Office of Economic Policy

Working Paper 2020-01

December 2020

The Job-Preservation Effects of Paycheck

Protection Program Loans

Michael Faulkender, Robert Jackman, and Stephen I. Miran

This Office of Economic Policy Working Paper presents original research by the staff of the Office of Economic

Policy. It is intended to generate discussion and critical comment while informing and improving the quality

of the analysis conducted by the Office. The paper is a work in progress and subject to revision. Views and

opinions expressed are those of the authors and do not necessarily represent official Treasury positions or

policy. Comments are welcome, as are suggestions for improvements, and should be directed to the authors.

This Working Paper may be quoted without additional permission.
The Job-Preservation Effects of Paycheck Protection

Program Loans∗

Michael Faulkender†

Robert Jackman‡

Stephen I. Miran§

December 16, 2020

Abstract

The Paycheck Protection Program (PPP) covered in excess of 80% of eligible U.S. small business

employment, supporting 51 million American jobs through the program’s close on August 8th,

2020. Of those supported jobs, how many would have been lost in the absence of PPP loans?

To answer this question, we propose an empirical strategy to identify the effects of PPP loans

on county-level unemployment insurance claims. Specifically, we exploit variation in the timing

of loan receipt caused by differences in local banking market structure across US counties. On

the margin, we estimate that a 10 percentage point increase in eligible payroll covered by PPP

resulted in a 1 to 2 percentage point smaller jump in weekly initial unemployment insurance

(UI) claims, as a share of employment covered by UI. With a lag, that same 10 percentage

point increase in PPP coverage resulted in an estimated 5 percentage point smaller increase

in the insured unemployment rate. In order to compare our estimates with related studies,

we calculate an aggregate employment effect of PPP loans. Moving from the 25th percentile

to 75th percentile of counties by early PPP coverage causes an improvement in the insured

unemployment rate of over 12 percentage points, or, extrapolated nationally, 18.6 million jobs.

We note meaningful caveats to interpreting this paper’s aggregate number; the same caveats that

apply to other papers evaluating PPP loans. This paper’s estimates are an order of magnitude

larger than previous evaluations of PPP, which have tended to find small employment effects or

none at all.

∗We are especially grateful to Steven Johnson and Matthijs Schendstok for many helpful comments and assistance

with analysis. We are also grateful to Tanner Black, Jonathan Greenstein, Mitchell Petersen, Jim Poterba, and Josh

Rauh for helpful feedback, and to Leah Damelin for research assistance. We thank the Bureau of Labor Statistics

for generating customized statistics and providing expert advice in support of this work. Special thanks are given to

Economist Jeremy Oreper and Associate Commissioner Julie Hatch Maxfield, as well as their partners at the State

Labor Market Information offices. We disclose that Michael Faulkender and Stephen Miran contributed to Treasury’s

efforts to support the Small Business Administration’s implementation of PPP. The findings and views are those of

the authors and do not necessarily reflect the position of the United States Department of the Treasury.

†Assistant Secretary, Economic Policy, United States Treasury.

‡Economist, Economic Policy, United States Treasury.

§Senior Advisor, Economic Policy, United States Treasury & corresponding author: stephen.miran@treasury.gov.

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I. Introduction

When most American small businesses faced mandatory Covid-19 related closures and a drastic reduction in

revenue, the Paycheck Protection Program (PPP) offered a financial lifeline. With the economy contracting

at an annualized 31.4% rate in the second quarter of 2020, widespread and permanent small business closure

and mass layoffs seemed likely. PPP was implemented to temporarily sustain America’s small businesses

and the jobs they provide. PPP coverage was broad: over 80% of eligible small business employment was

supported, totaling 51 million jobs when the second tranche closed. However, calculating the number of jobs

supported by the PPP is relatively simple; calculating the number of jobs preserved poses a harder task.

This paper addresses that thornier question: how many jobs would have been lost without this support, and

were therefore saved by the Paycheck Protection Program?

As a thought exercise and to provide a ballpark for where one might expect estimates to land, we produce

back-of-the-envelope calculations, which suggest that 13.4 to 17.3 million jobs may have been saved by PPP.

While these back-of-the-envelope calculations are not rigorous empirical work and do not merit being taken

at face value, they do serve to provide a credibility check on empirical studies conducted by others as well

as on our own work. The larger back-of-the-envelope calculation is quite similar to our own estimate.

We propose an empirical strategy to identify the employment effects of PPP loans using the fact that

different firms received loans at different times, based on which type of lender they used. However, finances

and employment are not independent, so we cannot use differences in the timing of loans directly. To address

this problem, we exploit geographic variation in the market structure of local banking. These differences

in banking market structure are useful, since community banks were quicker to approve and disburse first

round PPP funds than national banks or non-bank lenders. Under the condition that the composition of

the local banking market has no effect on early pandemic changes in local employment save through PPP

loans, we can use this variation to identify the effects of PPP loans on employment. We hypothesize that

areas which received PPP money early should have better labor market outcomes at the pandemic’s outset,

but that as PPP money reached near-full saturation, labor market outcomes should converge. This dynamic

convergence is the crux of our hypothesis, and an important test of our identification strategy. Since our

hypothesis is dynamic, any potential bias in the instrument must also be dynamic, relevant for one week but

not another.

Estimates based on county-level data suggest that, at the margin, a 10 percentage point increase in PPP

coverage of eligible payroll resulted in a smaller jump in initial Unemployment Insurance (UI) claims by

between 1 and 2 percentage points of covered employment.1 With a lag in time, the same increase in PPP

coverage led to a smaller increase in continuing UI claims (the IUR) by 5 percentage points. Consistent with

our expectations, the comparative benefits of early loan receipt disappear over time as lagging states catch

up on PPP penetration.

In order to provide a reference point for aggregate employment, this implies that moving from the 25th

percentile to 75th percentile of counties by PPP coverage causes an improvement in the unemployment rate

of over 12 percentage points, or, extrapolated nationally, 18.6 million jobs. This calculation assumes that

the estimated marginal effects equal the average effect within the interquartile range. While this strong

1Technically, we are referring to an initial claims analog to the Insured Unemployment Rate (IUR). In other words,

the underlying percentage is (number of initial UI claims)/(Covered Employment).

2
assumption is unlikely to hold literally, we make it in order to compare our estimates with other studies on

the topic, which require the same assumption to compute aggregate employment effects.

II. Previous Estimates

Three significant studies, Chetty et al. (2020), Autor et al. (2020), and Hubbard and Strain (2020) use

innovative high frequency, large datasets to study the effects of PPP. In most industries, only firms with

fewer than 500 employees were eligible to receive PPP money, and these studies leverage this eligibility cutoff,

comparing employment at firms just above and below that threshold. Autor et al. (2020) estimate that PPP

preserved 1.4 to 3.2 million jobs, while Chetty et al. (2020) find “no evidence” of a significant impact of the

Program, and Hubbard and Strain (2020) find statistically significant evidence of small employment effects.

While the empirical design of Autor et al. (2020) and Hubbard and Strain (2020) may be relevant in the

neighborhood of the eligibility cutoff, we question the external validity of these kinds of estimates. Using

reports from PPP loans themselves, we display in Figure I the cumulative density of PPP loan recipients

based on firm size, as measured by self-reported employment on loan applications. Where not reported

(6.8% of all loans), we leave firm size at 0, to keep with the Small Business Administration (SBA) standard.

However, imputation would not make a noticeable difference to our numbers. Figure I shows that 57.6% of

loans went to firms with three or fewer employees, 78.8% of loans went to firms with nine or fewer employees,

and 95% of loans went to firms with 38 or fewer employees.2 Table 1 details the fraction of PPP loans that

went to mid-sized firms and establishments, by number of loans, number of workers covered, and total PPP

dollars. The cutoffs were chosen to facilitate comparison with the findings of Autor et al. (2020). For that

paper’s largest treatment group, it captures 13% of all workers covered by PPP loans, representing only

11.1% of the total dollars lent and a mere 0.34% of total loans.

In order to credibly evaluate the aggregate employment effect of the PPP, we need an empirical strategy to

identify the effect on 100% of firms and workers. Estimates based on much larger firms, which constitute a

small share of PPP recipients, apply to smaller firms only if we believe large and small firms were equally

vulnerable to the pandemic’s economic shock. If, however, the assumption of equal vulnerability is violated,

these studies only tell us what happened in the neighborhood of the eligibility cutoff, for firms roughly 500

employees in size. Finding modest effects in large firms does not entail the same modest impact at firms

with 5 or 10 employees. Since small firms are more financially fragile, these firms stood to gain the most

from a PPP loan.

2

If we instead exclude loans with an unrecorded number of jobs, 54.7% of loans would be to firms with 3 or fewer

employees, 77.4% to firms with fewer than 10, and the 95th percentile of firm size would be 40 employees.

3
TABLE I

PPP Loans to Mid-Sized Firms, by Three Measures

Firm Size No. of Loans Total Workers Total Dollars

> 450 5,160 (0.10%) 2.5 million (5.0%) $21.7 billion (4.1%)

> 400 7,103 (0.14%) 3.36 million. (6.6%) $29.0 billion (5.5%)

> 350 9,551 (0.18%) 4.28 million (8.4%) $37.0 billion (7.1%)

> 250 17,508 (0.34%) 6.63 million (13.0%) $58.1 billion (11.1%)

Source: SBA and author’s calculations. Data as of 8 August, 2020. Firm Size is as reported by the

SBA, without any additional imputation. “Mid-sized firms” as measured by employee head-count.

13% of Program dollars went to firms with more than 250 employees

Figure I

Cumulative density of firm sizes by number of employees. Inferences drawn from firms with hundreds of

employees tell us little about the bulk of loan recipients. These lines are on top of each other, which tells us

our estimation for firms with missing worker counts matters little.

Moreover, Autor et al. (2020) note that neither of these studies observes firm-level loan data, which clouds

interpretation. Because the studies mingle eligible firms which received PPP loans with eligible firms which

did not receive PPP loans, the estimates themselves are biased downward; each estimate is a mixture of

no effect for eligible firms near the cut-off which did not receive a loan, and—potentially small—effects for

eligible firms near the cut-off which did receive a loan.

More problematic, as these studies move further away from the eligibility cutoff in order to increase sample

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size, their identification strategy grows murkier. The larger the difference in employee size, the weaker the

comparison. Larger firms tend to have more sophisticated managers, larger cash buffers, more ready access

to established lines of credit, fixed contracts with customers, and more bargaining power with suppliers. This

puts larger firms in a stronger position to weather economic shocks than small business. Because of their

superior ability to access lending and financial markets, larger firms were able to benefit from other treatments

not available to smaller firms, like the Federal Reserve’s interventions. Distance from the eligibility cutoff is

particularly a problem in Chetty et al. (2020), who use firms with hundreds or thousands of employees as

counterfactuals for firms with 50 employees; it is unclear what to make of such a comparison.

If businesses with fewer than ten employees had similar Covid-era employment outcomes as businesses with

more than 250 employees (as suggested by Chetty et al. (2020)), this speaks to the success of PPP, not its

failure. Without PPP’s safety net to buffer against this unprecedented aggregate shock, economic theory

and historical experience strongly suggest widespread small business permanent shutdowns and layoffs would

have occurred; larger businesses are better able to survive shocks. We review evidence for these differences

in fragility between small and mid-sized firms below.

A fourth study, conducted by Granja et al. (2020), finds precisely estimated tiny effects of PPP on hours

worked, business shutdowns and UI claims. With respect to hours worked and business shutdowns, we note

that the goal of PPP was to facilitate employees being paid while not physically showing up to workplaces,

essentially working zero hours, and for businesses to shut down without laying off workers, closing permanently or filing for bankruptcy. Finding widespread business shutdowns and reductions in hours worked

without corresponding bankruptcies doesn’t imply failure of the Program. Indeed if those employees went

back to work as the economy reopens, such outcomes will speak to the Program having worked as intended.

In normal recessions, hours worked and business shutdowns are a good indicator of severity. However, a

public-health recession is unusual in many ways. Policy was not traditionally Keynesian, and instead it was

explicitly aimed at reducing business activity and face-to-face interaction by providing household liquidity

that facilitated compliance with stay-at-home orders.

Further, Granja et al. (2020) study employment outcomes as a function of ultimate PPP penetration. However, since final PPP penetration is very high everywhere—in excess of 80% of eligible employment is covered,

nationally—there might not be enough variation in the underlying independent variable of interest to identify

the key effects. Granja et al. (2020) instead use an alternative independent variable, which is a function of

bank market shares like ours below, and which likely induces more variation than ultimate PPP penetration

itself. Our approach is distinct, precisely because it exploits the dynamic variation in PPP receipt.

Using survey data, Bartik et al. (2020c) estimates a larger employment effect for the PPP. However, their core

employment estimates are not statistically significant. Bartik et al. (2020c) uses pre-existing relationships

with banks to instrument for loan receipt, marking a similarity with this paper’s use of banking market

structure. The two banking-relationship instruments yield point estimates of 3.31 and 4.99 jobs saved per

loan, though the associated standard errors are also large. If we multiply the point estimates by the roughly

five million loans, this would imply between 16.6 and 25 million jobs preserved by the PPP, consistent with

our main results below considering the size of the standard errors.

Finally, general equilibrium effects exist, and studies drawn from microdata do not capture them. If layoffs

at treated businesses reduce aggregate demand and cause layoffs at other businesses, then outcomes in any

5
control group are caused in part by outcomes at the treatment group. This would lead to a classic violation

of the Stable Unit Treatment Value Assumption (SUTVA) in a Difference-in-Differences design, and the

observed differences between treatment and control groups will understate the true effects. If the study

includes time fixed effects to measure overall employment outcomes at a moment in time, the time fixed

effects will capture many of these general equilibrium effects. In that case, the true variable of interest will be

some combination of the estimated treatment effect and the time fixed effect. However, since the time fixed

effects will also capture the effects of other policies and shocks—for instance, Federal Reserve interventions,

transfers to households, medical developments—identification of the full treatment effect will be difficult.

Since PPP ultimately covered over 80% of national eligible employment, these general equilibrium effects will

be nontrivial, and estimated treatment effects likely understate the true effects of PPP, though there may

be biases in the other direction as well. Since our approach below uses counties as observational units with

state-week fixed effects, this problem is likely to be less pronounced, though we will still miss cross-county

economic linkages.

III. Back-of-the-Envelope Calculations

While most academic estimates of PPP’s effects are very low, perceptions in markets and industry are

much more optimistic. Standard and Poors (Fox et al. (2020)) estimates that PPP saved 13.6 million jobs,

while JP Morgan’s Jamie Dimon believes that PPP saved 35 million jobs (Ruhle, Miranda, and Capetta

(2020)). In March, there were near-universal market expectations for a world of widespread, permanent

small business closures and the mass layoffs that would result. For example, in February and March, many

stock market indices declined by over 40%, including the Russell Microcap Index. The subsequent recovery

in broad equity prices, including microcap stocks—which are only about 6% off their February high as of

August 19th, 2020—is prima facie incompatible with miniscule effects of PPP loans and other government

interventions. There is a wide disconnect between academic and market opinions.

At the outset of the pandemic, small businesses suddenly found themselves facing significant revenue declines

that would likely cause insolvency. Smaller businesses are less likely to have access to an established credit

line; and indeed if all small businesses sought credit simultaneously, the ability of the banking system to

respond would likely increase the price and decrease the availability of credit.3 A study by Farrell and Wheat

(2016) found that the median small business had cash buffers to last only 27 days without revenue, and only

25% of small businesses could last more than 62 days without income. Their sample comprised roughly

600,000 small businesses; of these, 70% had five or fewer employees, which closely matches the universe of

PPP loans wherein 70% of recipients had seven or fewer employees.

Against this backdrop, economic theory and experience from past recessions would indicate a surge in small

business bankruptcies given the magnitude of the economic shock, as has occurred in previous recessions.

Nevertheless, increases in small business bankruptcies have been modest, despite the largest economic shock

in nearly a century, according to data from the Justice Department and analysis by the Council of Economic

Advisers (2020). Indeed, contrary to expectations that small business bankruptcies should surge in this

3Although the Federal Reserve eased conditions in financial markets, it did not provide regulatory relief that would

have explicitly eased the ability or willingness of banks to directly make loans to the number of small businesses under

stress.

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economic environment, Figure II shows that after spiking in February and March due to regulatory changes

around Chapter 11 filings,4increases in small business bankruptcies in April through June were lower than

they were before pandemic struck, and that small business bankruptcies increased by less than larger-sized

business bankruptcies. Moreover, we observe that Chapter 11 bankruptcies picked up in July and remained

elevated, consistent with the timing of when PPP funds began depleting. Figure III shows that even with

the spike in the summer, overall Chapter 11 bankruptcy filings are well within historical norms and far from

levels seen in the last recession. Figure IV shows that Chapter 7 bankruptcies for the entirety of FY 2020 are

below any level since 2008; unfortunately, the Department of Justice does not provide CEA with separate

Chapter 7 filings data by business size, or data at frequencies higher than fiscal years. While many small

businesses may simply shut down instead of filing for Chapter 7 or Chapter 11, one might expect a jump in

simple shut-downs to be correlated with jumps in filings, and we haven’t yet observed any historic jump in

filings to correspond to the historic economic shock.

Indeed, industry economists have expected economy-wide bankruptcies throughout the recession. According

to David Mericle, head of US economics at Goldman Sachs, “This really seemed like it had the potential

to be a huge collapse. For most people, and I would include myself, [the lack of bankruptcies] has come

as a pleasant surprise.” For Mericle, the Paycheck Protection Program was the top item explaining the

lower-than-expected number of bankruptcy filings (Coy (2020)).

Figure II

US Total and Small Business Chapter 11 Bankruptcy Filings. Source: Council of Economic Advisers

(2020).

4For a discussion of the rules changes, see Ekvall and Evanston (2020). Small businesses took advantage of the

easing criteria for Chapter 11 reorganizations, in the weeks before the pandemic hit. The Small Business Reorganization Act was signed in August 2019, and thus forward-looking firms had plenty of time to plan for the implementation

in February.

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Figure III

Small business Chapter 11 bankruptcy filings, March 2006 - Oct 2020. Source: Council of Economic

Advisers (2020).

Figure IV

Total Chapter 7 bankruptcy filings, FY 2008 - FY 2020. Source: Council of Economic Advisers (2020).

We propose two simple back-of-the-envelope calculations to guide thinking about job preservation through

the PPP. The first takes the threat of large-scale, permanent small business closure seriously, and estimates

the job losses that might arise from those permanent closures. The second uses the small business survey

responses from Bartik et al. (2020b) and estimates potential job losses based on those data. While these

back-of-the-envelope calculations are not rigorous empirical work, they do serve to provide a ballpark of

where one might expect empirical results to land, and thus a credibility check on empirical studies including

the previous work we discussed above and our own empirical work below.

First, we consider the consequences of permanent, large-scale small business closure. Although we cannot

directly observe the cash flows of small firms, we draw on findings in Farrell and Wheat (2016) that fewer

8
than 75% of small firms hold cash buffers to cover more than 62 days of expenses. The PPP has given these

financially vulnerable small businesses a vital cash flow lifeline to fill in for the drastic reduction in revenues

caused by the pandemic and economic shutdown, thereby facilitating their survival and shoring up millions

of jobs. The fragility of small businesses found in Farrell and Wheat (2016) is corroborated in work by Bartik

et al. (2020a), which finds in an independent survey that fewer than 30% of firms had cash on hand to cover

more than two months’ expenses. Additionally, the Census Bureau’s Small Business Pulse Survey found in

the week ending May 2 that only 16.7% of small businesses had enough cash on hand to cover three or more

months of business operations.

We start the first back-of-the-envelope calculation by making the assumption that 75% of small businesses

covered by PPP would permanently shut down and layoff their workers in the Program’s absence. This

assumption is strong, but the evidence cited on small business fragility provides some justification. While it

is possible small businesses could secure liquidity from private sources to help them manage shocks, evidence

from the Joint Small Business Credit Survey in Federal Reserve Banks of New York, Atlanta, Cleveland,

and Philadelphia (2014) suggests otherwise. According to the Fed Banks’ report, only 32% of firms with 1-9

employees (which correspond to 75% of PPP recipients) received any credit, and a majority of firms with

less than $1 million in revenue were unable to secure any credit whatsoever. Moreover, 40% of firms seeking

credit said the primary purpose was for expansion, which suggests that fewer than 13% of firms with 1-9

employees had a line of credit which could be used to buffet revenue shocks.

The Fed report further finds that the primary means of financing of firms with less than $250,000 in revenues

is personal savings; that the average time it takes a small business to fill out a credit application is 24 hours;

and that typical wait times for approval are on the order of months. And credit has since the report become

harder to get: according to the Federal Reserve’s Senior Loan Officer Survey, the net percentage of banks

tightening lending conditions for commercial and industrial loans to small firms reached 70% in the third

quarter, only a few percentage points away from the previous peak in the series in the fourth quarter of 2008.

All this suggests that most small businesses rely on their cash buffers to smooth expenditures.

In recognizing that some firms will have access to alternate lines of credit, and that revenues did not fall to

zero for many businesses, we make a more conservative assumption regarding which firms go out of business.

We assume that only the smallest firms who received PPP close: as argued, the smallest firms are the most

likely to close. Further, insofar as this particular assumption misses the mark, it leads us to understate the

true number of jobs preserved. Therefore, this back-of-the-envelope calculation assumes that the smallest

75% of PPP recipients would have had to lay off their workers without PPP, because their cash buffer stocks

wouldn’t allow them to meet expenses for longer than 62 days. Implementation of this assumption suggests

that 13.4 million workers had their jobs preserved due to PPP.

To facilitate comparison with the estimates in Autor et al. (2020), consider that if all jobs lost were as a result

of shutdowns at the smallest businesses, the loss of 1.4 to 3.2 million jobs is approximately consistent with

the employment of the smallest 1.1 to 1.9 million firms in the sample of PPP loans. In contrast, the smallest

75% of firms we use for our back-of-the-envelope calculation corresponds to approximately 3.5 million small

business closures.

An alternative back-of-the-envelope calculation can be derived from surveys in Bartik et al. (2020b). These

surveys indicated that small firms in early April expected to have employment levels relative to January to

9
fall 40% by year-end. However, when told about the forgiveness provisions in the CARES Act loans, the

survey results reduced that forecast to a 6% reduction (firms told about loans but not forgiveness expected

reductions of 14%). Extrapolated over the 51 million small business employees supported by PPP, this

implicit reduction in unemployment is equal to 17.3 million workers.

One final point of reference is results in Barlett and Morse (2020), who find in a survey of 278 small businesses

in Oakland, CA, that PPP receipt reduced the subjective risk of medium-term small business closure by an

average 20.5 percentage points; if 20.5% of firms with fewer than 500 employees were forced to shut down

due to the recession, that would destroy approximately 12.4 million jobs. Because the sample is small and

localized, the results from this survey may generalize less readily than those of our other calculations above.

While these back-of-the-envelope calculations are useful for providing context, they are not careful empirical

work. We now turn to an empirical strategy to identify and estimate the employment effects of PPP.

IV. Data

In our empirical strategy discussed below, our primary variation is PPP Loan data courtesy of the Small

Business Administration. We observe the universe of approved SBA loans through August 8th, when the

Program closed to new applicants. These data include loan recipient, address, exact loan amount and date,

as well as a self-reported number of jobs covered. Further, we observe the lender for each loan, which we

match up to FDIC data in order to identify community banks.

We exclude loan applications that were canceled or never disbursed. The coverage of self-reported jobs is

incomplete, and we impute missing observations based on the size of the loans as described above. While these

data include limited demographic information of the applicant—race, sex, and veteran status—applicants

left these fields unanswered much more often than not, and we make no attempt to analyze the data using

those variables.

County-level Unemployment Insurance (UI) claims weekly data are furnished by the Bureau of Labor Statistics. Initial claims data reflect the number of initial claims filed and eventually accepted by the state. Note

that this is a different measure of initial UI claims than the widely-used weekly state-level UI release, which

reports the number of claims filed, regardless of whether or not they are ultimately accepted.5 Continuing

UI claims data are also furnished by the BLS. Unlike the data on initial UI claims, we observe these data

monthly, not weekly. We impute the data for the interim weeks using both accepted initial claims and the

attrition rate implied by the monthly difference in continuing claims. These data reliably cover 40 states and

DC. The remaining 10 states either do not report county-level statistics to the BLS, or their county-level

data suffer from inconsistent reporting.6 A state-level analysis in Appendix B, which draws on all 50 states

and DC, corroborates our county-level findings, albeit it cannot make use of cross-sectional fixed effects.

We construct data for community bank penetration from the Federal Deposit Insurance Company’s Summary

5Given that our data include only claims which are accepted, they avoid some of the double-counting issues

involved in applications for ordinary state programs and Pandemic Unemployment Assistance funds, discussed in

Cajner et al. (2020) and elsewhere. Our metric of initial UI claims is arguably cleaner than the state-level releases

made available to the public.

6The omitted states are California, Florida, Hawaii, Iowa, Kansas, Michigan, Minnesota, North Carolina, Oregon,

and South Dakota.

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of Deposits data, and we defined community banks following the FDIC’s Institution Directory. Data on

county population size and land area are from the US Census Bureau. Additionally, measures of eligible

county-level payroll for firms with fewer than 500 employees are estimated based on Census’ 2017 Statistics

of U.S. Business. The reported payrolls have been adjusted for three years of estimated growth. These

estimates are further adjusted to account for the fact that larger firms in the Accommodation and Food

Services industry were also permitted loans. With the exception of Nevada, this adjustment was relatively

small. County-level data on Covid-19 cases and deaths are from the New York Times, based on reports from

state and local health agencies.

Summary statistics for all our key variables can be found in Appendix A.

V. Identification from variation in timing of loan receipt

To estimate the employment effects of PPP loans, we propose to exploit variation in the timing of PPP

receipt. This contrasts with a strategy that examines firms near the eligibility cutoff (Chetty et al. (2020),

Autor et al. (2020) or ultimate PPP receipt Granja et al. (2020)). Since PPP penetration was ultimately very

high throughout the country, firms which received money earlier should have better employment outcomes

earlier, with other firms catching up as PPP approached saturation everywhere.7

However, the timing of loan receipt cannot, by itself, identify the employment effect of these loans; potential

confounders abound. Perhaps better management led to quicker loans, or loans flowed to firms which were

more (or less) effected by Covid-19 and its associated demand effects. It could be that banks had more

confidence in larger clients, or clients with a long-term banking relationship, and hence favored these firms

to receive first tranche loans. To address these endogeneity concerns, we need to find an additional source

of variation before we can leverage the variation in timing of loan receipt.

We propose community bank shares as this additional source of variation. There is evidence that some banks

were very fast in disbursing PPP funds, while other banks were slower. Liu and Volker (2020) documented

that community banks were among the fastest; this is due in part to both differences in pre-existing banking

relationships and the regulatory complications of larger banks, which slowed their PPP lending. Early in the

Program, therefore, there were significant differences across markets in PPP penetration, based in part on

significant variation in community bank market share. In figure V, we illustrate that 56.6% of all PPP loans

extended in the program’s first 9 days came from community banks, as well as 46.8% of all loans approved

during the first tranche of PPP funding. In contrast, by August 8th, this cumulative share fell to 30.6%, as

other lenders more fully participated in the second round of funding . With clear evidence that community

banks were quicker to disburse PPP loans, we use geographic variation in community bank market share to

instrument for early loan receipt.

While some endogeneity concerns remain—community bank market share may itself be correlated with other

relevant variables—we note that our hypothesis is about the timing of such outcomes. This means that any

potential confounder would not merely have to correlate with community bank market shares, but would

7We are implicitly assuming that anticipation of funds from PPP was not enough for small firms to delay layoffs,

either because firms didn’t have enough cash to allow them to do so, because they were not certain the PPP funds

would arrive, or because small firms are less sophisticated and less forward-looking in their behavior.

11
have to be dynamically correlated with employment outcomes. In other words, endogeneity issues would

have to introduce bias in April and May, and then reverse the sign of this bias in July and August. In OLS

results in Appendix A, community bank shares show a similar pattern to our IV estimates: any alternative

explanation of these effects must discuss why these community bank shares are relevant in some weeks but

not others.

Figure V

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Figure VI

V.A. Empirical Model

We will return to discuss potential violations of the exclusion restriction, but first we describe the familiar

two-equation IV model. In our initial empirical approach, we instrument county-level early PPP loan receipt

with community bank shares, using the variation in local banking markets to predict early PPP receipt. If

the PPP preserved jobs at scale, earlier loans should lead to superior early employment outcomes. However,

since PPP penetration everywhere was ultimately so high—PPP covered over 80% of eligible small business

employment, nationally—those employment effects of early treatment should disappear over time; states’

labor market conditions should converge as PPP coverage converges. In OLS form, our primary specification

is repeated cross-sections of

Yct = β0,s(c)t + β1,tP P Pct0 + X0

ctβ2,t + ct (1)

where Yct is an employment outcome in county c during week t, s(c) is the state of county c, and Xct is a

vector of controls. The first explanatory variable is a state-week fixed effect.

Our key variable of interest, P P P0

ct, is the cumulative percentage of small business payroll covered by a PPP

loan, as of t

0 = April 11th, 2020. We estimate this regression separately for each week in order to illustrate

the dynamic effects of early loan receipt. Note that t

0

is held constant in each regression, while t varies,

so the regression studies the dynamic relationship of employment with early PPP receipt. The treatment

is not that a county received PPP money, but that it received its PPP money early; and the subsequent

convergence of outcomes across counties is the core of our hypothesis. As (t − t

0

) gets large, β1,t should

13
(a) Early Variation in Loan Coverage (b) Late Variation in Loan Coverage

Figure VII

Early in the program, there was significant geographical heterogeneity in PPP receipt. Later in the

program, there was very little, limiting the ability of ultimate PPP receipt to identify the Program’s effects.

converge to 0.

We have two primary outcome variables. The first is the insured unemployment rate (IUR) for regular state

programs (i.e. continuing claims as a share of covered employment) at the county level. Using the IUR avoids

the double counting problems involved in the Pandemic Unemployment Assistance program and captures

high frequency reentry into the workplace. The second outcome variable is approved initial claims at the

county level. In Appendix B, we also study the Census’ Household Pulse Survey at the state level, which

asks whether adults live in a household which has lost employment income during the pandemic, and find

statistically significant results of similar magnitude.

In terms of cumulative small business payroll covered by PPP loans, we observe the greatest county-level

variation early in the program’s first round. By a variety of measures, the highest degree of dispersion in

PPP coverage occurs during the week of April 11th, shortly after the program begins and when the first

round of funds was nearly exhausted. On April 16th, the first tranche of $349 billion dollars had been fully

depleted, and the program closed for 11 days. The resulting delay until the SBA resumed accepting loans

bolsters our timing strategy, as it opened a substantial temporal gap between first-round and second-round

loan recipients. For the first stage, we isolate this week in order to measure early receipt of PPP loans

due to variation in banking relationships. Figure VI shows that early in the program, there was a large

degree of dispersion across counties in how much of eligible payroll covered by PPP loans; by the end of the

program, that dispersion had mostly disappeared, reducing the amount of variation available for studying

the Program’s effects.

Due to the endogeneity concerns discussed above, we instrument P P P0

ct with the share of county-level

deposits held in community banks. We model the endogenous variable at t

0 as

P P Pct0 = α0,s(c)t + α1,tCB Sharec + X0

ctα2,t + ηct (2)

The exclusion restriction is that CB Sharec does not enter (1), save through (2). Figure IX shows the bin

14
scatter plot of this first-stage regression, estimated with no controls for exposition.8 The F-Stat for this

no-controls first stage is 141.4. With the full battery of controls and standard errors clustered at the state

level, the F-Stat is 8.7.9 The full set of controls include state-time fixed effects, county population density,

the pre-Covid county-level Insured Unemployment Rate (IUR), new Covid cases in the past week and four

weeks, and finally new Covid deaths in the past week and four weeks—all at the county-level.

By including state-week fixed effects, this specification exploits within-state variation to identify the parameter of interest, β1,t. States controlled many of the decisions regarding lockdowns and economic restrictions

during March and April, and these states took a variety of approaches to combating the virus, both in terms

of severity and timing. In many cases these measures had a direct impact on employment, for example

when restaurants were prohibited from offering dine-in service. States also determine the major subnational

regulations that affect the local banking markets, as well as the relative capacity of localities to cope with

economic shocks. Cognizant of these heterogeneous economic restrictions, we focus on within-state-week

variation at the county-level.

The spread of the virus itself is another obvious threat to our identification strategy. As Granja et al. (2020)

notes, first-round PPP loans tended to flow to areas that were not as hard hit by the pandemic. Since

Community Banks tend to have stronger market shares in rural areas, it’s possible that our instrument is

negatively correlated with early virus prevalence—which is in turn negatively correlated with unemployment.

We address this concern via two types of control variables. First, we include new Covid-19 cases reported in

county c over the week, as well as the preceding 4 weeks. Additional controls for Covid-19 deaths are included

for the same timeframe. Second, we control for county population density. Finally, we include controls for

the insured unemployment rate for the week of February 5th, 2020, to avoid mistaking pre-existing level

differences for early effects.

Having described the controls and the identification threats they are meant to address, critical endogeneity

and selection concerns remain in the OLS specification. For example, we might be concerned that businesses

in higher distress would be more likely to apply for loans. Alternatively, we might be concerned that

businesses with stronger management will be quicker to apply for loans, or banks might be more inclined

to deal with stronger businesses, regardless of government assurances. Nonetheless, we present OLS results

before turning to our IV results. Figure VIII and Table A.3 present results from an ordinary least squares

estimation of Equation 1. The OLS results show a small negative relationship between PPP Loans and initial

UI claims in mid-April, though the relationship quickly peters out. There is also a negative relationship

between PPP Loans and continuing UI claims, growing slightly over time.

Finally, note that our study cannot fully address the general equilibrium concerns we discussed earlier: if

there are cross-county or cross-firm spillovers because layoffs in one location can cause layoffs in another

location, then no county-level or firm-level analysis will really isolate treatment effects. If the “control

group” is affected by outcomes at a treatment group, any empirical strategy will be problematic. However,

8Because there are over 3,000 counties in the United States, we present all scatter plots averaged by decile bin.

Full scatter plots for all figures are available in Appendix A.

9The associated p-value for this F-stat is 0.0052. Since we do not assume our errors are i.i.d. – we cluster at the

state-level – we also report tests developed in Kleibergen and Paap (2006). For underidentification, the KleibergenPaap rk LM statistic has a p-value of 0.0249 (Chi-Square(1) = 5.03). For weak identification, the Kleibergen-Paap

Wald rk F-statistic is 13.3. For reference, the Anderson-Rubin Wald test has a p-value of 0.000 (Chi-Square(1) =

39.61).

15
this is a core problem in empirical economics, and we won’t be able to surmount it here; indeed the other

studies cited above have not addressed it either.

16
(a) Initial UI Claims

(b) Continuing UI Claims

Figure VIII

OLS estimates of UI Initial and Continuing claims, regressed on early PPP receipt, with control variables.

Both initial and continuing claims are measured as a share of covered employment, at regular state UI

programs, i.e. the insured unemployment rate. 95% confidence intervals displayed.

17
In Appendix B, we replicate this empirical strategy at the state level in both unemployment insurance claims

and in the Census’ Household Pulse Surveys. We find similar results in the latter dataset to our main results

below, and the replication in a different dataset and at a different level of aggregation increases our confidence

in our main results.

V.B. IV Estimates

The dynamic results for the second stage coefficient (β1,t from equation 1) are plotted in Figure X and

presented in Table A.4. The estimated effects of PPP Loans on initial UI claims are at their largest in

early-April, when the first loans are approved. Those effects diminish over the next month and a half, and

are no longer statistically significant in late-May. By June, the results reflect precisely estimated near-zero

effects. Continuing UI claims are naturally slower to change than the initial UI claims. The estimated effect

of PPP loans on continuing claims is large and statistically significant in mid-April. As loan saturation

increased, the effect on continuing UI claims is reduced, starting in mid-May.

These results correspond to the predictions of our hypothesis: counties which received loans at the program’s

outset lost fewer jobs in April and early-May. During the second round of funding, PPP adoption became

near-universal among small businesses, reducing the scale of those effects, and UI claims rates in counties

which received loans early became statistically indistinguishable from claims rates in counties which did

not receive loans early. This dynamic is immediately detectable with initial UI claims, and is echoed in

continuing UI claims. Convergence in continuing claims may be slower for standard reasons of labor market

dynamics; rehiring a job loser, even on temporary layoff, is less effective at keeping employment levels high

than avoiding job losses at the outset.

The units of the regression are all percentages. If properly identified, we can read the coefficient from the

continuing claims regression on April 18th as “a 1 percentage point increase in early payroll coverage by

PPP leads to a 0.43 percentage point decrease in continuing UI claims”. Likewise, the coefficient from the

initial claims regression on the same date can be read as “a 1 percentage point increase in early payroll

coverage by PPP leads to a 0.15 percentage point decrease in the number of approved initial claims for state

UI programs.” Both continuing and initial claims are measured as a fraction of covered employment within

the county. Since we study outcomes during the pandemic recession, these effects should be thought of as

resulting in smaller jumps in claims, rather than in outright reductions.

18
(a) Binscatter of OLS, with No Controls

(b) Binscatter of OLS, on Residuals from Regression of Community Bank Share on all Controls

Figure IX

First stage, with and without controls. As of April 11th, community bank market share is strongly

correlated with PPP coverage of eligible payroll.

19
(a) Initial UI Claims

(b) Continuing UI Claims

Figure X

Instrumental variables regression estimates of the effect of PPP penetration as of April 11th on

employment outcomes. 95% confidence bands shown. Initial and continuing claims at regular state UI

programs are reported as a share of covered employment, i.e. the insured unemployment rate. Regressions

include control variables as described above.

20
To illustrate our “vanishing-effects” hypothesis, Figure XI displays scatter plots of the employment outcome

on fitted first-stage values of PPP. This figure’s purpose is to illustrate the contrast of early effects against

the later absence of effect. On May 2nd , there is a clear, inverse correlation between county-level loans and

county-level IUR; by October 17th, the statistical relationship is much weaker. These results are what was

predicted by our hypothesis.

(a) Early Effects (b) Late Effects

Figure XI

Insured Unemployment Rate vs. fitted first stage values of PPP penetration. Includes control variables.

Dots represent averages per bin in 5-percentile increments.

We emphasize that these estimates should be interpreted as a local average treatment effect (LATE). The

firms which complied with the treatment are those which received a PPP loan, but would have been less

likely to do so if they were domiciled in a county with fewer Community Banks. As we argued in the

introduction, the estimates from Autor et al. (2020) and Chetty et al. (2020) should also be interpreted in

a similar manner to a LATE. Their identification centered on the eligibility cutoff of 500 employees, and

therefore their estimates speak to the effect on mid-sized firms. In contrast, ours speak to the effect on

smaller firms in a weaker financial situation. This is an advantage insofar as this group was the PPP’s

target, and comprised a much larger portion of the PPP’s loan recipients.

Extrapolating local estimates to calculate aggregate effects is valid only in particular circumstances, and

our study does not satisfy them. However, the same is true for the estimates in Autor et al. (2020) and

Chetty et al. (2020). Therefore, we conduct a similar aggregation exercise to offer a comparable result. For

a county to move from the 25th percentile to the 75th percentile of early PPP receipt (a swing of PPP

penetration from 26.7% of eligible payroll to 51.6% of eligible payroll10) would imply a move in the insured

unemployment rate of 12.8 percentage points on May 2nd. By contrast, the difference between the 25th

and 75th percentiles in observed county unemployment rates was 6.0 percentage points during the same

week, suggesting there may be some overstatement in this exercise. Assessed over total national covered

employment of 145.7 million11, moving from the 25th to 75th percentile county leads to a difference of 18.6

million jobs. This number is far closer to our back-of-the-envelope calculations than the numbers furnished

10For the 40 states & D.C. of our sample. For all counties, the interquartile range of April 11th is 23.5% - 51.6%.

11NB: “covered employment” here refers to national employment covered by unemployment insurance, not by PPP,

since the point estimates refer to unemployment insurance claimant rates. County-level employment outcomes are

the unit of observation in our study.

21
in Autor et al. (2020) and Chetty et al. (2020). We focus on moving from the 25th to 75th percentiles

because of the problems associated with LATE; the usefulness of the local study in a linear specification

breaks down further from the treatment and from the mean.

The interpretative exercise relies on an assumption that counties with less community bank market share

would have evolved similarly to counties with more community bank market share. Further, it assumes

that the estimated LATE is in fact an ATE, which we believe is unlikely. However, we reiterate that this

exercise is to offer an aggregate jobs number comparable to Autor et al. (2020) and Chetty et al. (2020). The

estimated aggregate jobs saved in those two papers relies on the same assumption about LATEs and ATEs.

While the county-level assumption is not applicable, an analogous assumption is required: firms with fewer

than 20 employees would evolve similarly to those with around 500 employees; we believe our extrapolation

assumption is more reasonable than theirs.

Extrapolated out, these estimates imply that the current 80% PPP penetration rate nationally would reduce

unemployment by 41 percentage points. To discourage such extrapolation, we note: 1) this is a global

extrapolation from an estimate at the mean and a linear model is obviously not the “true” data-generating

process; 2) as above, our estimate can be interpreted as a local average treatment effect, where compliance

is determined by preexisting banking relationships, and extrapolation from a LATE to a population is

fraught; and 3) since the observations are counties of heterogeneous size and the we have estimated an

average treatment effect, an extrapolation to the population of the US is once again unlikely to be linear.

Finally, it is worth observing that given the confidence bands around the estimate, a 41 percentage point

swing in unemployment is statistically indistinguishable from both Jamie Dimon’s estimate above and the

34 percentage point swing we find in the same calculation in state-level data in the Census’ Household Pulse

in Appendix B.

To summarize, we estimate the employment effects of PPP loans by exploiting variation between counties

which received more of their loans in early April, vs. those who received more of their loans in May and

later. By using the county-level market share of Community Banks as an instrument for early loan receipt,

we avoid the endogeneity issues that arise with both self-selection into loans, as well as lender discrimination

based on firm-performance. We show that early PPP loans had a large and statistically significant effect

on UI claims in April and May. These effects faded later, as the second-round of PPP funding allowed for

near-universal loan coverage, consistent with our hypothesis. We underline that because our study focuses

on the dynamic effects of early receipt, any alleged bias in the instrument must be present in April and May,

but then disappear.

This empirical strategy serves to estimate the employment effect of loans to all firms in a county, which we

highlight as one of this paper’s primary contributions. Identification using the 500-employee eligibility cutoff

is clean, and has a clear-cut argument in its favor. However, it estimates the impact of loans on firms which

were much likelier to have alternate sources of funding, established revenue bases, and a host of other buffers

against a harsh downturn. In contrast, our strategy incorporates the effects on small firms, which were more

likely to cut jobs and go out of business in the absence of PPP. Most importantly, these firms were both the

primary targets of the program, as well as its primary recipients: 96% of all approved PPP loans went to

firms with fewer than 50 employees, in contrast with firms of over 250 employees, which received 0.4% of all

approved loans. When analyzing the employment effects of the PPP, estimating the employment effects on

these smaller firms is of first-order importance.

22
VI. Conclusion

We have discussed drawbacks of some of the early estimates evaluating the Paycheck Protection Program,

and offered back-of-the-envelope calculations that are an order of magnitude greater. We have proposed

an identification strategy which exploits variation in the timing of PPP loan receipt, and implemented it

at the county-level. Such an approach yields evidence for statistically significant, superior labor market

outcomes in counties which received PPP funds faster; these advantages fade over time as PPP penetration

converges toward 80% everywhere. We calculate that moving from the 25th percentile to the 75th percentile

county of PPP penetration leads to an improvement in the unemployment rate of over 12 percentage points,

or, extrapolated nationally, roughly 18.6 million jobs, a number much closer to our back-of-the-envelope

calculations than those of previous studies.

23
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25
Appendix A: Additional Figures

A. Descriptive Statistics & Table of Estimates

TABLE A.1

County-Level Descriptive Statistics at Three Points in Time

11 April 25 April 8 August

Dependent Variables

Initial Claims (% of Covered Employment) 1.6% 1.1% 0.3%

(1.8%) (1.2%) (0,4%)

Continuing Claims (% of Covered Employment) 5.4% 7.2% 4.8%

(4.1%) (5.0%) (3.4%)

Instrument and Independent Variables

Eligible Payroll Covered by PPP Loans (%) 39.2% 58.6% 85.1%

(22.0%) (25.7%) (29.7%)

Community Banks Deposits (% of Total $) 57.5% 57.5% 57.5%

Covid-19 Cases (33.2%) (33.2%) (33.2%)

In the Past Week (per 100K) 31.4 39.8 117.6

(71.5) (160.5) (135.6)

In the Past 4 Weeks (per 100K) 67.9 125.8 476.7

(163.4) (287.1) (472.2)

Covid-19 Deaths

In the Past Week (per 100K) 1.4 2.0 2.5

(6.2) (6.2) (7.6)

In the Past 4 Weeks (per 100K) 2.5 5.9 8.3

(9.6) (18.5) (16.5)

Population Density (People/Square Miles) 296.5 296.5 296.5

(1980.5) (1980.5) (1980.5)

Insured Unemployment Rate on 15 Feb. 2020 1.1% 1.1% 1.1%

(1.2%) (1.2%) (1.2%)

Number of Counties 2,436 2,436 2,436

All descriptive statistics describe county-level means, with standard deviation in parentheses.

Because these statistics describe county-level means, they do not equal national averages. All

counties excluded from the core empirical specifications are also excluded from these data.

Excluded states are CA, FL, HI, IA, KS, MI, MN, NC, OR, and SD.

26
TABLE A.2

Community Bank (CB) and Non-Community Bank (Non-CB) Loan Measures at Three Points in Time

11 April 25 April 8 August

CB Non-CB CB Non-CB CB Non-CB

Loan Measure

Number of Loans 455,951 348,911 758,223 861,996 1,592,393 3,619,742

Total Loan $ (Billions) $83.4 $108.1 $116.4 $203.5 $163.1 $361.9

Jobs Reported (Millions) 8.5 9.2 12.0 17.8 17.4 33.5

Figure XII

27
TABLE A.3

OLS Regressions: Employment Outcomes on Cumulative Share of Small Business Payroll Covered by PPP, as of

the week of April 11th

(1) (2)

Initial UI Claims Continuing UI Claims

February 22 -0.0002 0.0000

(0.0002) (0.0002)

February 29 -0.0005* -0.0001

(0.0002) (0.0003)

March 7 -0.0002* -0.0004

(0.0001) (0.0004)

March 14 -0.0003*** -0.0006

(0.0001) (0.0005)

March 21 0.0007 -0.0007

(0.0009) (0.0010)

March 28 0.0017 0.0003

(0.0023) (0.0020)

April 4 -0.0024 0.0021

(0.0041) (0.0028)

April 11 -0.0025 0.0004

(0.0030) (0.0041)

April 18 -0.0037** -0.0018

(0.0017) (0.0059)

April 25 -0.0025** -0.0057

(0.0012) (0.0062)

May 2 -0.0016 -0.0078

(0.0010) (0.0059)

May 9 -0.0026 -0.0088

(0.0017) (0.0057)

May 16 -0.0008 -0.0108*

(0.0008) (0.0058)

May 23 -0.0004 -0.0098*

(0.0006) (0.0055)

May 30 -0.0009 -0.0084

(0.0012) (0.0053)

June 6 -0.0003 -0.0078

(0.0006) (0.0055)

June 13 -0.0004 -0.0065

(0.0006) (0.0052)

June 20 -0.0006 -0.0066

(0.0006) (0.0048)

June 27 0.0003 -0.0069

(0.0005) (0.0045)

July 4 -0.0001 -0.0065

(0.0005) (0.0041)

July 11 -0.0006 -0.0059

(0.0006) (0.0038)

July 18 -0.0000 -0.0062

(0.0005) (0.0038)

July 25 -0.0002 -0.0066*

(0.0003) (0.0037)

August 1 0.0001 -0.0065*

(0.0002) (0.0033)

August 8 -0.0006** -0.0064**

(0.0003) (0.0029)

August 15 -0.0004 -0.0069**

(0.0003) (0.0027)

August 22 -0.0003 -0.0068**

(0.0003) (0.0026)

August 29 -0.0000 -0.0066**

(0.0002) (0.0025)

September 5 -0.0004* -0.0056**

(0.0002) (0.0025)

September 12 0.0000 -0.0052**

(0.0002) (0.0025)

September 19 -0.0003 -0.0044**

(0.0002) (0.0021)

September 26 -0.0002 -0.0039*

(0.0001) (0.0020)

October 3 -0.0001* -0.0033

(0.0001) (0.0021)

October 10 -0.0004*** -0.0026

(0.0001) (0.0024)

October 17 -0.0002 -0.0021

(0.0002) (0.0028)

Controls Y Y

Notes: Standard errors in parentheses, clustered at the state level.

Each week is regressed separately. *,**, and *** represent significance

at the 10, 5, and 1% level, respectively.

28
TABLE A.4

IV Regressions: Early PPP Receipt on Employment Outcomes

(1) (2)

Initial UI Claims Continuing UI Claims

February 22 -0.009 0.003

(0.010) (0.003)

February 29 0.001 -0.003

(0.001) (0.006)

March 7 -0.001 0.002

(0.001) (0.005)

March 14 -0.003** 0.003

(0.001) (0.006)

March 21 -0.099** 0.038**

(0.041) (0.018)

March 28 -0.156*** -0.025

(0.047) (0.052)

April 4 -0.210*** -0.141*

(0.065) (0.081)

April 11 -0.129*** -0.320***

(0.040) (0.102)

April 18 -0.148*** -0.429***

(0.053) (0.129)

April 25 -0.114** -0.477***

(0.047) (0.136)

May 2 -0.083** -0.515***

(0.033) (0.141)

May 9 -0.089** -0.515***

(0.043) (0.145)

May 16 -0.069* -0.516***

(0.035) (0.147)

May 23 -0.039 -0.503***

(0.024) (0.145)

May 30 -0.040 -0.468***

(0.024) (0.137)

June 6 -0.021 -0.429***

(0.016) (0.126)

June 13 -0.034* -0.368***

(0.020) (0.110)

June 20 -0.036** -0.366***

(0.015) (0.110)

June 27 -0.024** -0.366***

(0.011) (0.103)

July 4 -0.020 -0.360***

(0.013) (0.100)

July 11 -0.025* -0.306***

(0.014) (0.082)

July 18 -0.034** -0.295***

(0.015) (0.076)

July 25 -0.019* -0.270***

(0.010) (0.070)

August 1 -0.020* -0.256***

(0.010) (0.068)

August 8 -0.016* -0.239***

(0.008) (0.060)

August 15 -0.019** -0.200***

(0.009) (0.057)

August 22 -0.015** -0.192***

(0.007) (0.052)

August 29 -0.011* -0.193***

(0.006) (0.054)

September 5 -0.013** -0.179***

(0.005) (0.051)

September 12 -0.007* -0.165***

(0.004) (0.047)

September 19 -0.005* -0.134***

(0.003) (0.037)

September 26 -0.004* -0.094***

(0.002) (0.026)

October 3 -0.005 -0.062***

(0.003) (0.023)

October 10 -0.008* -0.020

(0.005) (0.023)

October 17 -0.003 0.009

(0.002) (0.030)

Controls Y Y

Notes: Standard errors in parentheses, clustered at the state level.

Each week is regressed separately. *,**, and *** represent significance

at the 10, 5, and 1% level, respectively.

29
B. IV Estimates with Alternate Definition of PPP Loan Coverage

These figures are constructed identically to those in the main text, but the main right hand side variable is

defined not as a share of eligible employment supported by PPP. Instead, the main independent variable is

each county’s share of its own total loans received by April 11th, i.e. for each county (dollar value of loans

made through April 11 / dollar value of loans made through August 8).

(a) Initial UI Claims

(b) Continuing UI Claims

Figure XIII

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C. OLS: UI Outcomes Regressions Directly on Community Bank Shares

(a) Initial UI Claims

(b) Continuing UI Claims

Figure XIV

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D. Scatter Plots for All Counties (Censored and Uncensored Vertical Axis)

(a) First Stage Scatter, No Controls (b) First Stage Scatter (Residualized)

(c) Second Stage Scatter, 2 May (d) Second Stage Scatter, 17 October

Figure XV

Appendix B: State-Level Results in the Census’ Household Pulse Survey

In this section we replicate our exploitation of temporal variation in loan receipt at the state level in unemployment claims and in the Census’ Household Pulse survey.

In terms of cumulative small business payroll covered by PPP loans, we observe the greatest state-level

variation early in the program’s first round. By a variety of measures, the most dispersion we see in PPP

coverage occurs during the week of April 11th, shortly after the program begins and when the first round

of funds was nearly exhausted. Note that the program was then closed for ten days, resulting in a delay

before further funds were disbursed, and allowing us to estimate employment variation. For the first stage,

we isolate this week in order to measure early receipt of PPP loans due to variation in banking relationships.

In OLS form, our primary specification is repeated cross-sections of

Yst = β0,t + β1,tP P Pst0 + X0

stβ2,t + st (3)

where Yst is an employment outcome in state s during week t, and Xst is a vector of controls. Our key

32
variable of interest, P P P0

st, is the cumulative percentage of small business payroll covered by a PPP loan,

as of t

0 = April 11th, 2020. We estimate this regression separately for each week in order to illustrate the

dynamic effects of early loan receipt. As in the main text, t

0

is held constant in each regression, while t varies,

so the regression studies the dynamic relationship of employment with early PPP receipt. The treatment

is not that a state received PPP money, but that it received its PPP money early; and the subsequent

convergence of outcomes across states is the core of our hypothesis.

For outcomes, we study the insured unemployment rate at regular state programs (i.e. continuing claims

relative to covered employment), to avoid the double counting problems involved in the Pandemic Unemployment Assistance program, as well as to capture high frequency reentry into the workplace. We also

study the Census’ Household Pulse Survey, which asks whether adults live in a household which has lost

employment income during the pandemic.

The control vector consists of the new Covid-19 cases reported in state s during the week, as well as the

insured unemployment rate for the week of February 5th, 2020. We include these controls due to concerns

that our instrument is correlated with the virus prevalence and pre-Covid unemployment conditions. As

Granja et al (2020) notes, first-round PPP loans tended to flow to states that were not as hard hit by the

pandemic.

Critical endogeneity and selection concerns remain in this OLS regression. For example, we might be

concerned that businesses in higher distress would be more likely to apply for loans. Alternatively, we might

be concerned that businesses with stronger management will be quicker to apply for loans, or banks might be

more inclined to deal with stronger businesses, regardless of government assurances. Nonetheless, we present

OLS results before turning to our IV results. Figures XVI and XVII present results from an ordinary least

squares estimation of Equation 3. There is a weak and statistically insignificant benefit to receiving early

PPP loans.

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(a) Initial UI Claims

(b) Continuing UI Claims

Figure XVI

OLS estimates of UI Continuing claims and Census household pulse employment income expectations,

regressed on early PPP receipt, with control variables. Continuing claims are measured as a share of

covered employment, at regular state UI programs, i.e. the insured unemployment rate. 95% confidence

intervals displayed.

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Figure XVII

OLS regressions of employment outcomes on early PPP receipt.

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Figure XVIII

First stage, no controls. As of April 14th, community bank market share is strongly correlated with PPP

penetration. 95% confidence band displayed.

Due to the endogeneity concerns just discussed, we instrument P P Pst0 with the share of state-level deposits

held in community banks. We model the endogenous variable as

P P Pst0 = α0,t + α1,tCB Shares + X0

stα2,t + ηst (4)

Figure XVIII shows the plot of this first stage regression, estimated with no controls for exposition. The

F-Stat for this no-controls first stage is 50.6. With controls included, the F-Stat moderately declines from

26.5 in early-April to 23.7 in early-May, then remains relatively consistent.

The dynamic results for the second stage coefficient (β1,t from equation 3) are plotted in Figure XIX and

presented in Figure XX. The estimated effects, both in reducing continuing UI claims and reducing loss in

employment income, grow stronger until late-April, then start to attenuate in mid-May. While the effects in

UI data are not statistically significant, the effects in the Census Pulse survey are. Early PPP loan receipt,

as instrumented by community bank shares, leads to lower unemployment rates and less loss in labor income

during April and May. During the second round of funding, PPP adoption became near-universal among

small businesses, reducing the scale of those effects. While the Census Pulse survey provides some evidence

for our hypothesis that there would be strong initial effects of early PPP receipt which would then fade as

PPP saturated the small business economy, that evidence is in the form of diminishing statistical significance

36
over time and would be clearer in the form of precisely estimated zeroes. We do note, however, that this is a

dataset of 51 observations, so diminishing significance as the treatment grows distant in time is unsurprising.

Why are there no significant results in the UI data at the state level, when we do find some at the county

level in the main text? First, as mentioned, there are only 51 observations here, prior expectations should

be dim for finding anything at the state level. Second, for the reasons discussed in Cajner et al. (2020),

there are many reasons why ordinary UI claims data might be much noisier now than in normal times, and

since these programs are administered at the state level, this variation will be correlated with the state fixed

effects we control for at the county-level but cannot control for at the state level.

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(a) Initial UI Claims

(b) Continuing UI Claims

Figure XIX

Instrumental variables regression estimates of the effect of PPP penetration as of April 14th on

employment outcomes. 95% confidence bands shown. ”Loss in income” corresponds to the Pulse

”Percentage of adults in households where someone had a loss in employment income.” Continuing claims

at regular state UI programs are reported as a share of covered employment, i.e. the insured

unemployment rate. Regressions include control variables as described above.

38
Figure XX

IV estimates of early PPP receipt effect on employment outcomes.

The units of the regression are all percentages. If properly identified, we can read the coefficient from the

continuing claims regression on April 22nd as “a 1 percentage point increase in early payroll coverage by

PPP leads to a 0.14 percentage point decrease in continuing UI claims”. Likewise, the coefficient from the

loss in employment income regression on May 20th can be read as “a 1 percentage point increase in early

payroll coverage by PPP leads to a 0.27 percentage point decrease in the number of adults in households

reporting employment income lost during the pandemic.”

39
Figure XXI

Percentage of adults in households which lost income vs. fitted first stage values of PPP penetration.

Includes control variables. Note: the confidence bands in this chart are for two-dimensional representations

of a multivariate analysis, and therefore suffer a generated regressor problem, i.e. the standard errors are

biased downward and the error bands are misleadingly tight. The standard errors and confidence bands in

Figure XIX and XX are calculated using two-stage least squares in repeated cross sections.

To interpret these results, consider that a 10% increase in PPP receipt would reduce the number of adults in

households experiencing income loss by 2.7%. If, as Chetty et al. (2020) argue, the vast majority of reduction

in employment is at the extensive margin, that is equivalent to a change in the unemployment rate of 4.2

percentage points, since only 63% of adults participated in the labor force pre-COVID.

We reiterate that these estimates should be interpreted as a local average treatment effect (LATE). The firms

which complied with the treatment are those which received a PPP loan, but would have been less likely

to if they were domiciled in a state with fewer Community Banks. As we argued in the introduction, the

estimates from Autor et al. (2020) and Chetty et al. (2020) should also be interpreted in a similar manner to

a LATE. Their identification centered on the eligibility cutoff of 500 employees, and therefore their estimates

speak to the effect on mid-sized firms. In contrast, ours speak to the effect on smaller firms in a weaker

financial situation. This is an advantage insofar as this group was the PPP’s target, and comprised a much

larger portion of the PPP’s loan recipients.

Extrapolating local estimates to calculate aggregate effects is valid only in particular circumstances, and our

study does not perfectly satisfy them. However, the same is true for the estimates in Autor et al. (2020)

and Chetty et al. (2020). Therefore, we conduct a similar aggregation exercise to offer a comparable result.

For a state to move from the 25th percentile to the 75th percentile of early PPP receipt (a swing of PPP

penetration from 40% of eligible payroll to 57% of eligible payroll) would imply a move in the unemployment

rate of 7.2 percentage points. By contrast, as of the May 12th survey week, the difference between the 25th

and 75th percentiles in observed state unemployment rates was 5.0 percentage points. Assessed over the

civilian noninstitutional population twenty years and older of 244 million, moving from the 25th to 75th

40
percentile state leads to a difference of 11 million jobs. This analysis assumes that layoffs are distributed

one per household; if multiple people per household lost jobs, the estimate would be biased upward. The

interpretative inference relies on an assumption that states with less community bank market share would

have evolved similarly to states with more community bank market share.

Extrapolated out, these estimates imply that the current 80% PPP penetration rate nationally would reduce unemployment by 34 percentage points. To discourage such extrapolation, we note: 1) this is a global

extrapolation from an estimate at the mean and a linear model is obviously not the “true” data-generating

process; 2) this estimate can be interpreted as a local average treatment effect, where compliance is determined by preexisting banking relationships, and extrapolation from a LATE to a population is fraught;

3) since the observations are states of heterogeneous size and the we have estimated an average treatment

effect, an extrapolation to the population of the US is once again unlikely to be linear; and 4) with such large

extrapolations, it is unlikely the assumption of one layoff per household would continue to hold. Finally,

it is worth observing that given the confidence bands around the estimate, a 34 percentage point swing in

unemployment is statistically indistinguishable from Jamie Dimon’s estimate above.

To consider the potential endogeneity of the instrument, in Figure XXII we contrast pre-COVID-19 trends

in the top tercile of states by community bank market share with the bottom tercile. While the trends

are not parallel before COVID-19, it is the low-community bank shares which are outperformers. Thus our

instrument is not picking up prior outperformance, and the pre-existing trends concern (though there are

others, mostly around regulatory flexibility) would seem to bias our main estimates downward.

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Figure XXII

Pre-COVID employment trends by community bank market share, by tercile.

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