

Q8. Does the evidence support collateral constraints as a binding barrier to entrepreneurial entry?

No. The paper tests the hypothesis, standard in the liquidity-constraints literature, that entry rates should be higher for homeowners experiencing house-price appreciation (which raises collateral value). The IRS data do not support this prediction. Separately, comparing asset incomes (interest, dividends, capital gains) of current entrants and future entrants with the same characteristics, the paper finds that most current entrants have lower asset incomes and less liquid wealth than those who switch later, which also argues against a liquid-wealth precondition for entry.

Q9. What does prior labor income reveal about why people enter self-employment?

Current entrants have higher prior labor income than matched future entrants with the same characteristics, indicating they enter with accumulated on-the-job experience rather than being pushed into self-employment as a fallback after failure in paid work. This is consistent with self-employment being a deliberate, experience-driven career transition for most entrants rather than a last resort for low earners. The paper interprets this as positive evidence for the role of experience-based human capital in driving entrepreneurial choice.

Q10. How do founders finance startup costs if most have negative business net income in early years?

Almost all founders in the sample report positive income on their personal (individual) tax form in the first year of operation, even though most report negative business net income and carry no external debt financing. This pattern suggests founders rely on personal income sources — prior savings, part-time paid employment, or spousal income — to cover startup costs rather than external debt, implying that formal credit-market financing constraints are not the primary barrier to entry for most entrants in the sample.

Q11. What are the scope conditions and key limitations?

The sample covers pass-through owners (sole proprietors, partners, S corporation owners) and excludes C corporation shareholders, whose entrepreneurial income does not flow to individual returns until distributed. Income measures exclude most employer fringe benefits; capital gains are excluded from self-employment income, and the authors note their inclusion would strengthen the main findings. The analysis covers 2000–2015 for cohorts born 1950–1975, and income is reported before taxes and transfers. Baseline estimates are not adjusted for misreporting, though BEA-implied adjustments of 46 percent for unincorporated owners and 14 percent for S corporation owners would widen the income gaps further.

Pass-through business owner: An individual who owns a sole proprietorship, partnership, or S corporation, such that business net income flows directly onto the owner’s personal tax return; excludes C corporation shareholders whose income appears only upon dividend or capital-gains distributions.

Tried self-employment: The paper’s primary self-employed comparison group within the “primarily employed” category — individuals with any years in self-employment (including frequent switchers and those with most years in self-employment) — who collectively earn virtually all self-employment income.

Group-specific age effect: The paper’s estimate of how individual income changes with age within a defined subgroup (determined by education, skill, industry, demographics, and employment history), identified by exploiting overlapping birth cohorts in the 16-year panel and separated from individual fixed effects and business-cycle time effects.

Primarily employed: Individuals with at least 12 of 16 sample years in either self- or paid-employment, with at most one intermediate year of non-employment; the paper’s main analytical focus for life-cycle income comparisons.

SOI Databank: The Statistics of Income Databank, a de-identified balanced panel combining SSA demographic records with IRS tax filing data for all living U.S. individuals with a Social Security number over 1996–2015; the paper’s primary data source providing Schedule C, K-1, W-2, and related filing information.

Kelly skewness: A robust measure of distributional asymmetry used by the paper to characterize income growth; the paper reports that Kelly skewness of self-employed income changes exceeds that of paid-employed by approximately 0.1, indicating greater right-skewness in self-employment income dynamics.

Non-pecuniary motivation subgroup: Primarily self-employed individuals who earn less on average than primarily paid-employed peers matched on observable characteristics, taken by the paper as consistent with non-wage job amenities (autonomy, flexibility) driving occupational choice; found to be 57 percent of primarily self-employed by count but earning only 16 percent of total self-employment income.

Optimal Taxation and Market Power

Mon, 01 Jan 0001 00:00:00 +0000

Overview

This paper asks whether and how optimal income taxation should change when firms have market power. The question is motivated by the documented rise in economy-wide markups since 1980, which has compressed the labor share, widened the gap between worker and entrepreneurial income, and generated allocative inefficiency through excessive pricing.

The authors develop a Mirrleesian optimal taxation framework augmented with three features absent from the canonical literature: (i) oligopolistic intermediate goods markets with endogenous, variable markups, (ii) heterogeneous firm productivities, and (iii) two occupational groups—wage-earning workers and profit-earning entrepreneurs—whose abilities are private information. Entrepreneurs strategically set prices under Cournot competition, which means that the tax system affects profits both through a firm’s own behavior and through the responses of its competitors. This strategic interaction is the critical novelty relative to prior work that assumes monopolistic competition.

The main theoretical contribution is the derivation of optimal tax formulas for both labor income and profit income that decompose into four named components: (i) the Mirrleesian incentive component, which reflects the standard trade-off between redistribution and labor supply distortions; (ii) the Pigouvian component, which corrects for the externality from market power by subsidizing labor and entrepreneurial effort to offset the output shortfall from high markups; (iii) the Reallocation Effect (RE), which shifts the profit tax to redirect labor inputs from low-markup firms to high-markup firms where labor is inefficiently scarce, and which emerges only under heterogeneous markups; and (iv) the Indirect Redistribution Effect (IRE), which uses changes in competitors’ product prices—a channel present only under oligopolistic (not monopolistic) competition—to redistribute income between entrepreneurs.

For the labor income tax, the dominant force is the Pigouvian component. As average markups rise, the Pigouvian subsidy to labor supply grows, mechanically reducing optimal labor income tax rates. The profit tax is shaped by all four components in opposing directions; the net quantitative effect is resolved empirically.

The model is calibrated to match distributions of labor income (from the Current Population Survey), profits (from Compustat-based data in De Loecker, Eeckhout, and Unger 2020), and firm-level markups (also from De Loecker, Eeckhout, and Unger 2020, using the cost-minimization approach) for the US in 1980 and 2019. The cost-weighted average markup rose from 1.25 in 1980 to 1.33 in 2019, with the increase concentrated at the top of the markup distribution.

The central quantitative prescription is that the optimal labor income tax rate should decline by 7.7 percentage points between 1980 and 2019 (average optimal rate falls from 22.0 percent to 14.3 percent), while the optimal profit tax rate should rise by 2.2 percentage points on average (from 58.4 percent to 60.5 percent) and by 29.1 percentage points at the top. The decline in the labor income tax is driven primarily by the rise in average markups reducing the Pigouvian component. The increase in the profit tax, especially at the top, is driven primarily by the Mirrleesian component operating through the skill gap, which rises because higher markups reduce profit elasticity. The Pigouvian and reallocation components push in the opposite direction on the profit tax, but the Mirrleesian effect dominates.

The optimal profit tax structure is regressive for large, high-markup firms—reflecting the RE, which requires lower tax rates for high-markup firms to incentivize labor reallocation toward them—but less regressive in 2019 than in 1980, reflecting the distributional tightening from rising markup inequality.

Robustness checks across parameter values for the social welfare curvature k, the span of control ξ, and the elasticity of substitution σ confirm that the directional results hold: labor income tax rates decrease and profit tax rates increase from 1980 to 2019 across all parameter configurations. Extensions to nonlinear sales taxes and conditioning on markups confirm that even when the planner can observe markups directly, the first-best is not achievable because markups are endogenous to entrepreneurs’ unobservable decisions.

In depth

Q1. What is the fundamental difference between this paper’s model and prior work on optimal taxation with market power?

Prior work using monopolistic competition (e.g., Gürer 2021; Boar and Midrigan 2019) assumes each entrepreneur holds monopoly power in its own market, so no strategic interaction exists between firms. Under monopolistic competition, entrepreneurs price to maximize utility given competitors’ choices, and the envelope theorem implies that tax changes have no first-order effect on prices or utility through the pricing channel—the Indirect Redistribution Effect (IRE) disappears. In this paper, entrepreneurs compete in Cournot oligopolistic markets with a finite number of firms I, so each firm’s pricing depends on competitors’ output. A change in one firm’s output (induced by taxation) shifts competitors’ prices, opening a redistribution channel through product markets that is entirely absent in monopolistic competition. Additionally, the Reallocation Effect (RE) emerges only when firm-level markups are heterogeneous, which requires oligopolistic (not perfectly competitive) markets.

Q2. What are the four components of the optimal tax formula and how does each relate to market power?

The optimal tax wedge for both labor and profit income decomposes into four components. First, the Mirrleesian component reflects the standard trade-off between redistribution and the efficiency cost of taxation; in the presence of market power, it is modified because the skill gap for entrepreneurs depends on markups through the profit elasticity. Second, the Pigouvian component corrects the externality from market power, which causes prices to exceed marginal cost and output to be inefficiently low; it implies a subsidy to both worker and entrepreneurial effort, scaled by the reciprocal of the average markup (for the labor tax) or firm-level markup (for the profit tax). Third, the Reallocation Effect (RE) applies only to the profit tax and reflects that labor should be shifted toward high-markup firms where it is inefficiently underemployed; it reduces the tax rate for firms whose markup exceeds the average. Fourth, the Indirect Redistribution Effect (IRE) captures redistribution through competitor price changes under oligopolistic interaction; it can either raise or lower the profit tax rate depending on the distribution of social welfare weights and the cross-inverse demand elasticity.

Q3. What happens to the labor income tax formula as average markups rise?

The labor income tax formula contains a Pigouvian component equal to the reciprocal of the employment-weighted average markup. As average markups rise, this reciprocal falls, reducing the optimal labor income tax rate. Quantitatively, the optimal average labor income tax rate declines from 22.0 percent in 1980 to 14.3 percent in 2019, a decrease of 7.7 percentage points. In a purely competitive benchmark economy, the top labor income tax rate would be around 60 percent (consistent with Saez 2001); in the calibrated model with market power, it is 34.2 percent in 1980 and 28.7 percent in 2019. The Pigouvian component accounts for essentially the entire difference because the Mirrleesian component, when calibrated to the same labor income distribution, is unchanged.

Q4. How does the Mirrleesian component cause the top profit tax rate to rise with market power?

The Mirrleesian component of the profit tax is driven by the skill gap, defined as the proportional rate of change in the composite entrepreneur ability measure. The skill gap depends on markups through the profit elasticity: as markups rise, profit elasticity falls (since profit elasticity is approximately the reciprocal of markup minus the span-of-control parameter minus the inverse of the labor supply elasticity term), which increases the skill gap. A higher skill gap amplifies the income divergence across entrepreneur types, increasing the Mirrleesian incentive to redistribute at the top. Quantitatively, Figure 5 shows that the rise in the skill gap from 1980 to 2019 tracks almost exactly the change in the inverse of profit elasticity, confirming that markup changes—not changes in the ability distribution—are the primary driver of increased Mirrleesian pressure on top profit taxes.

Q5. How does the Reallocation Effect influence the structure (progressivity) of the profit tax?

The RE term equals the ratio of the average markup to the firm-level markup minus one: RE(θe) = μ/μ(θe) − 1. For firms with markups above the average, RE is negative, reducing their optimal tax rate; for firms below the average, RE is positive, increasing it. This implies that the optimal profit tax should be regressive relative to markup (i.e., high-markup firms face lower marginal tax rates), even though the overall profit tax rises on average. This provides a novel rationale for why the profit tax schedule in practice is less progressive—or even regressive—for large firms. As markups rise across the distribution, the reallocation effect pushes down the top profit tax but does not offset the larger increase from the Mirrleesian component in the quantitative exercise.

Q6. What is the Indirect Redistribution Effect and why does it disappear under monopolistic competition?

The IRE captures the change in entrepreneurial utility that arises because a tax reduction for one entrepreneur increases their output, which reduces the prices of substitute goods produced by competitors, thereby lowering competitors’ incomes. Under oligopolistic competition with I > 1 firms per market, the cross-inverse demand elasticity is nonzero, so competitor prices are sensitive to any one firm’s output decision, and this redistribution channel is open. Under monopolistic competition (I = 1), each entrepreneur is the sole producer in its market; competitors’ prices do not depend on the firm’s output, the cross-inverse demand elasticity is zero, and the IRE vanishes by the envelope theorem. The IRE is also absent in perfectly competitive economies. Empirical evidence for the US suggests the hazard ratio of profits is sufficiently high that the IRE generally pushes toward a lower top profit tax rate, but the Mirrleesian effect dominates in the quantitative results.

Q7. What is the quantitative effect of rising markups on the optimal tax rates, and what drives the net change in the profit tax?

The model calibrated to 1980 and 2019 US data prescribes a decline in the optimal average labor income tax rate of 7.7 percentage points (from 22.0 to 14.3 percent) and an increase in the optimal average profit tax rate of 2.2 percentage points (from 58.4 to 60.5 percent). At the top of the profit distribution, the increase is 29.1 percentage points. The net profit tax increase results from four opposing forces: the Pigouvian component falls (pushing toward lower taxes) and the RE decreases for high-markup firms (also pushing down the top rate), while the IRE and especially the Mirrleesian component rise (pushing up top rates). The Mirrleesian effect is the dominant force, driven by rising markup inequality reducing profit elasticity and widening the skill gap for top entrepreneurs.

Q8. How does the counterfactual analysis isolate the role of markups from productivity changes?

The counterfactual fixes the markup distribution at its 1980 level while holding the 2019 productivity distribution constant, then solves for optimal taxes. The result is that high-profit entrepreneurs would face lower optimal tax rates under 1980 markups than under 2019 markups, while low-profit entrepreneurs would face higher rates. Decomposing the difference, the Pigouvian component and the RE are larger for high incomes under 1980 (lower) markups, making the profit tax more regressive, while the IRE and the Mirrleesian component are smaller under 1980 markups, producing a lower top rate. The increase in the Mirrleesian component due to the markup increase from 1980 to 2019 is identified as the primary reason top profit taxes rise. This isolates the markup channel from the productivity channel in accounting for changes in optimal taxes.

Q9. What does the robustness analysis reveal about parameter sensitivity?

The main qualitative result—labor income taxes decline and profit taxes rise from 1980 to 2019—holds across a broad parameter space. The optimal profit tax rate is largely insensitive to the social welfare curvature parameter k: across k ∈ {0.77, 1, 3}, the average optimal profit tax rate is approximately 58 percent in 1980 and 61 percent in 2019. The optimal average labor income tax rate is more sensitive to k: for k = 0.7, 1, and 3, the 1980 rates are 20.3, 26.7, and 44.6 percent, and the 2019 rates are 12.5, 19.4, and 39.1 percent, respectively. Changes in the span-of-control parameter ξ and the substitution elasticity σ do not affect the labor income tax wedge schedule directly but do influence it indirectly through the markup distribution. The directional results are confirmed for all tested parameter configurations.

Q10. What is the role of the “additivity property” from prior externality literature, and why does it fail here?

The additivity property from the Pigouvian externality literature (see Kopczuk 2003; Sandmo 1975) states that the Pigouvian correction is separable from other components of the optimal tax formula, implying that rising markups would simply decrease the optimal tax rate (since 1/μ falls). This property holds under simplifying assumptions that abstract from the general equilibrium and incentive effects of market power. In the present model, the additivity property does not hold because markups enter all four components of the optimal tax formula—not just the Pigouvian term—through the skill gap (Mirrleesian component), the RE, and the IRE. As a result, rising markups can increase the optimal profit tax rate even though the Pigouvian component falls, because the skill gap and Mirrleesian force dominate.

Q11. Can the government attain the first-best by conditioning taxes on markups?

No. The paper demonstrates that even if the planner can observe and condition taxes on firm-level markups, the first-best is not achievable. The reason is that markups are endogenous to the entrepreneurs’ unobservable decisions: an entrepreneur’s markup depends on their privately known type and chosen output. When the planner designs a mechanism that conditions on markup, the incentive constraint facing entrepreneurs remains the same as in the benchmark model, because the promise-keeping constraints are independent of the entrepreneur’s true type when markups are observable. The optimal allocation with markup-conditioned taxes is shown to be equivalent to the second-best with nonlinear sales taxes, which still falls short of the first-best.

Q12. What are the policy implications for the design of the profit tax schedule?

The model yields three concrete prescriptions for the joint design of labor and profit income taxes in the context of rising market power. First, labor income taxes should be reduced and top profit taxes should be increased as market power rises. Second, for large, high-productivity firms the profit tax should be designed to be appropriately regressive to enhance allocative efficiency through the Reallocation Effect—this provides a new normative justification for why profit tax schedules observed in practice are often less progressive than labor income taxes. Third, while profit taxes should be regressive for large firms, the degree of regressivity should decrease as market power rises, reflecting the trade-off between efficiency and equality: higher markups increase the Mirrleesian pressure for redistribution at the top, reducing the optimal regressivity.

Key Concepts

Mirrleesian component (of the optimal tax formula): The standard incentive component of the optimal tax, capturing the trade-off between direct redistribution and the efficiency cost of taxation. In the presence of market power, this component is modified because the skill gap for entrepreneurs depends on markups through the profit elasticity: higher markups reduce profit elasticity, widen the skill gap, and amplify the Mirrleesian force toward higher top profit taxes.

Pigouvian component: The correction in the optimal tax formula for the externality from market power. Because oligopolistic pricing causes output to be inefficiently low, the optimal tax subsidizes both worker and entrepreneurial labor supply. In the labor income tax formula, the Pigouvian component is the reciprocal of the employment-weighted average markup; in the profit tax formula, it is the reciprocal of the firm-level markup. As average markups rise, the Pigouvian component reduces the optimal labor income tax rate.

Reallocation Effect (RE): A component of the optimal profit tax formula that captures the efficiency gain from reallocating labor inputs from low-markup firms (where labor’s marginal product is high relative to value) to high-markup firms (where labor demand is inefficiently low). It equals the ratio of the average markup to the firm-level markup minus one. It implies a lower optimal marginal tax rate for firms with markups above the average, producing a regressive structure in the profit tax for large firms. This effect is absent under monopolistic competition (uniform markups) and in competitive markets.

Indirect Redistribution Effect (IRE): A component of the optimal profit tax formula specific to oligopolistic competition, capturing redistribution through competitor prices. Lowering the marginal tax rate of a high-productivity entrepreneur raises their output, which reduces the prices of substitutable goods produced by their competitors, thereby lowering competitors’ incomes and redistributing toward workers who benefit from lower prices. This effect is present only when the cross-inverse demand elasticity is nonzero—i.e., only under oligopolistic (Cournot) competition with multiple firms per market—and vanishes under monopolistic competition and in the limit as the number of firms grows to infinity.

Skill gap (for entrepreneurs): The proportional rate of change in the composite entrepreneur ability measure with respect to entrepreneur type, analogous to the Mirrleesian skill gap for workers. Under market power, the entrepreneur skill gap depends on the markup through the profit elasticity: as firm-level markups rise, profit elasticity falls, the skill gap increases, and the income dispersion across entrepreneurs widens, which amplifies the Mirrleesian incentive to redistribute at the top and raises the optimal top profit tax rate.

Symmetric Cournot Competitive Tax Equilibrium (SCCTE): The equilibrium concept used in the paper. It is a combination of a tax system, symmetric allocation, and symmetric price system such that all agents (final goods producer, entrepreneurs of each type, workers) are optimizing, strategic interaction in the intermediate goods market is a Cournot Nash equilibrium within each granular market, and all commodity and labor markets clear. Strategic interaction is restricted to within each granular market (firms in the same market compete), so decisions across markets are taken as given.

Composite ability: A combined measure of entrepreneur productivity that determines equilibrium allocations and optimal taxation in the nested-CES economy. It aggregates the entrepreneur’s raw ability (affecting output capacity) and the demand parameter (affecting the market-level markup). The markup-relevant component and the quantity-relevant component are not perfect substitutes in the composite, since equilibrium prices depend on their specific composition while equilibrium quantities depend only on their combined value.

Taxes Depress Corporate Borrowing: Evidence from Private Firms

Mon, 01 Jan 0001 00:00:00 +0000

Layer 1 — Overview

Research Question

Does corporate income taxation raise or lower corporate leverage? The canonical Modigliani-Miller (1963) view holds that the interest tax deduction makes debt more attractive, predicting a positive taxes-to-leverage relationship. Most prior empirical work using large public firms confirms this prediction. This paper re-examines the question using data on small private U.S. firms and finds the opposite: higher corporate taxes depress leverage, at least for small, financially constrained private firms.

Data and Identification

The primary dataset is the Federal Reserve’s Y-14Q supervisory collection (2011–2017), which covers the loan portfolios of the 33 largest U.S. banks and includes firm-level income statements and balance sheets for privately held, bank-dependent borrowers. The sample is restricted to domestic private C-corporations with prior-year assets above $100 million (to screen for pass-through entities), yielding 39,363 non-singleton firm-year observations. The median firm has $288 million in book assets and total debt-to-assets of approximately 38%. A supplementary dataset from the Shared National Credit (SNC) Program (1993–2018, 50,203 firm-year observations) provides a longer time series on syndicated loan commitments. Public firm comparisons use CRSP-Compustat (91,314 observations, 1989–2017).

The empirical strategy is a difference-in-differences event study using variation in state corporate income tax rates. A novel contribution is the manual collection of both enactment dates (when legislation was signed into law) and effective dates for each state tax change since 1975. Identification follows the narrative approach of Romer and Romer (2010) and Giroud and Rauh (2019) to exclude tax changes endogenous to local economic conditions. The specification includes firm and industry-by-year fixed effects, and the analysis uses heterogeneity-robust estimators (Borusyak et al. 2024; de Chaisemartin and D’Haultfoeuille 2020) to address staggered treatment timing.

Main Empirical Findings

For small private firms (below-median total assets, i.e., below $288 million), long-term debt-to-assets rises by approximately 4% in the year of tax cut enactment and remains elevated—at approximately 2%—four or more years later, indicating a permanent increase in leverage. This anticipation effect arises because firms respond to the law’s passage, not its effective date; results using effective dates are noisy and largely insignificant. The average tax cut during the sample period was 1.2 percentage points, representing approximately a 6% reduction in firms’ tax bills (given an average private-firm tax rate of 21%), and the implied leverage change of about 6% at year four is correspondingly large, consistent with a low-interest-rate environment in which small changes in marginal q translate into large investment and borrowing responses.

For large private firms (above-median assets), leverage shows no significant response to tax cuts in any event year. For public firms, evidence of any effect is scant, with at most transient significance and pre-trend issues that complicate interpretation.

Mechanism

The paper argues two tax-sensitive costs of debt offset the standard interest tax shield. First, a higher tax rate reduces after-tax profits, raising default probabilities and credit spreads endogenously; a tax cut thus lowers credit spreads and incentivizes more borrowing. Second, because external equity finance is either unavailable or very costly for small private firms, debt and capital are complements in financing investment: a tax cut raises the marginal product of capital, inducing firms to invest and borrow more. For small firms with low capital adjustment costs, this capital-debt complementarity dominates the direct loss of interest tax shield value. For large firms with high capital adjustment costs (estimated at nine times the small-firm value), investment responds sluggishly to tax changes, the complementarity effect is muted, and the traditional tax shield effect becomes relatively more important—producing the standard, slightly positive taxes-to-leverage relationship.

Bank-assessed default probabilities fall by 20–30 basis points (roughly a 10% decline from an average of approximately 2%) in the year of enactment or one year later for small borrowers, directly supporting the model’s credit spread mechanism.

Welfare Counterfactual

Removing the interest tax deduction from the estimated model (while retaining profit taxation and restricted equity access) causes leverage to fall from 0.36 to −0.26. Firms substitute into cash holdings, shrinking the capital stock. In equilibrium, hours worked rise, the real wage falls, and consumer welfare drops by approximately 1.8%. The interest deduction thus raises welfare in a second-best sense by offsetting other frictions that impede optimal capital accumulation.

In depth

Q1. Why do prior studies find a positive taxes-to-leverage relationship, and how does this paper differ?

Prior studies—including Titman and Wessels (1988), Heider and Ljungqvist (2015), and Faccio and Xu (2015)—predominantly use large public firms, for which the interest tax shield is the quantitatively dominant consideration. The present paper focuses on small private firms that face greater financial frictions (restricted equity access, higher default risk), in which two additional tax-sensitive costs of debt become quantitatively important. A further methodological difference from Heider and Ljungqvist (2015) is the use of firm fixed effects rather than first differences, which the authors argue is appropriate in a staggered DiD design.

Q2. Why use enactment dates rather than effective dates as the event?

Tax legislation is often signed into law one to two years before taking effect; in the sample of 125 tax packages since 1975, 33 became effective the following year and 13 became effective two or more years later. Firms that anticipate future tax changes will adjust leverage immediately upon enactment, not at the effective date. Results confirm this: event studies using enactment dates yield precise positive estimates for small firms (ranging from ~4% at year 0 to ~2% at year 4+), while results using effective dates are noisy and mostly insignificant. The paper therefore treats the enactment date as the economically relevant event and collects these dates as a novel contribution.

Q3. What is the economic magnitude of the leverage response for small private firms?

Small firms’ long-term debt-to-assets rises by almost 4% in the enactment year and remains elevated at approximately 2% four or more years after enactment, consistent with a permanent adjustment. The average tax cut during the period was 1.2 percentage points, representing roughly a 6% reduction in the average tax bill (given an average effective rate of 21% for private firms, per Zwick et al. 2016). The estimated coefficient of 0.021 in year four also implies approximately a 6% change in leverage, a large response that the paper attributes to the low interest rate environment amplifying the marginal q effect of even modest tax changes.

Q4. Do large private firms respond differently to tax cuts, and why?

Large private firms (above the median of $288 million in total assets) show no statistically significant leverage response to tax cuts in any event year, and this null is not attributable to wider confidence intervals. The model estimation explains this via capital adjustment costs: the adjustment cost parameter for large firms is estimated to be nine times larger than for small firms. With high adjustment costs, investment responds sluggishly to a tax cut, so the complementarity channel (more investment requires more debt) is suppressed. The traditional tax shield effect then becomes relatively more important, producing a slightly positive (or zero net) taxes-to-leverage relationship consistent with the large-firm data moment.

Q5. How does the model generate a negative relationship between taxes and leverage when the interest tax deduction is present?

Two mechanisms offset the tax shield. First, higher taxes reduce after-tax profits, pushing firms closer to the default threshold; this is capitalized into equilibrium credit spreads, raising the cost of debt. Specifically, for small firms, the model shows that once leverage exceeds approximately 0.47 of assets, the after-tax risky interest rate rises monotonically with the tax rate (rather than falling via the deduction effect). Second, capital and debt are complements in financing investment: because a tax cut raises the marginal product of capital, and because external equity is unavailable, firms substitute into capital by using more leverage. For small firms with low capital adjustment costs, both mechanisms outweigh the loss of interest tax shield value when taxes fall.

Q6. How are the model parameters estimated, and what are the key parameter values?

The model is estimated by simulated method of moments on the Y-14 small-firm sample, minimizing the distance between nine data moments and their model-simulated counterparts. The nine moments include the means and standard deviations of debt, investment, and operating income (all as ratios of assets), the serial correlations of investment and operating income, and the coefficient from a two-way fixed-effects regression of leverage on a tax-change dummy. The deadweight loss in default (ξ) is estimated at 0.6 for small firms and 0.32 for large firms, consistent with elevated financial frictions for small firms and in line with average recovery rates in Kermani and Ma (2023). Fixed operating costs (f) are approximately 0.15 for both samples, amounting to just under half of steady-state operating profits. The serial correlation of the tax process is estimated at 0.662, with innovation standard deviation of 0.022.

Q7. What is the model’s welfare counterfactual, and what does it imply?

The paper compares two economies both with profit taxation: one with the interest tax deduction and one without. Removing the deduction in the small-firm model causes leverage to fall from 0.36 to −0.26, as firms hold net cash rather than net debt. The capital stock shrinks, output falls, hours worked rise, and both the real wage and consumption decline. Consumer welfare drops by approximately 1.8%. Capital misallocation (measured following Hsieh and Klenow 2009) worsens from 0.89 to 0.88. The result has a second-best character: the interest deduction incentivizes debt-financed investment that partially offsets the distortion from restricted equity access.

Q8. What does the evidence on default probabilities add to the empirical case?

The Y-14 collection contains bank-assessed default probability estimates. In an event study covering Q1 2012–Q4 2018, the authors find that firms’ assessed default probabilities decline significantly by 20–30 basis points in the year of enactment or one year later for small borrowers (those with total loan commitments of $10–$100 million), representing approximately a 10% decline from the sample average default rate of around 2%. This decline peaks two years after enactment and persists for three years. No comparable decline is observed for larger loan size buckets. Separately, in SNC data, the probability of a non-pass (i.e., below-investment-grade supervisory) rating falls by 1.7–2.2 percentage points following tax cut enactments, persisting roughly three years. Together, these findings directly validate the model mechanism by which tax cuts lower default risk and credit spreads.

Q9. Are the results robust to alternative econometric methods that address heterogeneous treatment effects?

Yes. The paper applies the Borusyak et al. (2024) imputation estimator, which imputes fixed effects from untreated observations onto treated observations to remove negative weighting bias; for small firms and event years 0–3, it finds significant positive estimates comparable to the baseline. The de Chaisemartin and D’Haultfoeuille (2020, 2021) estimator, based solely on first-time switchers to treatment, yields an effect of 0.036 on leverage for small firms in the enactment year and no effect for large firms, consistent with the baseline. Results using the narrative approach (excluding Connecticut 2011 and 2015, New York 2014, and Rhode Island 2014 as potentially endogenous) produce slightly larger leverage estimates.

Q10. Are tax hike effects symmetric to tax cut effects?

Evidence on hikes is weaker because tax hikes are rare in the sample. In Y-14 data, hikes are associated with leverage declines for small firms in event year 4 and for large firms in event years 1, 2, and 4, but without sufficient pre-hike observations to identify pre-trends, these results are less credible than the cut results. In SNC data (which spans a longer period, 1992–2018), tax hikes are associated with large and significant reductions in total syndicated borrowing commitments of 6–7%, while cuts produce smaller and marginally significant increases. This asymmetry is consistent with the lower adjustment costs of reducing debt relative to increasing it.

Q11. What does the analysis of alternative model specifications reveal about the generality of the mechanism?

Three model extensions are considered. In a collateral-constrained model (no endogenous default), the cost of debt is lost financial flexibility (the future shadow cost of the borrowing constraint), which remains tax-sensitive. In a model with costly equity issuance (linear cost λ = 0.11 following Hennessy and Whited 2007), equity issuance is rare, so the model behaves nearly identically to the baseline. In a solvency-based default model (default when firm value turns negative rather than when liquidity is insufficient), the negative taxes-to-leverage result is preserved. A news-shock extension (Jaimovich-Rebelo 2009) incorporating the anticipation of future tax changes also produces lower leverage in response to higher anticipated taxes, consistent with the empirical anticipation effects, though with smaller magnitudes because the news shock variance is smaller than the total tax-change variance.

Q12. Why do contingent-claims models (Fischer-Leland-Goldstein class) always predict a positive taxes-to-leverage relationship?

In these models, shareholders have deep pockets, so negative cash flows can always be covered; this implies default is rare and the effect of taxes on the default put value is small relative to the direct interest tax deduction. Additionally, these models contain no capital stock, so there is no substitution mechanism between capital and a storage technology (i.e., cash/negative debt). Without endogenous investment, the only channel linking taxes to leverage is the tax shield, which necessarily implies a positive taxes-to-leverage relationship. This is why, as the paper notes, the result was “already hiding” in the Hennessy-Whited class of dynamic investment models but not visible in the contingent-claims literature.

Key Concepts

Interest Tax Deduction (Tax Shield) The paper uses this in the standard corporate finance sense: the after-tax cost of debt is reduced because interest payments are deductible against corporate income. In the model, debt proceeds are discounted at the after-tax interest rate, and the deduction is taken at the time of debt issuance. The paper’s contribution is to show this benefit can be outweighed by two tax-sensitive costs of debt, reversing the sign of the taxes-to-leverage relationship for small, constrained firms.

Tax-Sensitive Cost of Debt The paper defines two distinct tax-sensitive costs that offset the tax shield. First, taxes reduce after-tax profits, shifting the default threshold and raising equilibrium credit spreads; this is capitalized into the risky lending rate endogenously from the lender’s zero-profit condition. Second, taxes reduce the marginal product of capital, making debt-financed investment less attractive; because debt and capital are complements in a model without external equity, a higher tax rate lowers optimal capital and, with it, optimal debt.

Capital Adjustment Costs (ψ) Quadratic costs of changing the capital stock, parameterized as ψ(k’ − (1−δ)k)² / (2k). The paper identifies this parameter as the key determinant of whether leverage responds positively or negatively to taxes: for small firms, ψ is estimated to be near zero (insignificantly different from zero), enabling free substitution between capital and the storage technology (negative debt), so the complementarity channel dominates. For large firms, ψ is estimated to be nine times larger, suppressing this substitution.

Default Threshold In the model, default is triggered when the firm’s current after-tax profits plus recoverable capital are insufficient to repay debt: (1−τ)(y − wn − f) + (1−ξ)(1−δ)k < p. This threshold depends directly on the tax rate τ, so higher taxes move the threshold in the direction of default, raising credit spreads. The paper provides empirical support for this mechanism via the event study of bank-assessed default probabilities.

Enactment Date vs. Effective Date The paper distinguishes between the date tax legislation is signed into law (enactment date) and the date it becomes operative (effective date), which can differ by one to two years. The paper collects novel data on enactment dates from state legislative records. The empirical finding that firms respond to enactment rather than effective dates constitutes evidence of anticipation effects: firms adjust leverage upon observing future expected tax changes, not when the changes actually take hold.

Second-Best Welfare Effect of the Tax Deduction The paper uses this term to characterize the welfare result from the counterfactual: in an economy already distorted by profit taxation and restricted equity access, the interest deduction raises consumer welfare by incentivizing debt-financed capital accumulation. Removing the deduction causes firms to substitute into cash, shrinking the capital stock and lowering wages and consumption. This is a second-best result because the deduction is welfare-improving only because it partially offsets the distortions created by other frictions; in a frictionless world, no such second-best rationale would apply.

Y-14Q Supervisory Data The Federal Reserve’s supervisory collection from the 33 largest U.S. banks, covering loan portfolios and associated borrower financial statements for firms with commercial and industrial loans exceeding $1 million in commitment. The paper uses this dataset because it covers private, bank-dependent firms—a population not previously studied in the tax-leverage literature—and contains firm-level balance sheets, credit ratings, and default probability estimates.

The Welfare and Distributional Consequences of Corporate Tax Cuts in Open Economies

Mon, 01 Jan 0001 00:00:00 +0000

Layer 1: Overview

This paper uses an open-economy heterogeneous-household model with incomplete markets to evaluate the welfare and distributional consequences of the U.S. Tax Cuts and Jobs Act (TCJA) of 2017 — which reduced the U.S. corporate tax rate from 35 to 21 percent — both within the U.S. and in affected trading partners. The model features three economies (the U.S., a small open economy calibrated to Canada, and the rest of the world), free capital flows, progressive income taxes, and idiosyncratic uninsurable labor income shocks generating empirically realistic wealth Gini coefficients (0.80 for the U.S., 0.70 for Canada). Three main results are established. First, the TCJA is regressive in the U.S. — under a permanent cut, the bottom 5 percent of U.S. households by wealth experience welfare losses of 0.10–0.26 percent of lifetime consumption, while the top 1 percent gain 0.92 percent — and generates an even more regressive outcome in trading partners, where approximately the bottom 80 percent of the small open economy’s wealth distribution experience welfare losses averaging 1.28 percent at the bottom decile against gains of 2.57 percent at the top. Second, whether U.S. wealth-poor households benefit depends critically on the persistence of the tax cut: under a permanent cut, households above approximately the bottom 5 percent of the U.S. wealth distribution gain (driven by wage increases from capital inflows), but under an anticipated partial reversal from 21 to 28 percent after 7 years, approximately the bottom 75 percent of U.S. households experience welfare losses because the temporary wage gain is dominated by a persistent increase in the public debt burden. Third, when the small open economy reciprocates by matching the U.S. corporate tax reduction to 21 percent, the domestic distributional consequence reverses: all wealth quintiles in the small open economy gain (Table 5, Panel B shows gains of 0.52–1.19 percent across all groups), with the gain being roughly progressive within the SOE — a result driven by the wage increase from capital inflows exceeding the financing cost, which falls primarily on the wealth-rich through higher top marginal tax rates.

In depth

Q1. What is the model structure and how are the three economies connected?

The paper extends the Aiyagari (1994) incomplete-markets heterogeneous-household model to an open-economy setting with three countries — the U.S., a small open economy (SOE) modeled as Canada, and the rest of the world (ROW) modeled with Canadian parameters — linked by free capital flows that equalize after-tax returns to capital across countries: (1 − τc^US)r^US = (1 − τc^SOE)r^SOE = (1 − τc^ROW)r^ROW = r^b. Households in each economy face idiosyncratic uninsurable productivity shocks (three states: low s₁ = 0.167, medium s₂ = 0.839, high s₃ = 5.087, with a persistent Markov transition matrix following Domeij and Heathcote 2004) and save in internationally traded capital and government bonds, subject to borrowing constraints calibrated to match wealth Gini coefficients. The fiscal rule follows Bohn (1998) with the residence-based tax revenue responding to the debt-to-GDP ratio to ensure stationarity, and the top marginal tax rate τ₁ adjusts endogenously when corporate tax revenues change (consistent with Mertens and Montiel Olea 2018’s evidence on tax instrument choice). The SOE size is 10 percent of the U.S., enabling the paper to capture the asymmetric spillover mechanism by which U.S. corporate tax policy creates large distributional consequences abroad without generating offsetting fiscal adjustments in the SOE.

Q2. What are the distributional effects of the permanent TCJA in the U.S. and SOE?

Under a permanent reduction from 35 to 21 percent in the U.S. corporate tax rate, the average U.S. welfare gain is +0.146 percent of lifetime consumption, but this masks a strongly regressive distribution: households in the bottom 5 percent of the U.S. wealth distribution experience welfare losses of −0.045 to −0.101 percent, while those in the top 1 percent gain +0.920 percent, with gains monotonically increasing through the wealth distribution above the 5th percentile; in the SOE, the average welfare effect is −0.392 percent, and the losses are far larger and more broadly distributed, with approximately the bottom 80 percent (up to the 75th–95th percentile boundary) experiencing losses ranging from −0.582 to −1.282 percent while the top 1 percent gains +2.566 percent (Table 3). The mechanism for the U.S. involves capital inflows that raise wages (benefiting labor-income-reliant poor households at least partially) offset by increased tax burden from debt accumulation; in the SOE, capital outflows depress wages more severely, and wealth-rich households in the SOE gain even more than their U.S. counterparts because SOE households face no increase in their domestic tax burden to finance the U.S. corporate tax cut, making the SOE spillover a “free lunch” for SOE capital owners.

Q3. Why does the permanence of the tax cut matter for lower-wealth U.S. households?

When the TCJA is anticipated to be partially reversed — from 21 percent back to 28 percent after 7 years — households in approximately the bottom 75 percent of the U.S. wealth distribution experience welfare losses averaging from −0.091 to −0.259 percent; under a permanent cut only approximately the bottom 5 percent suffer losses, so the reversal shifts the crossover point from the 5th to the 75th percentile of the wealth distribution (Table 4, Panel A). The mechanism is that under a temporary tax cut the capital inflow is short-lived and so the wage increase is limited in duration, while the increase in U.S. government debt is persistent — because the government finances the cut through debt issuance and the debt level remains elevated even after the reversal from 21 to 28 percent, the resulting higher tax burden on labor income persists and dominates the temporary wage benefit for wealth-poor households who primarily earn labor income. This result has a direct policy implication: the distributional case for extending or making permanent the TCJA’s corporate rate reduction is much stronger than for a time-limited cut, because the wage-raising channel — the main argument for the cut’s benefits to workers — operates only persistently.

Q4. What happens when the small open economy reciprocates with its own corporate tax cut?

When the SOE reduces its corporate tax rate to match the U.S. at 21 percent (from 38 percent) simultaneously with the TCJA, all wealth groups in the SOE experience welfare gains (Table 5, Panel B shows average gains of +0.524 to +1.190 percent across wealth groups), with the distributional effect being progressive within the SOE: the incremental gain from reciprocation compared to not reciprocating is +1.807 percent for the bottom 1 percent of the SOE wealth distribution and −1.376 percent for the top 1 percent (Table 5, Panel C). The reason the SOE reciprocation is progressive is that the capital inflow triggered by the SOE’s cut raises wages across the SOE (benefiting labor-income-reliant poor households), while the financing cost of the cut — through debt accumulation and the eventual increase in top marginal tax rates — falls disproportionately on wealthy households. The paper notes this result depends on the SOE’s small size: because the SOE is only 10 percent of the U.S., its corporate tax cut creates a better investment opportunity for all global capital owners but the financing cost falls entirely on SOE residents, creating a distributional asymmetry between who benefits (all capital owners globally) and who pays (SOE income-rich households domestically).

Q5. How does the model fit the pre-TCJA data and what are the calibration targets?

The model closely matches its calibration targets: capital-to-output ratios of 2.50–2.52 (target 2.50), debt-to-GDP ratios of 0.827–0.882 (targets from Jordà-Schularick-Taylor 2017), and wealth Gini coefficients of 0.82 for the U.S. (target 0.80, from Budría-Rodríguez et al. 2002) and 0.71 for the SOE (target 0.70, from Brzozowski et al. 2010); and generates an untargeted prediction that the U.S. is a net borrower and Canada a net lender, consistent with data (Table 2, Panel B). The discount factors are calibrated to β^US = 0.968 and β^SOE = 0.969 to match the capital-output ratio, and the borrowing constraints are set at ψ = −1.65 for the U.S. and ψ = −0.88 for the SOE/ROW to match their respective wealth Gini coefficients. The model abstracts from terms-of-trade effects (consistent with Hanson et al. 2021’s evidence that US-Canada terms of trade are unaffected by US corporate tax changes) and aggregate uncertainty beyond corporate tax changes, and the SOE is set at 10 percent of the U.S. economy by population size.

Q6. How do the results change under alternative fiscal financing assumptions?

The key qualitative results — regressivity of the TCJA in the U.S. and its greater regressivity in the SOE — are robust across alternative fiscal financing assumptions: when the corporate tax cut is financed by immediately increasing the residence-based tax (χ = 1) rather than by debt (χ = 0 in the baseline), the losses at the bottom of the U.S. distribution become larger (approximately the bottom 70 percent lose rather than the bottom 5 percent), and when progressivity of the income tax (τ₃) rather than the top marginal rate (τ₁) adjusts, the additional tax burden falls more on wealth-poor households, making the cut even more regressive. The SOE reciprocation result is also robust: Appendix C.3 shows that financing the SOE corporate tax cut through increases in the residence-based tax (χ^SOE = 1) reduces the welfare gains for all SOE households but preserves the progressive distributional pattern within the SOE, while appendices C.1–C.2 show that the results are linear in the size of the SOE’s tax cut (at 30 and 18 percent, the distributional pattern is similar in direction).

Key Concepts

open-economy Aiyagari model : the paper’s framework — an extension of the Aiyagari (1994) incomplete-markets model with heterogeneous households and idiosyncratic uninsurable labor shocks to an international setting with free capital flows — used to capture how corporate tax changes distribute welfare across the wealth distribution in multiple countries simultaneously.

consumption equivalent variation : the proportional change in lifetime consumption required to make a household in the counterfactual no-TCJA economy as well off as in the economy with the TCJA; the welfare metric used in Tables 3–5, measured in percent of lifetime consumption, conditional on wealth and productivity state at the time of implementation.

TCJA persistence channel : the mechanism by which the distributional effect of the corporate tax cut for lower-wealth U.S. households depends on whether the cut is permanent: a permanent cut sustains capital inflows and wage gains long enough to dominate the increased tax burden, while a temporary cut leaves only a persistent debt overhang with limited wage benefits, turning even the bottom 75 percent of U.S. households into net losers.

SOE reciprocation progressivity : the finding that a small open economy that matches the U.S. corporate tax reduction achieves a progressive domestic distributional outcome because the wage increase from capital inflows benefits all households but the financing cost (through higher top marginal tax rates) falls mainly on the wealthy; this mechanism is size-dependent and reverses the regressivity that the U.S. cut generates domestically.

Summary of a forthcoming paper, AI-assisted. Draft pending human review. See the linked original for the authoritative claims and full conditions.

Who's Afraid of the Minimum Wage? Measuring the Impacts on Independent Businesses Using Matched U.S. Tax Returns

Mon, 01 Jan 0001 00:00:00 +0000

Layer 1 — Overview

Research Question

This paper asks how independent (pass-through) businesses in the United States accommodate minimum wage increases — specifically whether they reduce employment, compress profits, pass costs through to customers, or exit — and what happens to the low-earning workers and business owners affected by these adjustments.

Data and Methodology

The authors construct a novel linked firm-worker-owner panel dataset from the universe of U.S. tax returns, covering approximately 235,000 pass-through firms (S-corporations, partnerships, and LLCs) per year in highly exposed industries over 2010–2019. “Highly exposed” industries are defined as those where at least 15% of workers earned below the full-time equivalent of the federal minimum wage ($15,080 per year) in 2013. The dataset links annual business income tax returns to the individual income tax returns and W-2 information reports of all workers and owners.

The causal identification strategy exploits the six state minimum wage increases that took effect in 2014 (California, Connecticut, Delaware, Michigan, Minnesota, and New Jersey) relative to 24 states that did not change their wage floors at any point from 2012–2018. The empirical workhorse is a panel difference-in-differences event study (Equation 1), augmented by DFL re-weighting (DiNardo et al., 1996) to improve comparability of treatment and control firms on observables. The analysis covers cumulative effects through 2018, by which point the average minimum wage across treatment states had risen 30.6%.

Main Findings

Employment: The average exposed independent firm does not meaningfully reduce employment. The authors estimate an own-wage elasticity of -0.209 (s.e. = 0.0112). Employment adjustments manifest as moderately lower hiring rather than layoffs of existing workers. Reduced hiring is wholly concentrated among teenagers and very part-time jobs paying less than $3,900 annually (with 67% earning less than $1,000 per year).
Worker earnings: Despite the hiring reduction, low-earning workers employed at exposed independent firms experience average earnings gains of approximately $2,000 per year by 2018, relative to comparable workers in untreated states. Young individuals aged 20–26 without a 2013 job earn roughly $4,000 more per year by 2018; teenagers without a 2013 job gain approximately $1,000 per year. Workers in these groups are no less likely — and in some cases slightly more likely — to be employed five years after the minimum wage increase.
Wage bills: Average wage bills among surviving treated firms rose 7.03% (s.e. = 0.0153) by 2018. Earnings gains are concentrated among workers earning $15,600–$35,000 annually, with no evidence of reduced earnings for higher-paid workers. The 7% average wage bill increase amounts to only 1.4% of 2013 firm revenues, easing pass-through.
Revenue and profits: Revenues of surviving treated firms grew approximately 2.1% more than control firms by 2018. On average, this revenue increase fully offsets the higher wage bill, yielding a small net profit increase of roughly $3,360 (s.e. = $1,123) per owner by 2018, or about 2.7% of mean 2013 owner income.
Firm exit: On average across all highly exposed industries, minimum wages increased the five-year exit probability by 0.9 percentage points (s.e. = 0.0029), relative to a baseline raw exit rate of approximately 29%. Exit effects are driven entirely by restaurants: by 2018, restaurants in treated states were 1.85 percentage points (s.e. = 0.0039) more likely to have exited, while the exit response for non-restaurant exposed firms is a precisely estimated zero.
Heterogeneity by productivity within restaurants: Exit is concentrated entirely in the bottom productivity quartile (coefficient = 0.0254, s.e. = 0.0079), with no significant effect in the upper three quartiles. Profits among surviving small restaurants rise by $5,941 (s.e. = $1,546) by 2018 relative to 2013. Among small restaurants, the profit gains are larger for firms in the higher productivity quartiles (Q3: +$7,915; Q4: +$9,161). Surviving restaurants also increase non-labor input spending by 2.53% (s.e. = 0.0101), consistent with expanded output following competitor exits.
Entrant characteristics: Post-reform restaurant entrants in treatment states have higher wage bills (13.8% higher in logs), higher revenues (4.0% higher), higher value-added (8.4% higher), and higher productivity (net income/revenue ratio 2.24 percentage points higher) than entrants in control states, indicating the minimum wage raises the productivity floor for new entrants.
Owner outcomes after exit: Owners of small restaurants forced out by the minimum wage are significantly less likely to own an independent business five years later, but earn no less on average in wages plus business income. Policy-induced exiters are significantly less likely to report negative incomes, suggesting substitution away from risky or marginally profitable business ownership.

Theoretical Framework

The authors present a Cournot competition model with heterogeneous firm productivity and fixed production costs. A minimum wage cost shock raises marginal costs, narrowing margins for all firms. Firms whose cost increases exceed the market price increase cannot cover fixed costs and exit. Remaining firms gain higher markups and larger market shares as demand is reallocated from exiting firms. Selection on ex-ante productivity (the least productive firms exit) limits the distortion to market quantity and amplifies profit gains among productive survivors. The model predicts profit increases only in markets with firm exit, which matches the data: profits rise among restaurants (where exit occurs) but not among retailers (where exit is a precisely estimated zero).

Scope Conditions

Findings pertain to the short-to-medium run (up to five years post-legislation) of phased-in minimum wage increases averaging 30.6% in six U.S. states. The sample covers pass-through (independent) businesses in highly exposed industries. Longer-run effects may differ if entrants adopt production technologies that rely less on low-wage labor or incumbents reconfigure inputs. Border-county retailers appear to be less able to pass through costs than interior firms, suggesting product market competition is a key moderating factor.

In depth

Q1. Why do the authors focus on pass-through businesses rather than publicly traded corporations?

Pass-throughs (S-corporations, partnerships, and LLCs) comprise 78% of non-sole-proprietorship businesses and 79% of firms with fewer than 20 employees. They represent the majority organizational form for independent businesses in virtually all two-digit NAICS industry groups except utilities and enterprise management. Because minimum wage concerns are disproportionately raised on behalf of small independent businesses, and because most minimum wage workers in restaurants are employed at pass-throughs, studying pass-throughs directly addresses the policy debate. Additionally, pass-through tax returns link business income directly to the individual tax returns of each owner, enabling the authors to separately identify employee versus owner responses.

Q2. How do the authors define “highly exposed” industries and why does this matter for identification?

Highly exposed industries are defined as four-digit NAICS industries where at least 15% of workers earned below the full-time federal minimum wage equivalent ($15,080 per year) in 2013, using tax data to construct a proxy for minimum wage workers. The analysis focuses on these industries because minimum wage workers are extremely concentrated — the vast majority are in Leisure/Hospitality and Retail. Restricting to highly exposed industries allows the authors to estimate average effects within affected markets and conduct heterogeneity analysis across firm characteristics within those markets, including comparing firms with different baseline shares of low-earning workers that nonetheless all face the market-level cost shock.

Q3. How do the employment effects decompose into hiring versus retention?

The average firm subject to a higher wage floor does not lay off existing workers (the retention line is flat in event study estimates). By 2018, firms in treated states hire roughly one fewer worker on average than similar firms in control states, entirely through reduced hiring. This reduced hiring is wholly concentrated among teenagers in very part-time jobs: the missing hires consist entirely of workers who would have earned less than $3,900 annually, with 67% earning less than $1,000 per year. Simultaneously, workers already employed at exposed firms are 2 to 4 percentage points more likely to remain with their 2013 employer by 2016, with prime-age low-earning workers exhibiting the largest retention increases.

Q4. What happens to low-earning workers and young people in individual-level panels?

Low-earners (those earning below $25,000 in each year from 2012–2014) at exposed independent firms experience average earnings gains of approximately $2,000 per year by 2018 relative to similar workers in untreated states, including teenage low-earners. Young individuals aged 20–26 with no job in 2013 experience a relative earnings increase of approximately $4,000 per year by 2018; teenagers without jobs in 2013 gain approximately $1,000 per year. These workers are no less likely — and often slightly more likely — to be employed relative to their counterparts in control states, so the earnings gains are not offset by employment losses at the individual level.

Q5. What is the magnitude of the cost shock for firms and how does it compare to revenues?

By 2018, the average wage bill among surviving firms in treated states was 7.03% (s.e. = 0.0153) higher than comparable firms in control states. This is consistent with a back-of-envelope calculation: low-earning workers account for about 21% of wage bills at these firms, and states raised minimum wages by 30.6% on average (0.21 × 0.306 = 0.064). However, the 7% wage bill increase amounts to only approximately 1.4% of 2013 firm revenues, making cost pass-through relatively modest. Higher minimum wages have no discernible impact on pension contributions but slightly reduce deductions for other benefits including health insurance.

Q6. How do surviving firms finance the increased wage bill, and what happens to profits?

Surviving firms finance the wage increase primarily through higher revenues. By 2018, revenues of firms in treated states grew approximately 2.1% more than revenues of firms in control states. On average, this revenue increase outpaces the higher wage bill, resulting in a net profit increase of approximately $3,360 (s.e. = $1,123) per owner by 2018, representing about 2.7% of mean 2013 owner income. There is no evidence of redistribution from middle- or high-income workers within firms; wage bill increases are concentrated among workers earning $15,600–$35,000 annually, consistent with minimum wage spillovers to workers slightly above the statutory floor.

Q7. Why do restaurants experience exit effects but retailers do not?

The asymmetry stems from the intensity of low-wage labor in production. While low-earning workers account for a similar share of labor costs at restaurants (41.8%) and retailers (38.5%), labor costs overall are more than twice as large at restaurants relative to retailers. Wage bills account for 39% of variable costs and 27% of revenues at restaurants, but only 16% of variable costs and 13% of revenues at retailers. As a result, raising the minimum wage raises variable costs by 5.76% at restaurants. Non-restaurant exposed firms are able to fully pass through their smaller cost shock, yielding flat profits and neither employment nor exit impacts.

Q8. Why is firm exit concentrated in the lowest productivity quartile of restaurants rather than among the most exposed firms?

The Cournot framework predicts exits among firms with the lowest ex-ante productivity (highest marginal costs), the largest cost shock (highest share of low-wage labor per unit of output), or a combination. Empirically, productivity is the primary determinant: restaurants across all productivity quartiles use similar shares of low-earning workers (40–44% of wage bills for Q1 through Q4). Exit rises significantly only among restaurants in the bottom productivity quartile (coefficient = 0.0254, s.e. = 0.0079), with no significant effects in Q2–Q4. Among the lowest-productivity restaurants, those most dependent on low-earning labor face the largest exit rates.

Q9. How do the model’s predictions about profit heterogeneity match the data?

The Cournot model predicts profits should rise only in markets with firm exit (via increased margins and market share reallocation to survivors). This is exactly what the data show. Among restaurants, where exit is concentrated in the bottom productivity quartile, profits among surviving small restaurants rise by $5,941 (s.e. = $1,546) by 2018. Among small restaurants specifically, profit gains increase with productivity: Q3 restaurants gain $7,915 (s.e. = $3,326) and Q4 restaurants gain $9,161 (s.e. = $2,127), while Q1 and Q2 gains are statistically indistinguishable from zero. In non-restaurant exposed industries where the exit effect is a precise zero, profits are also flat — exactly as the model predicts.

Q10. What happens to the characteristics of new restaurant entrants after the minimum wage increase?

Post-reform restaurant entrants in treatment states are systematically more productive than entrants in control states. They have wage bills 13.8% higher (in logs), revenues 4.0% higher, value-added 8.4% higher, and productivity ratios (net income/revenue) 2.24 percentage points higher than new entrants in control markets. This implies the minimum wage raises the minimum viable productivity threshold for entrant restaurants, consistent with Sorkin (2015)’s insight that minimum wages shape the capital and technology choices of entering firms. The restaurant industry thus becomes more productive on average through both the exit of the least productive incumbents and the entry of more productive new firms.

Q11. How do worker transition patterns reflect the reallocation of output to surviving firms?

Workers at large independent businesses (top revenue quartile) are 3.52 percentage points more likely to remain with their 2013 employer in 2018 and 2.36 percentage points less likely to switch to another large firm. The large firms that retain more of their existing workforce also reduce their hiring of very part-time teenagers the most — in the top revenue quartile, firms shed roughly 4.5 employment relationships on average, comprising higher retention of 4.15 existing workers offset by reduced hiring of 8.67 very part-time teenage workers. Workers originally at smaller exposed firms are more likely to be found working at larger firms five years out, consistent with demand reallocation from exiting and shrinking small firms toward larger, more productive survivors.

Q12. What happens to owners of restaurants that exit due to the minimum wage?

Policy-induced exiters of small restaurants are significantly less likely to own an independent business five years later and less likely to receive all earnings from business ownership, relative to owners of restaurants that exited for other reasons in control states. However, their average incomes (wage income plus ordinary business income) are no lower. This income stability is partly explained by the fact that policy-induced exiters are significantly less likely to report negative incomes five years out, suggesting they substitute away from potentially risky or marginally profitable business ownership toward wage employment or other activities. The utility implications are ambiguous: these former owners may have preferred business ownership even if it did not yield higher income.

Q13. What is the role of product market competition in mediating pass-through, as evidenced by border-county analysis?

The border county robustness analysis reveals that product market competition is central to pass-through success. Retailers near state borders, where consumers can cross-state-border shop, face more elastic demand and are less able to finance the wage cost shock with new revenues, exhibiting reduced profits and higher exit rates (though estimates are imprecise). Further from the border, where the cost shock is more commonly felt by all potential substitutes (making market demand elasticity rather than firm demand elasticity the relevant parameter), results are very similar to the full-sample aggregate findings. This confirms that the common nature of the minimum wage cost shock — shared by all competing firms in the market — is a key reason firms can pass through costs to consumers.

Q14. How do the findings address the divide among independent business owners on minimum wage policy?

The heterogeneous outcomes rationalize why surveys consistently find business owners divided. Among restaurants, some owners (those operating the least productive small restaurants) face exit and loss of business ownership, while surviving productive restaurateurs see higher profits of $5,941–$9,161 per year. Among non-restaurant exposed businesses, owners are broadly unaffected in terms of profits and viability. Uncertainty about whether a given firm’s demand is elastic enough to bear cost pass-through — given that owners may be more familiar with the elasticity of firm-level demand from prior unilateral price changes, rather than the relevant market-level demand elasticity applying to a common cost shock — may broaden opposition to include even owners who would ultimately benefit.

Key Concepts

Pass-through businesses (independent businesses): Privately owned firms organized as S-corporations, partnerships, or LLCs, taxed by passing income through to the individual returns of owners rather than at the entity level. In 2015, these comprised 78% of non-sole-proprietorship U.S. businesses and 46% of employment. The paper uses “pass-through” and “independent business” interchangeably as the unit of analysis.

Highly exposed industries: Four-digit NAICS industries where at least 15% of workers earned below the annual full-time equivalent of the federal minimum wage ($15,080) in 2013, as measured in the authors’ administrative tax data. This threshold proxies the concentration of minimum-wage workers across industries and drives the sample selection for firm-level analysis.

Own-wage elasticity of employment: The estimated percentage change in employment at a firm associated with a given percentage change in the firm’s minimum wage. The authors estimate this as -0.209 (s.e. = 0.0112), reflecting the average effect across all exposed independent businesses, conditional on the firm’s industry, size, and local market characteristics.

DFL re-weighting (DiNardo-Fortin-Lemieux): A non-parametric reweighting procedure that adjusts the distribution of control-group firms to match the distribution of treatment-group firms on observables (specifically, two-year lagged value-added within three-digit NAICS industries). Used to improve pre-reform comparability of treatment and control firm samples without parametric functional form assumptions.

Firm productivity (in this paper’s sense): Measured as the ratio of net profits to revenues (net income/revenue) at the firm level in the base year 2013, used to assign firms to productivity quartiles for heterogeneity analysis. This is a firm-level profitability measure constructed from pass-through tax returns, not a total factor productivity estimate requiring production function estimation.

Firm exit: An indicator for a firm that filed a tax return in 2013 but did not file a return in a subsequent year t. The average one-year exit rate for highly exposed independent businesses is 5.2%; the cumulative five-year raw exit rate is approximately 29% across treatment and control states.

Cournot competition with heterogeneous productivity and fixed costs: The paper’s conceptual framework, in which N firms compete in quantities with asymmetric marginal costs (reflecting heterogeneous productivity), a common output price, and a fixed cost of production. Under this framework, a minimum wage cost shock narrows margins unevenly, induces exit among firms that cannot cover fixed costs, and generates both demand reallocation and market share gains for productive survivors — rationalizing simultaneous exit and profit increases in the same industry.

Common cost shock: The property that a minimum wage increase raises production costs for all firms employing low-wage workers in the same market simultaneously. Because all competing firms face higher costs, the relevant pass-through parameter is the elasticity of market demand rather than the (higher) elasticity of individual firm demand, facilitating cost pass-through to consumers and distinguishing minimum wages from unilateral price changes by a single firm.