J65 | Macro Paper Warehouse

Disincentive effects of unemployment insurance benefits

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

This paper isolates the disincentive effects of pandemic unemployment insurance (UI) benefits on employment recovery, separating them from the simultaneously operating stimulative (demand) effects that previous studies conflate. The authors study the largest UI expansion in U.S. history — the CARES Act of March 2020 — which introduced three simultaneous provisions: a $600 weekly income supplement (FPUC) through end of July 2020, a 13-week extension of maximum benefit duration (PEUC), and expanded eligibility to workers previously ineligible for UI (PUA), together raising the median replacement rate to 145% and more than doubling the number of UI recipients.

The empirical strategy uses high-frequency establishment-level data from Homebase (HB), a scheduling and payroll provider covering approximately 140,000 small U.S. businesses — predominantly restaurants and retailers — matched to Yelp price-tier data and Safegraph foot-traffic and spending data. The final estimation sample is 4,595 businesses within 1,195 local-industry cells, observed at weekly frequency from January 2019 to December 2020.

The identification rests on comparing employment recovery of low-wage versus high-wage businesses within the same narrow local labor market (four-digit zip code), industry (two-digit NAICS), and price tier. Because neighboring businesses largely share the local demand stimulus from UI, differencing within local-industry cells removes common demand effects. The key variation is the expiration of the $600 supplement, which differentially compresses the replacement-rate gap between low- and high-wage businesses depending on local average wages — labor markets where the gap falls more sharply are the treated group.

The main empirical finding is that a 100 percentage point decline in the replacement rate gap is associated with a 5.7 percentage point rise in low-wage business employment recovery relative to high-wage business employment recovery at 12 weeks after the $600 expiration. For the average labor market, the expiration of the $600 supplement decreased the replacement rate gap by 46 percentage points, implying a 2.6 percentage point closing of the low-versus-high-wage employment gap within 12 weeks. Importantly, hours per employee and hourly wages grew faster in low-wage businesses over the same period, consistent with a labor supply rather than a demand mechanism. When the comparison is conducted at the U.S. state level rather than within local-industry cells — as in Finamor and Scott (2021) — the effect disappears and reverses sign, illustrating how local demand effects obscure disincentive effects at broader geographic aggregations.

To quantify the aggregate employment impact, the authors build and calibrate a McCall-style labor search model with heterogeneous firm wages, a UI-eligible and non-UI unemployed pool, and equilibrium reservation wages. The model is extended to include a probability (calibrated at 16.5%) that workers lose UI eligibility upon refusing a job offer, which reconciles the model with the empirical estimates; without this feature the baseline model substantially overstates the differential employment effect of the $600 expiration.

The full model-implied aggregate employment loss from all CARES Act UI provisions combined is 3.4 percentage points on average between April and December 2020, representing approximately 20% of the average employment shortfall in the Leisure and Hospitality sector over that period. When each provision is implemented in isolation, the effects are modest ($600 supplement: 0.2 pp; extended duration: 0.2 pp; expanded eligibility: 1.0 pp), but their interaction generates the large combined effect. Expanded eligibility is identified as the most disruptive provision, particularly for low-wage businesses, because it depletes the pool of non-UI unemployed who are the primary source of hires for these firms. The unemployment duration elasticities implied by the model are modest and in line with the low-to-middle range of pre-pandemic estimates.

The paper’s scope is restricted to the disincentive channel and deliberately excludes the stimulative effects of UI; it studies small, in-person service sector businesses and the April–December 2020 recovery period only.

Q: What is the core identification challenge this paper addresses? A: Prior empirical studies find only modest net effects of pandemic UI on employment, but it is unclear whether this reflects small disincentive effects or the near-cancellation of two opposing forces — UI suppressing labor supply while simultaneously stimulating local consumer demand. Identifying the disincentive effect alone requires a design that neutralizes the demand channel. The authors accomplish this by comparing low-wage and high-wage businesses within the same narrow local market, industry, and price tier, so that common local demand shifts from UI are differenced out.

Q: What data does the empirical analysis use, and how is the sample constructed? A: The primary data source is Homebase, covering approximately 140,000 small U.S. businesses with daily employment, hourly wages, and hours worked. The estimation sample is restricted to 4,595 businesses present throughout 2019, matched to Yelp price-tier classification and Safegraph weekly foot traffic and credit-card spending. Businesses are grouped into 1,195 local-industry cells defined by four-digit zip code, two-digit NAICS industry, and Yelp price tier (inexpensive vs. expensive). Within each cell, businesses are classified as low-wage or high-wage, with high-wage businesses paying on average $1.80 per hour more — about 8% above the average hourly wage of $10.90.

Q: How is the replacement rate defined in the empirical framework? A: The business-specific replacement rate is the ratio of average UI receipts (state benefit plus the pandemic supplement, converted to hourly units) to the pre-pandemic average hourly wage of that business. Because the supplement is uniform across workers, businesses with lower pre-pandemic wages face higher replacement rates; the replacement rate gap between low- and high-wage businesses within a local market is therefore a function of both state benefit levels and the local wage dispersion.

Q: What does the event-study analysis around the $600 expiration show? A: The event study exploits cross-labor-market variation in how much the replacement rate gap between low- and high-wage businesses declined when the $600 FPUC supplement expired at end of July 2020. Labor markets with a larger decline in the gap see faster relative recovery in low-wage business employment after expiration. A 100 percentage point decline in the replacement rate gap is associated with a 5.7 percentage point rise in the low-versus-high-wage employment recovery gap at 12 weeks post-expiration. For the average labor market, the $600 expiration reduced the replacement rate gap by 46 percentage points, implying a 2.6 percentage point narrowing of the employment recovery gap.

Q: Why does the estimated effect disappear when broader geographic aggregations are used? A: When businesses are compared within U.S. state borders rather than within local-industry cells, the estimated coefficient on the replacement rate gap turns positive and statistically insignificant. This occurs because at the state level, low-wage areas benefit disproportionately from the purchasing power increase that generous UI provides to local unemployed workers, so demand effects swamp and reverse the supply-side disincentive. This finding explains why Finamor and Scott (2021), using Homebase data with state fixed effects, find no negative association between replacement rates and labor market re-entry.

Q: What evidence supports a labor supply rather than demand interpretation of the differential recovery? A: During the period of the $600 supplement, hours per employee and hourly wages grew faster in low-wage businesses than in high-wage businesses, even as low-wage businesses lagged in employment levels. If the differential recovery reflected demand deficiencies at low-wage businesses, hours per employee and wages should have grown faster at high-wage businesses instead. The observed pattern is consistent with labor supply shortfalls at low-wage firms.

Q: What is the structure of the quantitative labor search model? A: The model features a unit measure of workers and a fixed measure of firms, each posting a constant idiosyncratic wage drawn from an exogenous distribution. Unemployed workers receive job offers at a rate determined by labor market tightness and accept offers above their reservation wage. Reservation wages are equilibrium objects because UI benefits depend on the worker’s previous wage. The unemployed are split into UI-eligible and non-UI pools; the non-UI pool accepts jobs from lower in the wage distribution and is the primary supply source for low-wage firms. The model is calibrated to pre-pandemic U.S. service sector averages, with a pre-pandemic UI replacement rate of 0.51, a UI recipiency probability of 14%, and a non-UI replacement rate of 0.15.

Q: Why does the baseline model overstate the empirical effect, and how is this reconciled? A: The baseline model dramatically overstates the differential employment impact of the $600 expiration because the CARES Act’s expanded eligibility (modeled as a rise in the recipiency probability from 14% to 70%) nearly empties the non-UI unemployed pool, which is the dominant labor supply source for low-wage firms. In the data, the share of unemployed receiving UI nearly tripled for in-person leisure and hospitality workers, but not to the degree that the model’s implied employment collapse would require. The model is reconciled by introducing a 16.5% probability that a worker loses UI eligibility upon refusing a suitable job offer — consistent with UI law — which reduces the effective outside option and raises acceptance rates for low-wage firms.

Q: What are the aggregate employment losses implied by the model? A: When all three CARES Act provisions are implemented jointly, the model estimates that the disincentive effects held back aggregate employment recovery by 3.4 percentage points on average between April and December 2020 — approximately 20% of the average employment shortfall in the Leisure and Hospitality sector. Implemented in isolation, each provision generates only modest losses: the $600 supplement alone accounts for 0.2 percentage points, extended duration for 0.2 percentage points, and expanded eligibility for 1.0 percentage points. The large combined effect arises from the interaction of all three provisions, not from any single one.

Q: What are the conditional (interaction) effects of each provision when the other two are in place? A: Conditional on the other two provisions being active, the income supplement holds back employment recovery by 1.6 percentage points, the extended duration by 1.5 percentage points, and expanded eligibility by 2.9 percentage points. This interaction effect is the central quantitative finding: individually modest provisions combine to produce effects far exceeding their sum when implemented simultaneously.

Q: What are the implied unemployment duration elasticities, and how do they compare to the literature? A: The $600 supplement alone raises average unemployment duration by 8% against a 343% rise in the replacement rate, implying an elasticity of 0.02. Extended duration alone raises unemployment duration by 6% against a 150% increase in potential benefit duration, implying an elasticity of 0.03. Expanded eligibility alone raises unemployment duration by 19%, implying an elasticity of 0.04. When each provision is activated on top of the other two, the implied elasticities rise substantially: 0.24 for the $600 supplement, 0.43 for extended duration, and 0.28 for expanded eligibility. These are in the low-to-middle range of pre-pandemic estimates (Katz and Meyer, 1990: 0.3–0.5; Johnston and Mas, 2018: 0.4–0.8; Rothstein, 2011: 0.06; Farber and Valletta, 2015: 0.15).

Q: What is the role of expanded eligibility specifically? A: Expanded eligibility is identified as the most disruptive CARES Act provision, accounting for 1.0 percentage points of employment loss alone and 2.9 percentage points conditional on the other provisions. Mechanically, expanded eligibility converts non-UI unemployed workers into UI-eligible workers, draining the pool of workers willing to accept low-wage job offers. Because low-wage firms depend disproportionately on the non-UI pool for hiring, this provision disproportionately depresses their employment. Using CPS data, the authors document that the share of unemployed workers receiving UI in the in-person leisure and hospitality sector nearly tripled in 2020 relative to the pre-pandemic period.

Q: What are the scope conditions and limitations of the analysis? A: The empirical analysis is restricted to small, in-person service sector businesses (restaurants and retailers) in the Homebase sample, which may not be representative of the broader labor market. The quantitative model is explicitly focused on disincentive effects only and does not capture the stimulative or demand effects of UI. The model also abstracts from re-opening restrictions and other pandemic-specific confounders. The analysis covers April to December 2020; the 2021 pandemic UI extensions are not studied. The job-refusal probability (chi = 16.5%) is a reduced-form calibration target rather than a structurally identified parameter.

Replacement rate gap: The difference in business-specific UI replacement rates between low-wage and high-wage businesses within the same local labor market; defined as UI benefits (state benefit plus supplement) divided by the business’s pre-pandemic average hourly wage. Larger gaps indicate greater relative disincentive for workers to accept jobs at low-wage firms.

Disincentive effect: The negative impact of higher UI replacement rates on workers’ willingness to accept job offers and thus on business employment recovery, isolated from the simultaneous stimulative demand effect of UI spending.

Non-UI unemployed pool: Workers who are ineligible for or have exhausted UI benefits and therefore receive only social benefits at a lower replacement rate (calibrated at 0.15 in the model). This group has a lower reservation wage and constitutes the primary labor supply source for low-wage firms.

Local-industry cell: The paper’s unit of comparison — businesses sharing the same four-digit zip code (covering on average four neighboring zip codes), two-digit NAICS industry, and Yelp price tier. Within-cell differencing is the mechanism that removes common local demand effects.

Benefit recipiency probability: The probability that a newly separated worker enters the UI-eligible unemployed pool, combining UI eligibility and takeup. Pre-pandemic this is calibrated at 14%; under the CARES Act it rises to 70%, targeting the observed near-tripling of UI recipients in the CPS data.

Job-refusal eligibility loss: A probability (calibrated at 16.5%) that a UI-eligible worker who rejects a job offer loses UI status and transitions to the non-UI pool. Motivated by UI law prohibiting refusal of suitable work; reduces the effective outside option and reconciles the model’s predicted employment gap with the empirical estimate.

Equilibrium residual wage dispersion: The wage dispersion observed in equilibrium conditional on worker observables. The model generates realistic dispersion by calibrating the non-UI replacement rate to match the lower half of the wage distribution and the firm wage offer variance to match the upper half; the presence of the non-UI state substantially increases residual dispersion relative to standard search models.

Unemployment Insurance, Starting Salaries, and Jobs

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

Seven U.S. states permanently cut unemployment insurance (UI) benefits by 30–64 percent between 2011 and 2014, providing the study’s quasi-experimental variation. North Carolina enacted the largest reform: maximum duration fell from 26 weeks to 12–20 weeks and maximum weekly benefits fell from $535 to $350, an average total reduction of 64 percent. Six “moderate reform” states (FL, GA, KS, MI, MO, SC) cut duration only, by an average of 30 percent (26→18 weeks). Using a multi-state firm identification strategy — comparing establishments of the same firm operating in reform states against the same firm’s establishments in non-reform states, with establishment and firm×year fixed effects — the paper estimates causal effects of UI cuts on employment (EEOC, 946K–1.4M establishment-years), starting salaries (Glassdoor, 500K–942K person-years), and posted wages (Burning Glass Technologies, 709K–1.18M establishment-job-quarters). The main results: NC establishments gain +1.3% employment on average relative to same-firm establishments in other states (ATT), reaching +2.1% by year 2; moderate reform states gain +0.8% (ATT). Starting salaries of new hires fall −5.5% in NC and −1.2% in moderate states. Posted wages for the same job within the same firm fall −3.5% in NC and −3.2% in moderate states. The negative co-movement of employment and wages identifies a labor supply shock: workers lower reservation wages in response to reduced outside options; firms take advantage by hiring more at lower wages. Labor demand elasticity: −0.36 (SE 0.21) for NC, −0.42 (SE 0.18) for moderate states. The larger effects in NC relative to moderate reform states are consistent with the larger total benefit reduction; effects are robust to controlling for concurrent right-to-work laws, minimum wage changes, Medicaid expansions, and corporate/personal tax reforms. The paper concludes that large, permanent UI reductions can raise employment but at the cost of lower starting wages.

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

In depth

Q1. How does the multi-state firm design separate UI effects from aggregate and local shocks?

The key innovation is including firm×year fixed effects alongside establishment fixed effects: within any given firm and year, the only remaining variation is which state the establishment is in — absorbing all firm-wide demand trends, management strategies, and capital-allocation decisions that would otherwise confound cross-state comparisons. Standard difference-in-differences compares reform states to non-reform states at the level of geographic unit or industry; this approach confounds UI changes with the economic conditions that prompted them. The multi-state firm design eliminates this confound because firms’ nationwide operational decisions are held constant. The identification concern is policy endogeneity — whether reform states had weaker economies motivating both the UI cuts and slow hiring. This is addressed in three ways: (1) the 27 other states whose UI trust funds became insolvent in the early 2010s did NOT cut benefits, ruling out insolvency per se as the trigger; (2) restricting the control group to only the insolvent states (Table 3 cols 2 and 5) leaves estimates nearly unchanged; (3) restricting further to insolvent states that experienced a Great Recession unemployment shock within ±2 percentage points of the reform states (Table 3 cols 3 and 6) again leaves estimates unchanged, ruling out mean reversion. The mean reversion hypothesis is additionally ruled out by the wage results: mean reversion predicts faster wage growth in reform states, but wages fall.

Q2. What are the quantitative employment effects, and how do they compare across specifications?

In the baseline specification (Table 3, column 1), NC establishments gain +1.26% employment on average over the post-reform period (ATT, SE 0.0052, p<0.05), with the effect growing from +1.2% in year 1 to +2.1% in year 2 (both p<0.01); moderate reform states gain +0.83% (ATT, SE 0.0022, p<0.01), reaching +1.5% by year 2. Alternative specifications (Table 5) using less-saturated fixed effects (firm+state+year FEs or establishment+year FEs only) produce estimates roughly twice as large — +2.5% for NC — confirming that firm×year fixed effects absorb a substantial share of cross-state employment variation that is not attributable to UI. This amplification underscores why controlling for firm-level trends matters: firms simultaneously expanding in many states would appear in the unconditioned data as UI-reform effects. Robustness to policy confounders (Table 4): excluding states with RTW law changes, minimum wage changes, Medicaid expansions, major corporate or personal tax reforms all leave ATTs statistically significant and economically similar (0.80%–1.25% for NC; 0.82%–1.18% for moderate states). A Fisher exact test places NC’s t-statistic in the top 2/42 (4.8%) of placebo assignments, consistent with a one-sided 5% test. Controlling for NC’s concurrent corporate tax cut, which bounds the maximum tax-driven employment effect at 0.76pp (Giroud and Rauh 2019), implies the UI reform accounts for between 0.5% and 1.26% of NC’s employment increase — broadly consistent with the 0.83% moderate reform estimate.

Q3. What do the wage results show, and how do posted wages rule out compositional explanations?

Table 7 (Glassdoor starting salaries, job and firm×year FEs): NC ATT = −5.5% (SE 0.021, p<0.01), moderate states ATT = −1.2% (SE 0.0051, p<0.05); the effect is concentrated in jobs with starting salaries at or below $100,000, where UI replacement rates are meaningfully binding, and is statistically insignificant for higher-wage jobs. Starting salary declines could in principle reflect worker composition (lower-skilled workers drawn into the labor force) or worse job matches (workers accepting jobs below their productivity) rather than firms lowering offer wages. Burning Glass Technologies (BGT) posted wages, which measure the wage advertised for the same job within the same firm over time (establishment-job and firm×year FEs), rule out both channels: Table 8 shows NC posted wage ATT = −3.5% (SE 0.013, p<0.01) and moderate states = −3.2% (SE 0.0071, p<0.01). The near-equality of posted and realized wage effects implies the wage decline is driven by firms lowering their wage offers — not by changes in worker composition or match quality. Occupational heterogeneity confirms the mechanism: high-exposure occupations (above-median fraction of workers with unemployment spells or employment tenures exceeding 20 weeks) exhibit NC posted wages −3.5% and moderate states −4.1%; low-exposure occupations show near-zero insignificant effects (Table 9). Posted wages also provide additional evidence against mean reversion: if reform states had faster-growing underlying wages, posted wages would rise relative to controls, but the opposite occurs.

Q4. How does the negative co-movement of employment and wages identify a labor supply shock and discipline the theoretical mechanism?

The simultaneous rise in employment (+1.26% NC, +0.83% moderate) and fall in posted wages (−3.5% NC, −3.2% moderate) is the signature of a labor supply shock under the standard Mortensen-Pissarides (1994) framework: when workers’ outside option (the value of UI) falls, their reservation wages fall, inducing firms to post more jobs at lower wages. A positive demand shock would raise both employment and wages; a positive supply shock raises employment while lowering wages. The posted wage channel further implies that firms’ labor demand responds to the wage reduction (not just to the supply expansion): if firms were passive price takers, posted wages would not change. The data imply that firms internalize workers’ changed outside options and lower their wage offers accordingly, consistent with the monopsonistic wage-setting in Mortensen-Pissarides with free entry. The labor demand elasticity calculated as (Δlog employment / Δlog posted wage) = 1.26/3.5 ≈ −0.36 (SE 0.21) for NC and 0.83/3.2 ≈ −0.26 or using preferred specification −0.42 (SE 0.18) for moderate states; these fall in the middle of the distribution of prior estimates from cross-country labor demand elasticity studies (Hamermesh 1996; Acemoglu et al. 2004). A Chodorow-Reich et al. (2019) decomposition suggests that if labor market tightness increased (fewer unemployed and more vacancies), the reservation wage (opportunity cost) effect dominates the tightness effect — since we observe posted wages falling.

Q5. What do the CPS results add, and how do employment duration effects inform the mechanism?

Using individual-level CPS data with state and year fixed effects (no within-firm comparison), combining all reform states: employment probability +1.0pp (SE 0.43pp, a 1.5% increase relative to the 65% baseline) [Table 10 col 1]; new-hire wages (tenure <1yr) −6.3% [col 2]; unemployment duration −2.8 weeks/year ATT (relative to 33.48-week control mean, an 8% reduction) [col 3]. The CPS results are qualitatively consistent with the multi-state firm findings and use an entirely different data source, sampling frame, and identification approach. The unemployment duration effects are instructive about timing and mechanism: the ATT is negligible in the first two post-reform years (−1.0 and −1.2 weeks, insignificant), rises to −1.7 weeks in year 3, −3.6 in year 4, −4.2 in year 5, and −5.6 in year 6 — consistent with gradual stock-flow dynamics (the stock of workers who began unemployment before the reform exhausts gradually, so average duration in the reform states drifts lower over time as a larger share of the unemployed pool faces the new rules). This pattern helps interpret the gradual employment growth in the event studies.

Q6. How does the paper explain divergence from prior literature finding small UI effects?

The paper argues that prior work finds small effects because reforms studied were smaller in size, temporary, and enacted during deep recessions — all conditions where the job creation channel from lower reservation wages is muted. Schmieder et al. (2010), Rothstein (2011), Farber-Valletta (2015), Chodorow-Reich et al. (2019) and others study UI extensions/expirations that are often 13–20% changes in duration (versus NC’s 44% duration cut and 64% total benefit cut), enacted during high unemployment (when moral hazard is lower) or temporary (so workers discount the change in outside options). A 13-week contrast off a high base of 83 weeks (the EUC expansions) differs fundamentally in moral hazard intensity from an 11.5-week cut off a low base of 26 weeks plus a benefit level reduction — the effective present value of UI falls far more in the NC reform. Additionally, the border county-pair design used in much prior work (Chodorow-Reich et al. 2019, Hagedorn et al. 2025) compares establishments on opposite sides of a state border within the same labor market; these competing establishments cannot fully exploit lower reservation wages because they compete for the same workers — suppressing both the employment and wage responses. Notable exceptions that do find sizable effects — Johnston-Mas (2018) and Karahan et al. (2025), both studying large permanent post-recession reforms — corroborate this paper’s findings.

Key concepts

multi-state firm design : the identification strategy that compares establishments of the same firm operating in reform states against the same firm’s establishments in non-reform states; with establishment and firm×year fixed effects, this absorbs firm-wide demand trends, product market shocks, and management decisions that affect all of a firm’s establishments equally, isolating state-level UI variation as the sole source of identification.

reservation wage : the minimum wage at which an unemployed worker is willing to accept a job offer, determined by the outside option value (UI benefits plus expected future wages from continued search); UI cuts reduce the outside option value, lowering the reservation wage and enabling firms to post and fill vacancies at lower wages.

posted wage : the wage listed in a job advertisement before any worker-firm negotiation or match quality sorting; measured here using Burning Glass Technologies (BGT) data at the establishment-job level, controlling for the same job across time within the same firm; distinct from realized starting salary in that it reflects the firm’s wage-setting decision independent of which worker accepts.

labor supply shock : an exogenous change in the willingness of workers to supply labor at given wages; identified here by the negative co-movement of employment (up 1.3–0.8%) and wages (down 3.5–3.2%), which is the opposite of what a positive labor demand shock would predict, ruling out confounding from corporate tax cuts or mean-reverting demand.

outside option : the payoff available to an unemployed worker from continued search rather than immediate job acceptance; UI benefits are the dominant component; when UI generosity falls, the outside option value falls and firms can hire more workers at lower wages — the core mechanism linking permanent UI cuts to simultaneous employment gains and wage reductions.