Consumer-Lending Discrimination in the FinTech Era*

Robert Bartlett 

School of Law 

UC Berkeley 

rbartlett@berkeley.edu

Adair Morse 

Haas School of 

Business 

UC Berkeley & NBER 

adair@berkeley.edu

Richard Stanton 

Haas School of 

Business 

UC Berkeley 

rhstanton@berkeley.edu

Nancy Wallace 

Haas School of 

Business 

UC Berkeley 

newallace@berkeley.edu

February 2019 

Abstract 

Ethnic discrimination in lending can occur in face-to-face decisions or in algorithmic scoring. The 

GSEs’ model for pricing credit risk provides us with an identified setting to estimate discrimination 

for FinTech and face-to-face lenders, as well as to offer a workable enforcement interpretation of 

U.S. fair -lending laws using the court’s justification of legitimate business necessity. We find that 

face-to-face and FinTech lenders charge Latinx/African-American borrowers 6-9 basis points 

higher interest rates for purchase mortgages, consistent with the extraction of monopoly rents in 

weaker competitive environments and from profiling borrowers on shopping behavior. In 

aggregate, Latinx/African-American pay $750M per year in extra mortgage interest. FinTech 

algorithms have not removed discrimination, but two silver linings emerge. Algorithmic lending 

seems to have increased competition or encouraged more shopping with the ease of applications. 

Also, while face-to-face lenders discriminate against minorities in application rejection, there are 

reasons to believe FinTechs may discriminate less. 

JEL classification: G21, G28, G23, J14, K22, K23, R30 

Keywords: FinTech; mortgages; credit scoring; algorithmic underwriting; big-data lending; platform loans; 

disparate impact 

_______________________________

*Financial support from the Fisher Center for Real Estate and Urban Economics is gratefully acknowledged. For helpful

comments and suggestions, we thank Sanjiv Das, Andreas Fuster, Andres Liberman, Raghu Rau, Amit Seru, Justin Wolfers, and

seminar participants at U.C. Berkeley, Cornell, the University of Michigan, USC, NYU Stern, Texas A&M, Michigan State

University, Chicago-Booth, Johns Hopkins University, the George Washington-Federal Reserve Board Seminar Series on

Household Finance, the Spring 2018 NBER Corporate Finance meeting, the 2018 NBER Law & Economics Summer Institute,

the 2018 Annual Conference on Empirical Legal Studies, the 2017 Boston JOIM Conference, the 2017 Annual Research

Conference of the Center for International Securities and Derivatives Markets, the 2018 NY Fed./NYU Stern Conference on

Financial Intermediation, the 2018 American Law and Economics Association (ALEA) annual meeting, the 2018 Financial

Intermediation Research Society (FIRS) annual meeting, and the 2018 EFA meeting.

1
I. Introduction 

Minority households hold 17.3% of the $13 trillion (or $2.25 trillion) in U.S. household debt as of 

2017.1

 Any discrimination with respect to payments on such a large balance is thus likely to be material, 

affecting both minority household well-being and lender profits. For example, consider home loans, which 

are the largest component of consumer debt, representing approximately $9.1 trillion of the total. Each extra 

basis point in mortgage interest charged due to discrimination costs minority mortgage holders 

approximately $100 million per year.2

 Likewise, any discrimination with respect to the accept/reject

decision on loan applications is also likely to be material. In 2017 census data, homeownership by white 

ethnicities was 72.4%, but only 48.4% and 43.0% for Latinx and African-American households, 

respectively. If any of this disparity is due to discrimination, it would materially bias the well-being of 

minority households in the United States. 

In this paper, we estimate the level of discrimination in the largest consumer-loan market, 

conventional conforming mortgages securitized by the Government Sponsored Enterprises (GSEs), Fannie 

Mae and Freddie Mac. Using GSE loans allows a novel identification strategy. The GSEs charge each loan 

a guarantee fee (which gets added into the base mortgage rate from the MBS market) that depends only on 

where in a grid of credit score and loan-to-value (LTV) ratios (both observable) the borrower lies. In return, 

lenders are guaranteed against credit risk. Thus, mortgage interest rate differences between loans within a 

given GSE grid cell of credit score and LTV cannot reflect differential credit risk, but must instead reflect 

strategic pricing decisions on the part of lenders. Strategic pricing is not illegal by any means, but, under 

the law, it cannot induce disparate impact.3

 Lenders cannot, even inadvertently, charge a higher mark-up to

protected ethnic groups. Using this novel identification, we find that discrimination is 6-9 basis points (bps) 

in purchase-mortgage markets and 2-4 bps in refinance mortgages. Averaging across the distribution of 

these products in the U.S., lending discrimination currently costs African-American and Latinx borrowers 

approximately $750 million in extra interest per year. 

Consumer lending in the United States is changing rapidly, with loan origination becoming almost 

exclusively algorithmic. A case in point is the Rocket Mortgage of the platform lender Quicken, which is 

the largest-volume mortgage product in the U.S. as of 2018. Algorithmic loan origination is not, however, 

just a feature of FinTech companies. We study the 2,098 largest mortgage lenders (inclusive of all the big 

banks) over the 2012-2018 period, finding that as of 2018, 45% of them offer complete online or app-based 

                                                            

1

 Percent-of-debt estimates are from the 2016 Survey of Consumer Finances. 2

 The calculation assumes at origination the mean mortgage is $175,000, paying 4.0 % on a 30 year fixed contract. 

3

 A disparate impact occurs under U.S. anti-discrimination law when a decision maker's practices do not expressly 

discriminate on a protected characteristic (e.g., race or ethnicity), but nevertheless disadvantage one or more groups 

without a legitimate business justification. Under disparate-impact theory, proof of discriminatory intent is not 

required to establish liability. As noted below, the U.S. Supreme Court has held that the disparate impact theory of 

liability extends to the Fair Housing Act. 

2
mortgage contracting. Value-weighted, that percentage is much higher. Indeed, nearly all of the big banks 

and most small lenders now act as FinTech. This is not simply a mortgage story; one only has to look to the 

emergence of personal lending platforms, such as Capital One and Goldman Sachs Marcus’s site, to see the 

broader transformation in consumer lending. 

With algorithmic credit scoring, the nature of discrimination changes from being primarily 

concerned with human biases – racism and in-group/out-group bias – to being primarily concerned with 

illegitimate applications of statistical discrimination. Even if agents performing statistical discrimination 

have no animus against minority groups, they can induce disparate impact by their use of Big Data variables. 

Whether these changes induce more or less discrimination was previously unknown. We find that FinTechs 

and face-to-face lenders are equally discriminatory in extracting rents from minorities. It is perhaps 

disappointing that discrimination remains even though face-to-face interactions are removed. However, the 

story has two silver linings for the role of FinTechs and algorithmic decision-making. First, in the loan 

accept/reject decision (as opposed to pricing), FinTech lenders (particularly those meeting our most 

restrictive definition of FinTech) reveal less evidence of discrimination than lenders within our overall 

sample. Second, our evidence on time patterns is consistent with discrimination declining for all lenders 

from 2008 to 2015 due to the advent of FinTech startups and the speed of being able to shop around at 

algorithm-based lenders of all sorts. 

At the core of our paper is the importance – in our identification and in the legal setting – of 

statistical discrimination in the new era of algorithmic loan decision-making. Regulators and courts face 

heightened hurdles to identify which Big Data variables can give rise to a successful claim of illegal 

discrimination under U.S. fair lending laws (see Inclusive Communities; 2015).4

 For economists, the courts’ 

struggle to untangle legitimate from illegitimate statistical discrimination is the same problem as handling 

omitted variables in estimating discrimination. Statistical discrimination arises as a solution to a signal 

extraction problem. The signal extraction setting in consumer lending emerges as follows. Economists can 

write down a macro-founded (life-cycle) model of repayment risk that applies to everyone.5

 The problem 

is that some variables in this macro-fundamental model are not observable. The goal of statistical 

discrimination is to reconstruct this hidden fundamental information using observable proxies. 

In the law, lenders can use proxy variables that produce a disparate impact on minority applicants 

but only if the lender can show that these variables have a legitimate business necessity. According to the 

courts, legitimate business necessity is the act of scoring credit risk. Furthermore, according to the courts, 

efforts to use proxy variables that produce a disparate impact for other purposes, including lenders’ earning 

                                                            

4

 The potential for illegitimate statistical discrimination toward protected classes of borrowers was a key aspect of 

Congressman Emanuel Cleaver’s 2017 investigation into FinTech lending. 

5

 Behavioral models may correctly profile individuals on average, but some individuals would be incorrectly profiled, 

which could be deemed discrimination by disparate impact under the law. 

3
of higher profit margins, do not meet this definition. In business terms, any strategic pricing that causes 

disparate impact, even inadvertently, is discrimination in the eyes of the law. 

Lenders have used the legitimate-business-necessity defense to argue that any variable that is 

correlated with default is acceptable. This definition of legitimate business necessity is necessary but not 

sufficient to comply with the court rulings. An example is illustrative. Surely, the high school that a person 

attended is an empirically relevant proxy for hidden wealth, where wealth is the endowment variable in a 

macro-fundamental model of repayment risk. High school, however, may be correlated with ethnicity even 

after orthogonalizing with respect to wealth. If so, using high school would punish, or have disparate impact 

on, some minority households. 

Our economic mapping of these court rulings on disparate impact to legitimacy in statistical 

discrimination yields three punchlines: (a) Scoring or pricing loans explicitly on credit-risk macrofundamental variables is legitimate. (b) Scoring or pricing on a Big Data variable that only correlates with 

ethnicity through hidden fundamental variables is legitimate. (c) Scoring or pricing on a Big Data variable 

that has residual correlation with ethnicity after orthogonalizing with respect to hidden fundamental credit 

risk variables is illegitimate. 

For policymakers, these punchlines suggest that regulators might take an approach of mandating 

that lenders provide proof of legitimacy of Big Data proxy variables using their proprietary data on 

otherwise-hidden fundamental variables such as wealth. This arrangement would be akin to putting the 

burden of value-at-risk modeling on banks, as is done in banking regulation. We discuss this more in the 

conclusion. 

For researchers, these punchlines imply that in the age of Big Data, econometricians require a 

setting in which all legitimate business necessity variables are observable in order to identify discrimination 

without concern for omitted-variable bias. We have been able to find just such a setting, covering a large 

fraction of consumer lending yet free from omitted-variable concerns. We use this setting to document the 

extent to which discrimination is happening in the largest consumer-loan market and to illustrate how 

algorithmic pricing of loans may yet result in discrimination. 

It is well known that, post-crisis, the GSEs (Fannie Mae and Freddie Mac) purchase, securitize, and 

guarantee more than 90% of the conventional conforming mortgage market in the U.S. It is less recognized 

that, post-crisis, the GSE actions fully determine the price of credit risk by their role as guarantors. In 

particular, the GSEs produce a predetermined grid pricing that prices credit risk across loan-to-value and 

credit score buckets. The pricing grid need not be the optimal model for predicting default among all 

4
application variables,6

 but is nevertheless the price lenders must pay the GSE to absorb risk for the MBS 

market. Thus, any deviation from this grid pricing reflects lenders’ competitive agenda in capturing volume 

or profit per mortgage. Because these non-credit-risk objectives are unrelated to creditworthiness, they fail 

to satisfy the legitimate business necessity as determined by the courts. Thus, within the grid, any additional 

correlation of loan pricing with ethnicity is discrimination. 

Our analysis uses a data set that includes never-before-linked information at the loan-level on 

income, ethnicity, loan-to-value ratios, debt-to-income ratios, all contract terms (such as coupon, loan 

amount, installment payment structure, amortization, maturity, loan purpose, and mortgage-origination 

month), and indicators for whether the lender-of-record primarily used algorithmic scoring. 

We find that accepted borrowers in Latinx and African-American ethnic groups pay 7.9 basis points 

more in interest for home purchase mortgages on account of discrimination. Although courts have explicitly 

held that credit risk is the only legitimate business necessity, as economists, we believe the spirit of these 

decisions may perhaps include room for lenders to differentiate loan pricing based on the fixed cost of 

providing a loan, by lender or by geography. We thus additionally include county and lender fixed effects. 

Overall, we find discrimination in these specifications of 5.3 to 7.9 bps for purchase mortgages, or 

approximately 11-16% of lenders’ average profit per loan.7

 Interest rate discrimination is almost identical 

for FinTech lenders (5.3 basis points extra paid by minorities) as for the overall set of lenders (5.6 basis 

points). In the paper text, we address the robustness of this result to other concerns of servicing rights and 

to the use of points. 

How discrimination happens is an important question. Although, we leave a full exploration to a 

separate research project, we can fix ideas. Lenders may be able to extract monopoly rents from minority 

borrowers because such borrowers might be prone to less shopping on average. In refinance loans, we find 

minorities pay just 2-4 bps in higher interest rates. This result is consistent with an interpretation that 

monopoly price extraction of rents is easier in transactions where the borrowers have less experience or are 

acting in a more urgent time frame. Additionally, because lenders may price loans to capture rents in lesscompetitive areas, prices might be higher in financial services deserts, which might have higher ethnic 

minority populations. 

In our setting of the GSE guarantee, if lenders were to discriminate in the accept/reject decision, it 

would imply that money is left on the table. Logic suggests that such unprofitable discrimination must 

                                                            

6

 The actuarially fair GSE guarantee fee (or G-fee) is also a central policy question in the determination of the future 

role of the GSEs in the U.S. mortgage markets (see Elenev, Landvoigt, and Van Nieuwerburgh, 2016; Vickery and 

Wright, 2013.) A standard G-fee is assessed on all mortgages as a percentage of the loan balance and is collected 

monthly (see Fuster, Goodman, Lucca, Madar, Molloy, and Willen, 2013). 7

 According to the Mortgage Bankers’ Association, the average mortgage profit is 50 basis points (see 

https://www.mba.org/x73719). 

5
reflect a human bias by loan officers. This is what we find. Face-to-face lenders reject Latinx and AfricanAmerican ethnicities 3%-5% more often for purchase mortgages. Overall, we find FinTech lenders 

discriminate at roughly the same rate; however, lenders meeting our most restrictive definition of FinTech 

lenders discriminate at far lower rates and, in some specifications, do not appear discriminate at all. Because 

our findings with respect to rejections must rely on proxies for certain variables utilized by the GSEs in 

approving loans, we note that these results are preliminary. But they nevertheless point toward the 

possibility that fully automated underwriting may reduce the incidence of discrimination in loan rejections. 

Our paper contributes to a small but growing literature on discrimination in lending. A large 

literature in labor contributes to the topic of wage discrimination, but even there, our commentary on how 

courts and regulators can consider Big Data use may be informative. The lending discrimination literature 

has lagged the wage literature primarily because of the lack of data on ethnicity or race combined with an 

identification strategy that handles omitted variables in scoring. 

Early studies looking at the raw HMDA data found that minority loan applicants were rejected 

much more often than white applicants even with higher incomes, but did not control for variables not 

collected by HMDA, such as credit history. In a widely cited paper, Munnell, Browne, McEneaney, and 

Tootel (1996) combined HMDA data on loan applications in Boston in 1990 with additional borrower data 

collected via survey by the Federal Reserve Bank of Boston, and found that after controlling for borrower 

characteristics, especially credit history and loan-to-value ratio, white applicants with the same property 

and personal characteristics as minorities would have experienced a rejection rate of 20% compared with 

the minority rate of 28%. 

Much of the more recent literature focuses on the pre-crisis period, usually looking at subprime 

lending. Ghent, Hernandez-Murillo, and Owyang (2014) examine subprime loans originated in 2005, and 

find that for 30-year, adjustable-rate mortgages, African-American and Latinx borrowers face interest rates 

12 and 29 basis points, respectively, higher than other borrowers. Bayer, Ferreira, and Ross (2018) find that 

after conditioning on credit characteristics, African American and Hispanic borrowers were 103% and 78% 

more likely than other borrowers to be in a high-cost mortgage between 2004 and 2007. Similar results 

were obtained by Reid, Bocian, Li and Quercia (2017). 

Cheng, Lin and Liu (2015) use data from the Survey of Consumer Finances to compare mortgage 

interest rates for minority and non-minority borrowers. They find that black borrowers on average pay about 

29 basis points more than comparable white borrowers, with the difference larger for young borrowers with 

low education, subprime borrowers, and women. 

 Focusing on the quality of consumer credit services, Begley and Purnanandam (2018) study the 

incidence of consumer complaints about financial institutions to the CFPB. They find that the level of 

6
complaints is significantly higher in markets with lower income and educational attainment, and especially 

in areas with a higher share of minorities, even after controlling for income and education. 

 In one of the few experimental papers in this area, Hanson, Hawley, Martin, and Liu (2016) show 

that when potential borrowers (differing only in their name) ask for information about mortgages, loan 

officers are more likely to respond, and give more information, to white borrowers. 

Finally, Fuster, Goldsmith-Pinkham, Ramadorai, and Walther (2018) show that the use of machinelearning techniques to evaluate credit quality may result in differential impact on loan provision to minority 

versus non-minority borrowers. This paper conveys important knowledge in how algorithms are utilized in 

mortgage markets. 

There are also related results from other consumer debt markets. For example, Dobbie, Liberman, 

Paravisini and Pathania (2018) look at data from a high-cost lender in the UK and find significant bias 

against immigrant and older loan applicants when measured using long-run profits. However, they find no 

bias when using the (short-run) measure actually used to evaluate loan examiners, suggesting that the bias 

is due primarily to a misalignment of firm and examiner incentives. 

The rest of the paper is organized as follows. In Section II we discuss our multifaceted data. We 

present our methodology for the measurement of mortgage discrimination in Section III, and provide 

statistics showing the role of the GSE pricing grid in practice. Our empirical results are reported in Section 

IV. Section V concludes and offers policy perspectives on moving forward from our findings. We also 

include Appendix A, where we discuss anti-discrimination regulations in U.S. mortgage lending. 

II. Data & Statistics

A key obstacle in prior empirical mortgage discrimination studies has been their reliance on the 

Home Mortgage Disclosure Act (HMDA) data. The HMDA compliance surveys cover 90% of mortgage 

origination in the U.S. (see Engel and McCoy, 2011),8

 and are the only data source with loan-level 

information on applicant ethnicity for both successful and unsuccessful loan applications.9

 HMDA also 

contains applicant income and (nonstandardized) information on lender name. What HMDA lacks is 

information on the contracting structure of the loan (exact date, interest rate, maturity, loan-to-value ratio), 

on the type of loan (fixed, ARM), on the property characteristics (e.g., address), and on the applicant’s 

credit data commonly used by the GSEs and other lenders (credit score, debt-to-income ratio, etc.). 

                                                            

8

 HMDA reporting is not required for institutions with assets (of the entity and its parent corporation) that are below 

$10 million on the preceding December 31 (see http://www.ffiec.gov/hmda/pdf/2010guide.pdf). 

9

 HMDA has missing values on ethnicity (Buchak et al., 2017). We first fill in these missing values using the ethnicname-categorization algorithm from Kerr and Lincoln (2010) and Kerr (2008). Then we check for consistency of our 

results including and excluding these fixes. 

7
A challenge with mortgage loan data in the U.S. has been the lack of unique loan identification 

number and thus the lack of a direct way to link the HMDA data and other datasets containing these missing 

data. We ameliorate this deficiency with a multi-year project of linking loan-level data across the following 

data providers: 

 HMDA (54.8 million single family residential loans accepted and 53.6 million single family 

loan rejections between 2008 and 2015). HMDA data include information on borrower income, 

ethnicity, and lender, as well as geography of the property only at the census tract. 

 ATTOM (77.2 million single family loans between 2008 through 2015). ATTOM data provide 

transaction and assessor information including loan performance data (i.e., prepayment and 

default), lender names and exact location, but very little information on mortgage contract 

terms other than the loan amount. 

 McDash (24.2 million single family residential loans between 2008 and 2015). McDash data 

provide loan-level data compiled by Black Night Financial Services and includes quite 

comprehensive information on the mortgage terms (including interest rates and the zip code of 

the mortgaged property). 

 Equifax (23.7 million single family residential loans that merged with McDash between 2008 

and 2015). Equifax data provide information on other consumer financing balances that are 

held by borrowers in addition to their mortgages. 

Using a machine-learning protocol, we exploit overlapping variables within these four datasets to construct 

a merged data set of accepted loans with performance information, contract terms, and borrower 

information. We describe our machine-learning merging algorithm in detail in Appendix B. A key 

component of the merging was the McDash-ATTOM link, which we accomplished by matching 

performance strings; i.e., matching loans on the flow of events reported in the property registers. The 

Equifax-McDash merge was done by Equifax in compliance with IRB standards, and our residual data are 

anonymized. 

An admitted weakness in our data is in the rejection data. Although the HMDA data includes 

rejection/accept decisions (with ethnicity, income and loan amount) so that we can perform tests of 

accept/reject discrimination, we must be more cautious in interpreting the accept/reject results (as opposed 

to the pricing results) because none of these auxiliary datasets covers loan application rejections. Thus, 

while we know the precise (observable) variables that the GSEs use to determine loan acceptance/rejection 

in their automated underwriting systems, we must augment the rejection data with proxies for credit score, 

debt, and loan-to-value at the census tract level. A census tract is on average 1600 households (4000 

inhabitants), designed by the Census Bureau to reflect relatively uniform economic situations. We describe 

the construction of these proxies in Appendix B. 

8
To standardize our analysis, we filter the data to focus on 30-year fixed rate, single-family 

residential loans, securitized by the GSEs over the period 2008 through 2015. We additionally eliminate 

from our sample any loans made within a census tract covered by the Community Reinvestment Act of 

1977 (CRA), given the potential bias these census tracts would introduce into our empirical analysis.10 The 

final sample consists of 6,865,349 accepted loan applications and 8,648,282 rejected applications with 

application dates between 2008 and 2015. 

We present summary statistics in Table 1. Among rejected and accepted loans, the median loan 

amount was slightly under $200,000, with accepted loans showing evidence of positive skew through the 

maximum loan size ($729,000) and the slightly higher mean loan amount. The mean and median interest 

rate on loans within our sample was approximately 4.3%. Not surprisingly, compared to the average 

accepted loan applicant, the average rejected loan applicant generally had lower income, a slightly lower 

credit score, and a slightly higher LTV. The median purpose for the loan application among accepted and 

rejected loans was to purchase a home, as opposed to refinance a home. 

Table 1 also reveals summary information concerning the types of lending institutions that received 

the loans applications in our sample. Using the list of firms identified as FinTech in Buchak et al. (2017), 

we find that FinTech lenders originated approximately 4.2% of accepted loans and were responsible for 

5.4% of all loan rejections in our sample. As noted previously, however, many conventional lenders now 

regularly utilize online application forms, and we separately identified the 2,098 largest mortgage lenders 

over the 2012-2018 period who made available an online application form as of the summer of 2018. Loans 

originated by these firms constituted over 45% of our accepted loans, and they rejected over 50% of the 

rejected loans in our sample. Table 1 also highlights the dominance of the largest originators in the 

mortgage lending industry: the top 25 originators (by origination volume in their respective loan origination 

year) originated over 50% of all accepted loans and processed nearly 50% of all of our rejected loan 

applications.11

Finally, among accepted loans in our sample, Table 1 shows that 13.2% were made to AfricanAmerican or Hispanic borrowers, while 20.8% of rejected loan applicants were African-American or 

Hispanic. 

III. Method: Using the GSE Lending Process 

When an individual applies for a conforming mortgage (where conforming means the loan size falls 

beneath a federally-set conforming loan limit), the GSE process begins. The lender feeds application 

                                                            

10 Under the CRA Act, financial institutions are required to provide a certain level of lending to CRA districts to 

counter the lack of financial services in lower-income districts. 11 We create a variable of the top 25 mortgage originators per year by matching HMDA lender names with mortgage 

origination statistics obtained from Inside Mortgage Finance. 

9
observables (the credit score, income, liquid reserves, debt-to-income ratio, loan-to-value ratio, property 

value, etc.), into the GSE ‘black box’, an automated underwriter system (Desktop Underwriter for Fannie 

Mae; Loan Prospector for Freddie Mac). The GSE black box produces the accept/reject decision based on 

a specified set of observables contained in the application. If the GSE accepts the loan, and the lender and 

borrower issue the loan, the lender sells the mortgage to the GSE. In return, the GSE compensates the lender 

with a cash transfer.12 The GSE then packages the loan with a pool of similar mortgages into a mortgagebacked security (MBS), issues a default-risk guarantee on this product, and sells it to the MBS market. 

For 30 days, the lender holds some credit risk, but after this time period, the lender is not exposed 

to any prepayment or default risk. The only risk that the lender faces is put-back risk. Put-backs can occur 

when the documentation on income (the IRS form), credit score (the credit history pull), loan purpose 

(residential vs. non-occupancy) or property value (the appraisal) is falsified or missing. After the 2008 

financial crisis and the half decade thereafter, because of put-backs13 and large fines for misrepresentation, 

lenders ceased no-documentation GSE loans and adjusted their policies to remove the potential for falsified 

documentation. The magnitudes of put-backs on post-2010 originations have become a trickle compared to 

the early 2000 issuances.14 Figure 1 plots put-backs over the time horizon of the loan for issuances 2000-

2010. This figure is taken from an Urban Institute publication by Goodman and Zhu (2013). The figure 

supports our assumption that put-back risk for post-2008 issuances had diminished as well as the lack of 

importance of the 30-day exposure of lenders to credit risk. 

III.a. GSE Process in Pricing and Identification of Rate Discrimination 

Within this GSE process, the lender has three decisions to make. The first is pricing. The mortgage 

interest rate that a household sees consists of three parts (see Fuster et al. (2013)). First, all mortgages face 

the same market price of capital, determined by the Base Mortgage Rate, which reflects the primary market 

interest rate for loans to be securitized by the GSEs. In effect, the Base Mortgage Rate reflects the 

compensation demanded by investors in the MBS market, or the credit risk-free rate. 

Second, when a GSE buys a mortgage from a lender, the GSE takes a guarantee fee (or g-fee) to 

cover projected borrower default and operational costs. Starting in March 2008 and adjusted a handful of 

                                                            

12 If the originator is a large volume lender, the lender will transfer loans to the GSE in bulk and, instead of receiving 

cash for the mortgages, the originator receives back an MBS with a pool of similar-characteristic mortgages produced 

by that lender. (Sometimes these MBS products have mortgages originated by other lenders to fill out the MBS, but 

one should think of this pool as primarily being the lender’s own issuances.) These MBS products are equally 

guaranteed by the GSE, but because the lender issuer also retains servicing rights, the lender may be exposed to the 

extra servicing costs (i.e., additional phone calls and outreach) that happen when loans become delinquent. For this 

reason, we show all of our results with and without the large volume lenders. 

13 The GSEs put back $4.2 billion of pre-crisis loans in 2010 alone (American Banker, July 14, 2016). 14 A large number of put-backs by a lender will induce the GSEs and the MBS market to add a discount to the price 

for products from this lender. These overall lender adjustments can be controlled for with firm fixed effects. 

10
times since then, this g-fee varies in an 8×9 matrix of LTVs and credit scores to reflect varying credit risk 

across the GSE grid. Figure 2 depicts a typical GSE grid of Fannie Mae, also called the Loan Level Price 

Adjustments (LLPAs) (see FHFA 2000; 2010; 2011; 2012; 2013 as well as Fuster and Willen, 2010). In 

practice these one-time fees are commonly converted into monthly “flow” payments, which are added into 

the interest rate as rate pass-throughs to borrowers. In quoting rates to customers, originators utilize rate 

sheets that expressly incorporate both the Base Mortgage Rate (generally reflected as the “par rate”) as well 

as LLPA adjustments (colloquially referred to as “hits” to the par rate). 

The third component of pricing comes from lenders discretion in quoting rates in excess of the par 

rate and any LLPA “hits” and rate quoting to incorporate strategic volume positioning and monopoly renttaking. Lenders might use monopoly-like pricing based on the competition environment of a location (e.g., 

in areas of collusion and in financial desert environments) or as a rent extraction strategy against borrowers 

who shop around less. 

In Figure 3, we graphically depict the importance of the LLPA “grid” for purposes of pricing loans. 

In panels (a) and (b) we display histograms of borrower interest rates among approved loans by treatment 

(treat=1 is the set of Latinx and African-American borrowers). Panel (a) shows a histogram of the raw data, 

revealing a wide distribution of rates for both control and treated loans, as one might expect given the length 

of our sample period and the large number of loans in the sample. However, when we level interest rates 

within the grid by subtracting out the year-grid cell mean, Panel (b) shows a dramatic reduction in the 

distribution of interest rates for both groups of borrowers, highlighting the central role of the LLPA grid in 

determining interest rates for GSE mortgages. Notably, visual inspection of Panel (b) shows greater mass 

in the higher rates for the ethnic treatment histogram, suggesting that (even controlling for GSE pricing 

grid) minority borrowers are paying higher interests. 

Much like this visual exercise, our identification relies on the observation that within a cell of the 

GSE grid, disparities in interest rates for similar grid borrowers reflect markup strategies arising from 

discretionary pricing, not omitted variables in credit-risk scoring. There is no disparity in fair pricing of 

credit risk (which equals the market rate plus any LLPA adjustments) inside the grid because the price is 

dictated by the GSE.15 In particular, for application i occurring in the month-year t, we estimate: 

interest rate௜ ൌ α LatinxAfricanAmerican௜ ൅ μீௌா௚௥௜ௗ ൅ μ௠௢௡௧௛_௬௘௔௥ ൅ ε௜௧, (1) 

                                                            

15 Fundamentally, the GSEs have elected to price credit based on two observable variables (LTV and credit score) that 

proxy for lifecycle variables in a macro-fundamental model of repayment risk. Under U.S. fair lending law, any 

discrimination in loan pricing that arises from this parsimonious pricing model should be justifiable as a legitimate 

business necessity given that these variables proxy for repayment risk. For the same reason, however, any disparate 

impact in loan pricing within the GSE grid is unexplainable by this model and represents illegitimate discrimination 

that is unrelated to the life cycle variables the GSEs have chosen to utilize in pricing credit risk. 

11
where the mortgage interest rate is regressed on the indicator for the applicant being a Latinx or African 

American, GSE grid fixed effects μீௌா௚௥௜ௗ and month-year fixed effects μ௠௢௡௧௛_௬௘௔௥. Because the grid 

fully prices credit risk, and the month-year fixed effects pick up changes in the risk-free rate, any correlation 

of prices with ethnicity inside the GSE grid is thus discrimination. 

 Notably, our identification strategy permits us to identify not only the incidence of discrimination 

in loan pricing but also the legal consequence of this discrimination under U.S. fair-lending laws. This 

stems from the fact that, as discussed in Appendix A, courts have consistently limited the legitimate 

business necessity defense to a lender’s use of structural variables and loan practices that seek to ascertain 

creditworthiness.16 Evidence of within-grid discrimination would thus lack a legitimate business necessity 

defense given the GSEs’ decision to price credit with respect to just two observable variables pertaining to 

creditworthiness. That said, the interpretation of U.S. fair lending laws continues to evolve, and we would 

find it unsurprising were a court to permit at some point a lender to advance as a legitimate business 

necessity defense the ability to recoup differential overhead costs for originating loans in different 

geographic locations (which may correlate with applicants’ ethnicity). Thus, we also utilize an alternative 

specification to account for this possibility which includes lender fixed effects, μ௟௘௡ௗ௘௥, and county fixed 

effects, μ௖௢௨௡௧௬: 

interest rate௜ ൌ α LatinxAfricanAmerican௜ ൅ μீௌா௚௥௜ௗ ൅ μ௟௘௡ௗ௘௥ ൅ μ௖௢௨௡௧௬ 

൅ μ௠௢௡௧௛೤೐ೌೝ

൅ ε௜௧,

(2) 

This specification forces the appealing interpretation on the discrimination coefficient to be the differences 

in average interest rate charged to a minority applicant “as compared to that offered to an ethnic majority 

applicant by the same lender” and “as compared to rates offered to others in that same county”. Although 

appealing econometrically, this rigor throws out some of the variation in which we are interested. Thus, we 

interpret our results in a range from equation (1) to equation (2). 

                                                            

16 See A.B. & S. Auto Service, Inc. v. South Shore Bank of Chicago, 962 F. Supp. 1056 (N.D. Ill. 1997) (“[In a 

disparate impact claim under the ECOA], once the plaintiff has made the prima facie case, the defendant-lender must 

demonstrate that any policy, procedure, or practice has a manifest relationship to the creditworthiness of the 

applicant...”). See also Lewis v. ACB Business Services, Inc., 135 F.3d 389, 406 (6th Cir. 1998) (“The [ECOA] was 

only intended to prohibit credit determinations based on ‘characteristics unrelated to creditworthiness.’”); Miller v. 

Countrywide Bank, NA, 571 F.Supp.2d 251, 258 (D. Mass 2008) (rejecting argument that discrimination in loan terms 

among African American and white borrowers was justified as the result of competitive “market forces,” noting that 

prior courts had rejected the “market forces” argument insofar that it would allow the pricing of consumer loans to be 

“based on subjective criteria beyond creditworthiness.”) 

12
III.b. GSE Process in Accept/Reject Decisions and Identification of Rejection Discrimination 

The second and third decisions the lender makes concern the accept/reject decision. Upon receiving 

an approval in the GSE underwriter system, the lender can still choose to reject the application. If there 

remains no credit risk post-transacting, money would be left on the table. Why would a lender choose to 

reject a GSE-accepted applicant? (i) The lender might feel that a particular borrower reflects additional putback risk. As we have argued, such put-back risk is so small, especially in the latter half of our sample, that 

even if this put-back risk were residually correlated with ethnicity (which is not established), it would not 

be able to explain any material differences that we find in rejection rates. Thus, this argument would amount 

to a biased belief affecting loan decisions. (ii) The lender might be directly racist or have other in-group 

biases. (iii) The lender might prefer to cater to white ethnicities for their clientele. None of these 

explanations falls under legitimate business necessity. 

The third decision that a lender makes is an ex ante one. A loan officer might deter a potential 

borrower from applying, and this influence might be unequal across ethnicities. This argument has two 

sides, however. Loan officers might discourage legitimate borrowers because of biases and racism. 

Conversely, loan officers may inform their in-group preferred clientele of weaknesses in an application to 

assist the potential applicant. If we had a perfect set of both accepted and rejected applicants’ data with all 

observable variable used in a GSE’s black box, we could recreate the GSE’s black-box algorithm and 

eliminate this possibility from biasing our results. Since our data and identification weakness concerns this 

point, we interpret our results with some caution. 

Our estimation of rejection discrimination for application i in the year y is: 

rejection௜ ൌ α LatinxAfricanAmerican௜ ൅ fሺHMDA income௜, HMDA loan amount௜ሻ

൅ μ௬௘௔௥ ൅ ε௜௧.

(3) 

Rejection is an indicator for an application being rejected by the GSE underwriter system. The fሺ⋅ሻ function 

is a non-parametric function of the original HMDA data for income and loan amount. Since we do not know 

the exact scoring function of lenders on these variables, we implement splines (21 splines for income and 

47 splines of loan amount) rather than a linear form to capture step functions, which we understand is the 

standard practice of lenders. We control for year fixed effects μ௬௘௔௥, rather than month fixed effects because 

HMDA does not provide precise dates for the rejections. This is our baseline estimation, with the complete 

set of data on rejections given by HMDA. However, we can do better by including a function of variables 

which the GSEs disclose as entering into the GSE black box underwriter system, proxied at the census tract 

(1600 households) level, where the census tract c refers to that of the property: 

13
rejection௜ ൌ α LatinxAfricanAmerican௜ ൅ fሺHMDA income௜, HMDA loan amount௜ሻ

൅ gሺLTV௖, credit score௖ሻ ൅ μ௟௘௡ௗ௘௥ ൅ μ௖௢௨௡௧௬ ൅ μ௬௘௔௥ ൅ ε௜௧.

(4) 

To capture the distribution within the census tract, we include the 25th, 50th, and 75th percentile of LTV and 

credit score in the census tract in gሺ⋅ሻ. As in our rates estimations, we include, in some specifications, fixed 

effects for lender and county to control for cost differentials in providing financial services. 

IV. Results 

IV.a. Interest Rate Discrimination Estimates - Main 

Table 2 presents within-GSE grid estimates of interest rate discrimination for 30-year fixed-rate 

mortgages that are approved and originated. Because lenders’ pricing strategies vary by mortgage type, we 

present estimates for purchase mortgages (columns (1) and (2)) separately from refinance mortgages 

(columns (3) and (4)). Columns (1) and (3) are a no-covariate model, reported to present the overall mean 

difference. Columns (2) and (4) are our full credit-risk model of equation (1), containing the 72 GSE grid 

fixed effects to capture the pricing in the grid plus the month-year fixed effects to capture market-price 

fluctuations.17

The overall mean difference in the purchase-mortgage interest rate between white ethnicities and 

Latinx/African-American ethnicities is 0.090%, 9.0 basis points. Of this amount, column (2) shows that 1.1 

basis points are explained by the credit-risk model (monthly dummy variables and the GSE grid effects), 

leaving 7.92 basis points of discrimination. Also of interest in column (2) is the ability of the credit-risk 

model to explain about 73% of the variation across nearly 1.5 million mortgages. 

The interpretation of these main-credit-risk-model results is that conditional on being given a loan, 

African-American and Latinx borrowers pay an average of 7.92 basis points more than other borrowers for 

their purchase mortgage. The unexplained (one minus the R2

) 27% of the variation may reflect strategic 

pricing on borrowers’ location (perhaps due to collusion or financial deserts) or perhaps due to borrowers’ 

behavioral characteristics (perhaps a lack of shopping). In the context of residential mortgage loans this 

pricing differential is economically meaningful for both borrowers and originators. For borrowers, 

imposing an additional 7.92 basis points per year in interest for a 30-year fixed-rate amortizing loan has a 

present value of approximately 1 percentage point of the loan balance assuming a benchmark annual interest 

rate of 3.5%. Likewise for originators, sourcing mortgages with an 8 basis point premium on the coupon 

would represent a material increase in profits over the Mortgage Bankers’ Association mean profit of 50 

basis points. 

                                                            

17 Our estimates are almost identical if we instead use GSE grid fixed effects interacted with the month-year dummies.

14
For refinance mortgages (columns (3) and (4)), we identify a still very significant but economically 

smaller amount of price discrimination. Column (4) reports that within the full credit risk model, lenders 

discriminate against minorities by 3.6 basis points. The disparity between purchase and refinance mortgage 

discrimination suggests that borrower sophistication and hurriedness matter. Refinancing borrowers are, by 

definition, experienced and may be in less of a hurry to re-contract than the average purchase-mortgage 

borrower, who may be time constrained to bid on a house on the market. 

To put these magnitudes in content, Panel B shows a back-of-the-envelope calculation of interest 

paid. The total U.S. mortgage market is $12.19 trillion.18 Assuming the existing float of mortgages 

comprises 75% refinance loans and 25% purchase loans and that African-Americans and Latinx borrowers 

make up 13.3% of the total (in our data), we find that discrimination in mortgage interest rate costs AfricanAmericans and Latinx approximately $750 million extra in interest annually. 

IV.b. Robustness of Interest Rate Discrimination Estimates 

Table 3 introduces lender and county fixed effects to the main credit-risk model specification of 

Table 2. The point of this table is two-fold. First, from a court’s perspective, we want to produce estimates 

that are immune from arguments that our discrimination estimates in Table 2 are a function of differing 

costs of providing loans by lender and by geography. The courts have not ruled that legitimate business 

necessity includes locational or lender fixed costs, but we can imagine such an argument might arise. 

Second, from an econometrician’s point of view, including lender fixed effects allows an appealing withinlender comparison as robustness. We find that lenders indeed discriminate within their organizations. Latinx 

and African-Americans pay 5.3 basis points more for purchase mortgages and 2 basis points more for 

refinance mortgages. Although it is important for the robustness of our result to show that our results hold 

within lender and within geography, it is likely that we are throwing away some true discrimination in these 

estimations. 

Three additional concerns might confound the interpretation of our estimates: the use of points, 

servicer risk, and the designation of ethnicity. Table 4 presents a series of tests for these concerns. All 

estimations are in the form of the full credit risk specification of Table 2, columns (2) and (4). 

The term “paying points” refers to the borrower’s act of paying a lump sum to a lender to reduce 

the interest rate (also called the yield spread premium). An interpretation consistent with our results is that 

ethnic minorities are more fully utilizing their cash for downpayment, leaving no cash to pay points to 

reduce the rates. If, on average, white ethnic groups were to be using points more, it would imply lower 

interest rates inside the credit risk grid model without implying discrimination. We test for this for purchase 

                                                            

18 Source: Federal Reserve. 

15
(panel A) and refinance (panel B) mortgages in column (1) of Table 4. In particular, we limit the sample to 

borrowers precisely at the 0.80 LTV threshold. This set is likely to contain borrowers who are scraping up 

funds to just make the downpayment required to meet an LTV of 0.80, which is the LTV threshold at which 

borrowers are exempt from purchasing mortgage loan insurance. Going over this threshold also results in a 

higher loan interest rate. Figure 3 illustrates how important the LTV threshold is. Column (1) show that 

discrimination is higher for these borrowers, inconsistent with our results being driven by (positive) points. 

Another points story consistent with our results is that Latinx and African-American borrowers 

may be paying negative points (incurring a higher interest rate) to get a rebate in cash to pay closing costs. 

Although we cannot see the pricing in our dataset, conversations with mortgage brokers suggest that the 

interest rate costs of taking these yield rate spreads are high. Thus, if a borrower could pay a slightly lower 

downpayment, retaining some cash for closing costs, without inducing any interest rate increase, this choice 

would be optimal for most borrowers. (This need not be true in the very short term, but applies only on 

average.) The borrowers who are likely to be able to slightly decrease their downpayments while not 

increasing their rates are those not facing the LTV=0.80 threshold. Thus, in column (2) of the panels of 

Table 4, we consider the robustness of our results to negative points by estimating our full credit risk model 

for the sample of LTV borrowers who are unconstrained in LTV in the grid. Specifically, we sample 

borrowers who have LTVs between (0.70, 0.74) and (0.75, 0.795), not at the edge of the GSE grid bucket 

in LTV. The results in these columns are very similar to those in Table 3. 

The third robustness concern taken up in Table 4 involves the role of residual risk for lenders who 

hold the GSE-guaranteed MBS or those who service these loans. GSE loans are special in that most lenders 

do not retain servicing rights in the GSE process nor do they hold the asset on their balance sheets once the 

loan is put through the GSE system. This is not always true for the very large lenders, who provide a GSE 

with a pool of mortgages, and thus are repaid in kind (the GSE pays the lender with MBS rather than in 

cash). These large lenders may hold the MBS on their balance sheets and may be servicers of their own 

mortgages. In such a situation, the GSE still guarantees the loan. However, servicing costs surely increase 

with delinquent or defaulting loans. Thus, if the GSE grid does not perfectly model the underlying credit 

risk (e.g., because it does not incorporate other fundamental variables such as wealth), a large lender might 

rationally implement a better pricing model using fundamental variables to estimate hidden servicing cost 

risk. Likewise, a large lender who plans to hold some balance sheet MBS prepayment risk that is not 

guaranteed by the GSE may implement a better-than-the-grid model to estimate prepayment risk. Since 

both of these actions would imply adjustment in prices for credit risk, they would be deemed legitimate 

business necessity by the courts.19 

                                                            

19 Note that minorities on average prepay less in aggregate statistics, which means our main estimate is probably 

conservative on this point of large lenders pricing prepayment risk differentially for minorities. 

16
In column (3) of Table 4, panel A and B, we estimate our full credit-risk specification, but limiting 

the sample to the non-top 25 volume lenders for every year. The results are not materially different from 

those in Table 2, with the magnitudes being somewhat larger in Column (3) of Table 4 for both purchase 

mortgages and refinance mortgages in the lender and county fixed-effects specification. 

Finally, we note that there is the potential for inaccuracies in our estimation in discrimination due 

to errors in identifying borrower ethnicity. The ethnicity of a borrower in our dataset is determined by 

combining self-reported data from HMDA and, for mortgages in HMDA that lack an indicator of borrower 

ethnicity, borrower’s likely ethnicity based on an ethnic-name-categorization algorithm from Kerr and 

Lincoln (2010) and Kerr (2008). Given the possibility that this algorithm misclassified some borrowers’ 

ethnicities, the final column of Table 4 estimates the same specifications utilized in Columns (2) and (4) of 

Table 4 using only observations where borrower ethnicity is provided by HMDA. For both purchase 

mortgages (panel A) and refinance mortgages (panel B), our estimates for discrimination in pricing are 

slightly higher than the estimates obtained in Columns (2) and (4) of Table 4. Our results are therefore not 

driven by potential errors in identifying borrower ethnicity using the ethnic-name-categorization algorithm. 

IV.c. Interest Rate Discrimination Estimates – FinTech Lenders 

With robustness concerns alleviated, we can move to one of our fundamental objectives, estimating 

price discrimination by face-to-face versus algorithmic lenders. Face-to-face lenders have additional 

freedom to discriminate, not just because of racism and in-group bias, but because they may use soft 

information differentially across borrowers. For example, imagine someone comes into a loan officer’s 

office saying that she must have a mortgage today to make an offer on a house she has seen. The loan 

officer, if maximizing for the lender, would quote a higher rate than if the applicant mentioned that she is 

shopping around. If loan officers apply this soft information differentially, or simply quote biased rates 

every time for a particular minority, higher discrimination should be seen in face-to-face lenders. 

Algorithmic lending requires using data from applicants on geography or other variables to induce 

the algorithm to quote a higher price. Face-to-face lenders may be doing this too, in their preparation of 

different rate sheets by branch (or, in the case of small mortgage issues, simply in having particular zone 

pricing). 

Table 5 shows the results for FinTech lenders. We do not have a perfect indicator as to which 

lenders are FinTechs, or, said more broadly, which lenders have platform loans that require no face-to-face 

contracting. The sample in columns (1) to (3) consists of the FinTech platforms from Buchak et al (2017). 

However, this sample is too restrictive. Although their sample covers 47 non-banks (6 of which are FinTech 

platforms, including Quicken) and 31 banks, the lenders only cover 41,832 purchase and 110,870 refinance 

mortgages. 

17
The sample in columns (2) to (4) adds to this group of “platform lenders” any lenders who, as of 

summer 2018, have a full mortgage application process online. We manually search for the existence of a 

full on-line mortgage application for each of the 2,098 largest volume lenders in our sample. A full 

mortgage application means that all pulling of data from the applicant (tax forms, bank information, credit 

report, etc.) is online and the borrower is provided an accept/reject decision as well as a rate. The only 

residual underwriting that remains for the lender concerns the house (appraisal, lien check, etc.). We found 

that 945 of these, 45%, including almost all of the big banks, have a fully on-line mortgage application. 

This set, however, is somewhat too inclusive, in that we could not discern when these lenders began to 

provide algorithmic mortgages. Thus, we are looking for consistency in these two measures of FinTech. 

Looking at columns (1) and (2) in Table 5, we find that FinTech purchase-mortgage discrimination 

is 5.34 bp for the restrictive definition and 6.8 bp for the inclusive FinTech definition. These numbers are 

similar to each other and are also very similar to those in Table 3. For refinance mortgages, likewise, our 

Table 5 estimates are consistent both with each other and with the roughly 2 bp of discrimination we report 

in Table 3. We conclude from this that the FinTechs do no better at discrimination than face-to-face 

borrowers. It must be the case that any loan-officer discrimination that is removed by not seeing faces is 

added back by algorithms that better predict which borrowers can be captured at higher rates. 

However, the story has a couple of silver linings. The first can be seen in Figure 5, which shows 

discrimination by two-year windows of loan issuance. Discrimination has monotonically declined between 

2008-2009 and 2014-2015. Although we cannot prove causality, this result could be due to competition 

from the platforms and/or the ease of shopping around made possible by online applications. To look for 

evidence consistent with this interpretation, we plotted (not displayed in the current draft) Figure 5 by 

FinTech or not. The pattern looks the same. Thus, the pattern seems to reflect growing competition edging 

out the possibility for rents. We also plotted the same pattern for FHA mortgages. The pattern is the 

opposite. The non-GSE market seems to be discriminating more, and the GSE market, whose processes 

have become easier with algorithms directly connected to the GSE underwriter systems have become more 

uniform in pricing. 

The second silver lining for the role of FinTechs comes in rejection decisions, to which we now 

turn our attention. 

18
IV.d. Rejection Discrimination Estimates – Face-to-face and FinTech Lenders 

Table 6 presents the discrimination estimated in the accept/reject decision. The dependent variable 

is an application rejection indicator, where a loan is deemed accepted if the lender offers a loan to a borrower 

even if the borrower does not take it up. The table presents six different samplings of the data, covering 

more than 15 million observations, starting with the full sample of data for columns (1) to (3). This is close 

to the full population of applications processed as 30-year, fixed-rate, single-family-home GSE mortgages, 

run through the GSE underwriting system, except for nonconforming loans using the system for 

convenience. For each sample, we present three models. The first includes the application data that HMDA 

provides – ethnicity, income, loan amount, and year. We include 21 piecewise linear functions of income 

and 47 of loan amount. The second model of each sample adds census-tract-level data to proxy for the other 

application data that HMDA does not record – LTV and credit score. We add three percentiles (the 25th, 

50th, and 75th) of these variables to capture the distributional spread in the census tract. We also add the 

census tract average loan amount. The third model adds lender and county fixed effects. 

In columns (1) to (3) of Panel A, we find results consistent with residual discrimination against 

Latinx and African-Americans. Focusing on column (3), lenders reject minority ethnicities 3.2% more often 

for purchase loans, after controlling for the variables (or proxies for the variables) entering the black-box 

GSE underwriting system and after doing within-lender and within-location analysis. Columns (4) to (6) 

focus on small (non-top-25) lenders. We believe these results are more robust, as these lenders are less 

likely to bear residual default risk in servicing delinquent borrowers. Column (6) shows that these lenders 

reject minority ethnicities 6.4% more often. 

Note that the R-squared is 0.33 in this model. It is well known that R-squareds are severely 

downward biased in Linear Probability Models (which we use because of all the fixed effects). A useful 

interpretation from Gronau (1998) is that the R-squared estimates the improvement in the probability that 

the model accurately predicts a rejection over a naïve model that would be correct fifty percent of the time. 

In this light, our model does very well in picking up much of the residual variation. 

A 6.4% higher rejection rate is very material. Table 1 reports that the mean rejection rate is 56%, 

so our results suggest that rejections increase by nearly 9% of their unconditional value. 

Table 7 considers discrimination by the FinTechs. As before, we use an under-inclusive (columns 

(1) to (3)) definition of algorithmic decision-making in the measure of FinTech platforms from Buchak et 

al (2017), and we use an over-inclusive (columns (4) to (6)) measure of the set of lenders who, as of summer 

2018, have a full mortgage application online. The truth is somewhere in between, so we average the 

coefficients to interpret. The average of the Latinx and African-American coefficients in Panel A for 

columns (3) (a coefficient of 0.0259) and (6) (a coefficient of 0.0344) is 0.0301. This rate is only slightly 

less than the 3.16% higher rejection rate for purchase mortgages across all lenders in Table 6, Panel A. 

19
Since face-to-face discrimination should play little role among FinTech lenders, the presence of any 

discrimination in rejection rates among FinTech lenders represents something of a puzzle, and potentially 

reflects our need to proxy for certain variables used in the GSEs loan underwriting software. Nonetheless, 

the fact that discrimination among FinTech is less than half that of small lenders as well as the much lower 

discrimination among lenders meeting the most restrictive definition of FinTech provides some reason to 

believe FinTech underwriting results in lower discrimination in rejection rates than conventional face-toface lending. 

Finally, our analysis of discrimination in rejection rates merits one additional consideration given 

that discrimination in rejections is not akin to discrimination in pricing for the lenders. Firms that 

discriminate in rejecting qualified applicants are losing money, because of the GSE guarantee and 

securitization process. Thus, it is rational that FinTechs might remove the potential for racism and ingroup/out-group biases that cost lenders money. What is strange is the face-to-face lenders still allow 

procedures that allow-for discrimination. However, we must revisit our caveat to provide an alternative 

interpretation. 

We stated that a potential selection issue may influence our estimates. In particular, a story 

consistent with our results is that white ethnicity loan officers are more active in informing white applicants 

of weaknesses in their profile ahead of the formal application process. This alone would not bias our results 

except that we do not have all the variables (in their exact form rather than in a proxy form) that enter into 

the GSE underwriter. Our results showing a higher (3%-6% higher, depending on the specification) 

rejection rate for Latinx and African-Americans is large for such a bias to explain, but we cannot rule out 

that at least part of this coefficient is due to selection rather than discrimination. We are content with this 

alternative explanation, noting that this is still a form of discrimination in that the “help” provided to white 

ethnicities is still welfare-improving. 

This potential selection/observability bias cannot, however, affect our FinTech results, in that 

algorithms do not perform these “help” functions of pre-informing potential applicants of profile 

weaknesses, or, if they do, they do not perform these functions differentially by ethnicity. Given the GSE 

guarantee, understanding the channel that can explain why FinTech lenders discriminate in rejection rates 

must therefore await further research and the release of additional borrower credit variables in revised 

HMDA data that the CFPB intends to make available to the public beginning in 2019.20 Based on our 

findings, we summarize our rejection results as follows. 

                                                            

20 The CFPB recently revised the HMDA to include additional variables concerning borrower credit characteristics. 

See https://www.consumerfinance.gov/about-us/newsroom/consumer-financial-protection-bureau-announces-policyguidance-disclosure-home-mortgage-data/. 

20
Our rejection results suggest two inferences. First, our results suggest that face-to-face lenders 

either over-reject Latinx and African-American borrower applications, or they help ethnic majority 

borrowers in pre-advising of application weaknesses, both of which imply a non-neutral welfare implication 

for minority applicants. Second, discrimination in rejection rates also exists among FinTech lenders when 

we define this category broadly; however, even with this definition, they discriminate less than small 

lenders, with platform lenders showing even lower levels of discrimination in the accept/reject decision. 

This represents a second silver lining in the role of algorithms in consumer lending. 

V. Conclusion 

Using a unique data set of mortgage loans that includes never-before-linked information at the loanlevel on income, ethnicity, loan-to-value and other contract terms, we exploit the unique structure of the 

GSE pricing grid to identify discrimination in mortgage loan pricing. Overall, we find that conditional on 

obtaining a loan, African-American and Latinx pay a higher (7.9 bps) interest rate for purchase mortgages 

and a 3.6 bps higher mortgage interest rate for refinance mortgages. Given the emergence of algorithmic 

credit scoring, we further investigate the level of discrimination among FinTech lenders, which we find to 

be roughly identical (6.8 basis points extra paid by minorities in purchase mortgages) to the overall set of 

lenders. In cross-sectional analysis of rejection rates, we find that African-American and Latinx borrowers 

are 3%-6% more likely to be rejected for a mortgage than other borrowers in the overall sample, but 

somewhat lower discrimination in rejection rates among FinTech lenders, suggesting these lenders may 

cater to borrowers discriminated against by face-to-face lenders. 

We also make contributions to debates within the economic literature concerning the identification 

of illegitimate discriminatory use of protected characteristics in credit screening and how the economist’s 

challenge maps to the legal concept of “disparate impact” liability under U.S. fair lending laws. 

Fundamentally, the GSEs have elected to price credit based on two observable variables (LTV and credit 

score) that proxy for lifecycle variables in a macro-fundamental model of repayment risk. Under U.S. fair 

lending law, any discrimination in loan pricing that arises from this parsimonious pricing model should be 

justifiable as a legitimate business necessity given that these variables proxy for repayment risk. However, 

for the same reason, any disparate impact in loan pricing that is unexplained by this model points to 

illegitimate discrimination in loan pricing that is unrelated to the legitimate life cycle variables the GSEs 

have chosen to utilize in pricing credit risk. This is precisely what we find. Having access to the GSE’s 

model for pricing credit risk thus permits us to identify empirically the incidence of discrimination as well 

21
as to make the further assertion that any such discrimination is not justified by a legitimate business 

necessity under U.S. fair lending laws.21

This framework has important implications for regulators and courts as lenders seek to exploit Big 

Data to improve credit scoring in consumer lending, particularly given the Supreme Court’s recent 

requirement that a plaintiff must establish a “robust” causal connection between a lending practice and an 

alleged statistical disparity in lending outcomes. As our analysis suggests, loan outcomes that depart from 

a lender’s credit scoring model that includes variables used to proxy for legitimate lifecycle variables in a 

macro-fundamental model of repayment risk is sufficient to identify empirically the existence of 

problematic discriminatory lending practices. This conclusion follows because, under the lender’s own 

model, any such discrimination is unjustified by the legitimate business necessity of evaluating 

creditworthiness. At the same time, courts’ cabining of the legitimate business necessity to evaluating 

borrower’s creditworthiness places the burden on lenders to justify empirically why the use of particular 

Big Data variables is an empirically valid proxy for a legitimate lifecycle variable even after 

orthogonalizing with respect to their estimate of this variable.22 

Finally, our findings with regard to the lower levels of discrimination of rejection rates in FinTech 

mortgage lending suggest that, in addition to the efficiency gains of these innovations, they may also serve 

to make the mortgage lending markets more accessible to African-American and Latinx borrowers. While 

we caution that this finding is tentative given that we lack access to all of the variables used by the GSEs 

to approve or reject loans, this positive evaluation of FinTech lending bodes wells for the needed expansion 

of U.S. residential mortgage markets as they continue to recover from the 2009 crisis. On a more cautionary 

note, however, the discipline imposed by the GSEs’ underwriting and pricing requirements may help 

explain why the incidence of discrimination in lending is not greater than we find within our sample of 

                                                            

21 Under the Supreme Court’s Inclusive Communities framework, our identification strategy should provide a means 

to satisfy the Court’s requirement that a plaintiff first establish a “robust” causal connection between a lending practice 

and an alleged statistical disparity before shifting the burden to the lender to establish a legitimate business necessity 

for the practice. While we lack information as to what specific practices are producing these discriminatory effects in 

GSE lending, the fact that the GSE underwriting and pricing structure relies entirely on observable life-cycle variables 

also permits the conclusion that, whatever practices produced these outcomes, they cannot be justified as lenders’ 

attempts to proxy for unobservable life-cycle variables. Accordingly, the practices producing these discriminatory 

effects must be lacking in a legitimate business necessity, to the extent courts limit this defense to factors related to a 

borrower/applicant’s credit risk. 22 This approach necessarily assumes a lender has taken the effort to justify the inclusion of a Big Data variable as a 

proxy for some lifecycle variable in a model of credit risk. Even if these variables are unobservable for a particular 

applicant (e.g., income growth), lending institutions can nevertheless observe how these variables are distributed in 

the population using historical data. For instance, a lender that chooses to use the high school of an applicant as a 

proxy for the applicant’s income growth would have presumably made this decision due to some historical relationship 

between observed high school and observed income growth among a sample of other individuals for which the lender 

has historical data. Under our approach, the burden is on the lender to demonstrate empirically why, within this 

sample of data, high school attended is not correlated with ethnicity after orthogonalizing high school and observed 

income growth. 

22
loans originated by FinTech lenders or among lenders overall. To date, this less-well-understood role of the 

GSEs has not been considered in GSE reform proposals, nor is it obvious how such a role could be supported 

within a fully privatized, conventional conforming secondary mortgage market. Likewise, outside of 

mortgage lending, the possibility remains that, lacking formal underwriting and pricing standards, lending 

algorithms might proxy for unobservable life-cycle variables by relying on observable characteristics (e.g., 

the name of an applicant’s high school) that produce illegitimate statistical discrimination. 

 

23
Appendix A: Legality of Explicit and Statistical Discrimination 

Discrimination in residential real estate lending is policed primarily by two federal statutes, the Fair 

Housing Act of 1968 (FHA) and the Equal Credit Opportunity Act of 1974 (ECOA). The FHA, which is 

administered by the Department of Housing and Urban Affairs (HUD), prohibits taking any action affecting 

the terms of the transaction on the basis of a borrower/applicant’s race, religion, national origin, gender, 

familial status (e.g., family size or marital status), or handicap. The ECOA builds on the FHA by expanding 

the set of protected classes of borrowers. Under these laws, a lender in the mortgage market may not engage 

in, among other things, (i) refusing to extend credit, (ii) using different standards in determining whether 

to extend credit, or (iii) varying the terms of credit offers (e.g., loan amount, interest rate, duration and type 

of loan) on the basis of one of the above-mentioned protected characteristics. 

Under the FHA and the ECOA, either a wronged borrower/applicant or the relevant administrative 

agency has the authority to bring a civil action against a lender. As in other areas of anti-discrimination law, 

however, the scope of these anti-discrimination mandates has been determined largely by the legal standards 

that courts have created for a successful claim. Historically, the primary method by which lenders were 

found to have engaged in prohibited discrimination was by a showing of disparate treatment --- specifically, 

a showing that a lender had treated borrowers or applicants differently because of a protected characteristic. 

For example, in Watson v. Pathway Financial, 702 F. Supp. 186 (N.D. Ill. 1988), an African-American 

couple successfully sued a lender under the FHA for rejecting their mortgage application because of 

delinquent credit card accounts. While the lender’s justification was a potentially legitimate basis for 

denying credit, the court found the lender had violated the FHA because the applicants produced evidence 

that the lender had approved at least six applications from white borrowers showing similar delinquencies. 

Disparate treatment claims have also been used to prohibit traditional redlining, in which a 

mortgage lender refuses to make loans to entire geographic areas because of their racial composition (Gano; 

2017). Thus, for economists, the two sets of variables that are explicitly illegal under disparate treatment 

are indicator variables of the protected category (e.g., ethnicity in our case) and geography. 

In addition to disparate treatment claims, a private party or governmental agency can bring a claim 

of lending discrimination under the FHA or ECOA under a disparate impact theory. In contrast to disparate 

treatment claims, claims of disparate impact do not involve any allegation of intentional discrimination in 

how a lender treats applicants/borrowers but rely instead on the fact that lending practices that are facially 

neutral in their treatment of different groups nonetheless fall more harshly on a protected category of 

applicants. For instance, in a joint policy statement on the enforcement of the ECOA and the FHA, the 

24
Department of Justice, HUD, and all federal banking regulators provided the following example as an 

illustration of a lending policy that could give rise to a disparate impact claim23: 

Example. A lender’s policy is to deny loan applications for single-family residences for less than 

$60,000. The policy has been in effect for ten years. This minimum loan amount policy is shown to 

disproportionately exclude potential minority applicants from consideration because of their income levels 

or the value of the houses in the areas in which they live. 

Despite agency approval of disparate impact theory, the ability of parties to pursue disparate impact 

claims has been hindered for two principal reasons. First, the existence of a disparate impact, by itself, is 

insufficient to prove illegal discrimination. Rather, after a plaintiff demonstrates that a lending practice 

produces a disparate impact, the lender can defend the practice as justified by a legitimate business 

necessity, provided no alternative policy or practice could achieve the same goal with less discriminatory 

effect. 

Second, neither statute expressly defines the standard for proving a disparate impact violation. In 

fact, it was not until the Supreme Court’s 2015 decision in Texas Department of Housing and Community 

Affairs v. Inclusive Communities Project that the Court even approved the disparate impact framework 

under the FHA. While the Court interpreted the FHA to permit disparate impact claims, it also required 

plaintiffs to establish a “robust” causal connection between a specific practice and an alleged statistical 

disparity. The Court imposed this requirement to ensure that “ ‘[r]acial imbalance ...does not, without more, 

establish a prima facie case of disparate impact’ and thus protects defendants from being held liable for 

racial disparities they did not create.” Since the decision, however, lower courts have struggled with how 

plaintiffs can establish proof of causality under this standard for a disparate impact case to move forward 

(Fulks, 2017). 

The economic literature has something to contribute to the question of specifying a standard of 

proof.24 Starting in the 1970s (Phelps, 1972; Arrow, 1973; Aigner and Cain, 1977), economic research 

shifted discrimination discussions to the statistical theory of discrimination rather than taste-based 

discrimination associated with Gary Becker. The language of statistical discrimination maps well to the 

legal theory of disparate impact. However, like courts considering disparate impact claims, economists have 

struggled with positing causality in statistical discrimination estimation. Criticisms of inadequate data and 

omitted-variable biases in estimation have plagued this literature.25

                                                            

23 See Policy Statement on Discrimination in Lending, 59 Fed. Reg. 18,266 (Apr. 15, 1994). 24 See commentary by <cite>GlassmanVerna:16</cite> and Winfield v. City of N.Y., No. 15CV5236-LTS-DCF, 2016 

WL 6208564, at *1 (S.D.N.Y. Oct. 24, 2016); Cty. of Cook v. HSBC N. Am. Holdings Inc., 136 F. Supp.3d 952, 955 

(N.D. Ill. 2015). 

25 Cited data omissions in empirical tests of discrimination include omitted information such as the loan to value ratio 

at origination, the debt to income ratio, all the contracting elements of the mortgage, the property characteristics and 

25
For example, Schafer and Ladd (1981) find some evidence of interracial pricing differences, and 

Black and Schweitzer (1985) find indications of differences in loan terms. Yet Sandler and Biran (1995) 

critique any use of statistics in legal proceedings of discrimination claims due to the prevalence of omittedvariable bias and poor identification strategies in these studies. Not surprisingly, the mortgage literature on 

discrimination parallels these broader patterns, primarily focusing either on the methodological difficulties 

in providing robust statistical evidence for discrimination (See Black, Schweitzer, and Mandell (1978); 

Kaye (1982); Maddala and Trost (1982); Rachlis and Yezer (1993) on the efficacy of specific types of 

legislation.)26

However, we think that more can be accomplished. Consistent with the Supreme Court’s decision 

in Inclusive Communities, the theory behind statistical discrimination provides guidance as to how one can 

demonstrate a “robust” causal connection between lending practices and racial or ethnic disparities in 

lending outcomes. In statistical discrimination, agents (in our case lenders seeking to screen applicants) 

have limited information and no animus against racial groups. Statistical discrimination arises as a solution 

to a signal extraction problem. Agents seek to reconstruct hidden information as to the expected 

creditworthiness of applicants using observable proxies. 

We have already established that the use of a protected variable (ethnicity) is illegal under disparate 

treatment. However, could other variables be legal proxies for hidden information, even if they are 

correlated with ethnicity? Under the legal theory of disparate impact, statistical discrimination is allowable 

for “legitimate business necessity.” Economic theory guides us that the meaning of this phrase is that a 

variable legitimately appears as a structural variable that maps the ability of a household to repay a loan to 

their economic fundamentals. In particular, one can write down a life-cycle model in which cash flow for 

repayments emerge from the current borrowing position (debt), cost of borrowing (credit score), income 

(in levels, growth, and risk), wealth, and regular expense levels (cost of living measures). 

Legally, limiting the legitimate business necessity defense to structural variables that predict 

creditworthiness is entirely consistent with legal precedent assessing whether the refusal to extend credit 

violates federal antidiscrimination law.27

                                                            

exact address, the applicant’s ethnicity, gender, credit history and debt burden levels, lender and regulatory 

characteristics, all at the loan level for successful and unsuccessful applicants. 

26 For example, the Fair Housing Act of 1968, the Equal Credit Opportunity Act of 1974, the Home Mortgage 

Disclosure Act of 1975, the Community Reinvestment Act of 1977, and the HMDA amendments to the Financial 

Institutions Reform Reregulation and Enforcement Act of 1989 

27 See A.B. & S. Auto Service, Inc. v. South Shore Bank of Chicago, 962 F. Supp. 1056 (N.D. Ill. 1997) (“[In a 

disparate impact claim under the ECOA], once the plaintiff has made the prima facie case, the defendant-lender must 

demonstrate that any policy, procedure, or practice has a manifest relationship to the creditworthiness of the 

applicant...”). See also Lewis v. ACB Business Services, Inc., 135 F.3d 389, 406 (6th Cir. 1998) (“The [ECOA] was 

only intended to prohibit credit determinations based on ‘characteristics unrelated to creditworthiness.’”). 

26
Further, cases involving “reverse redlining” --- the extension of credit to a protected class on 

inferior terms than those offered to non-protected applicants --- also suggest that only structural variables 

that map to repayment risk should be permitted under the “legitimate business necessity” defense in 

assessing disparate impact claims in rates, conditional on the extension of credit.28

Thus, whether in loan rejection or in loan pricing, the use of any of the foregoing structural variables 

should be considered “legitimate business necessity” under disparate impact theory, even if this variable 

statistically loads on (punishes) a particular protected category. As a matter of economic theory and legal 

doctrine, this is legitimate statistical discrimination. 

What if a lender cannot see one of these variables (wealth, say), but can see a variable (e.g., the 

name of the high school attended) that correlates with wealth? Under the theory of disparate impact, this 

variable should be allowable if it is only disparately impacting the pool of applicants in sorting on wealth. 

In other words, conditional on latent hidden wealth, high school is orthogonal to loan decisions. A slightly 

less stringent assumption would also be consistent: conditional on latent hidden wealth, the impact of high 

school on loan decisions must be the same for one ethnic group as the other (ignorability). 

 

                                                            

28 See, e.g., Miller v. Countrywide Bank, NA, 571 F.Supp.2d 251, 258 (D. Mass 2008) (rejecting argument that 

discrimination in loan terms among African American and white borrowers was justified as the result of competitive 

“market forces,” noting that prior courts had rejected the “market forces” argument insofar that it would allow the 

pricing of consumer loans to be “based on subjective criteria beyond creditworthiness.”) 

27
Appendix B: Algorithm for Merging Mortgage Data Sets 

Since there are no unique mortgage loan identifiers in the U.S., we develop an algorithm using 

machine-learning techniques to match loans found in two independent datasets: the McDash dataset, which 

contains loan-level data compiled by Black Night Financial Services, and the ATTOM dataset, which 

provides detailed property transaction and ownership information in addition to a time-series history of all 

recorded mortgage lien events such new mortgage originations, prepayments, REO, foreclosure, short sales, 

and arms-length sales and loan payoffs. The algorithm relies on matching distinct loan characteristics such 

as origination date, loan amount, and termination and distress events to precisely match each loan with its 

counterpart in the other dataset using a modified k-nearest neighbor classifier (see Hastie, Tibshirani, and 

Friedman, 2009; James, Witten, Hastie, and Tibshirani, 2015). 

Merge process for newly originated fixed-rate GSE loans 

There are two steps to our merging process. The first is to serialize the ATTOM data into a record 

format in which each property is assigned a full event-history string for each mortgage loan and the priority 

and performance of these mortgage loan positions. ATTOM provides very comprehensive geographic 

coverage for mortgage originations and terminations at the property-level for 90 percent of property 

Assessor’s Pin Numbers and all the mortgage and lien recording for each property all sourced from public 

records. The second stage of the merging process is to employ functions for k-nearest neighbor algorithms 

using sklearn.neighbors in Python to fit radial kernels using BallTree. The k-nearest-neighbors classifier 

implements learning based on the 25 nearest neighbors in the corresponding zip code within the McDash 

mortgage data that also records the loan contract features and a loan-level string of performance 

characteristics. We represent each loan in each data set with a thirteen element vector that includes: 1) the 

original loan balance; 2) the lien position, 3) the origination date of the loan, 4) the ending date of the loan, 

5) the foreclosure date of the loan (maybe null), the prepayment date of the loan (maybe null), 6) the 

appraised market value of the property, 7) the loan purpose (refinance or purchase), 8) loan distress dates 

(may be null), 9) loan REO date (may be null), 10) loan liquidation date (may be null), 11) short sale 

indicator variable (may be null), 12) interest rate type (fixed or variable loans), 13) property transaction 

value if there is a sale. Each of these elements is assigned a category subscore between 0 and 1. Each 

subscore is then squared to achieve a greater penalty for matches on key elements such as the loan amount. 

The category subscore is then scaled by a factor which represents the category’s importance to the match 

quality relative to other elements used in the match. Each category factor is an integer between 0 and 100, 

and the sum of the category factors is 100. Our scoring algorithm (get.score in Python’s sklearn) takes into 

account the 13 different elements of each matched pair of loans to calculate a score. The score roughly 

corresponds to the estimated error for each match, measured in hundredths of a percent. Thus, a match score 

28
of 1689 corresponds to a 16.89% chance of an incorrect match, or an 83.11% confidence in the match. We 

use only matches with scores of 2000 or less. For GSE loans originated between 2008 and 2015, we obtain 

a 90% merge rate. 

Our prior machine learning strategy is less applicable for the merge of the HMDA data to McDash 

data, because we have only origination data in HMDA and as well as a greatly reduced set of loan 

characteristics at origination including: 1) the regulator type, 2) the loan type (conventional), 3) property 

type (1 to 4 single family residential properties, 4) loan purpose (refinance or purchase), 5) occupancy 

status, 6) original loan amount, 7) MSA, state, county and census tract, 8) self-reported borrower and coborrower ethnicity (African-American, Hispanic, Asian, Caucasian, unknown), 9) borrower/coborrower 

gender, 10) borrower annual income, 11) year of origination, 12) denial reason if loan application is 

rejected, and 13) lender name. For this merge, we instead we standardize the lenders names between 

ATTOM and HMDA and then merge these data sets using lender names, loan amount, lien type, and the 

loan purpose fields. Of the 30.6 million originations in the HMDA data sets and about 10.4 million GSE 

loans, we successfully merged 60% of these loans. We then merged the ATTOM to McDash merged data 

to the ATTOM to HMDA merged data using the crosswalk developed with the k-nearest neighbor algorithm 

and we obtained a final data set of 7.73 million loans that are single family fixed rate GSE loans with 

maturities of 120 to 360 months originated between 2008 and 2012 (6.87 of those loans have maturities of 

360 months). 

The Equifax-enhanced subsample of originations 

To obtain a final data that includes the full spectrum of underwriting characteristics that would have 

been available to the lender, we again merge the HMDA/ATTOM/McDash data set of fixed rate GSE loans 

that were originated between 2008 and 2015 to the McDash loans that are merged to Equifax data. The 

Equifax-enhanced originated loan sample includes other consumer credit positions of the borrowers such 

as: the total sum of retail, consumer finance and bank card balances; total student loan debt, total auto loan 

debt (sum or auto finance and auto bank debt); age of the borrower, and Vantage 3.0 score. 

The HMDA sample of rejected conventional loans 

The second important class of loans in our data set includes all of the conventional conforming 

loans in HMDA for 1-4 family residential borrowers whose loan applications were either denied by the 

originator, approved but not accepted by the borrower, withdrawn by the applicant, or the loan application 

file was closed for incompleteness. These data include information on: 1) the regulator type, 2) the loan 

type (conventional), 3) property type (1 to 4 single family residential properties, 4) loan purpose (refinance 

or purchase), 5) occupancy status, 6) original loan amount, 7) MSA, state, county and census tract, 8) self29
reported borrower and co-borrower ethnicity (African-American, Hispanic, Asian, Caucasian, unknown), 

9) borrower/co-borrower gender, 10) borrower annual income, 11) year of origination, 12) denial reason if 

loan application is rejected, and 13) lender name. They also include information on demographic and 

minority representation in the census tract in which the collateral on the loan is located. These variables 

include: the Federal Financial Institutions Examination Council (FFIEC) tract median family income to 

MSA median family income as a percentage, median family income for tract in thousands of dollars, tract 

population in thousands, tract minority population as a percentage, tract number of owner occupied units 

in thousands, tract number of 1- to 4-Family units in thousands. 

HMDA does not include information on the credit score or the loan-to-value ratio of the rejected 

loan application files. For this reason, we proxy for the rejected loans’ unobserved loan-to-value ratios by 

computing the census tract-level first and third quartiles and the medians of the loan to value ratios for the 

GSE loans reported in McDash. Similarly, we proxy for the unoabserved FICO scores by computing the 

census tract-level median and the first and third quartiles for the Vantage 3.0 Scores reported for GSE loans 

in the McDash data. 

30
References 

Abowd, J., F. Kramarz, and D. Margolis, 1999, High wage workers and high wage firms, Econometrica 

67, 251–333. 

Aigner, D. J., and G. Cain, 1977, Statistical theories of discrimination in the labor market, Industry and 

Labor Relations Review 30, 175–187. 

Aldrich, S. P. B., W. R. Greenberg, and B. S. Payner, 2001, A capital markets view of mortgage servicing 

rights, Journal of Fixed Income 11, 37–53. 

Arnold, D., W. Dobbie, and C. S. Yang, 2018, Racial bias in bail decisions, Quarterly Journal of 

Economics (forthcoming). 

Arrow, K., 1973, Higher education as a filter, Journal of Public Economics 2, 193–216. Ayres, I., 2002, 

Outcome tests of racial disparities in police practices, Justice Research and Policy 4, 131–142. 

Bayer, P., F. Ferreira, and S. L. Ross, 2016, The vulnerability of minority homeowners in the housing 

boom and bust, American Economic Review: Economic Policy 8, 1–27. 

Bayer, P., F. Ferreira, and S. L. Ross, 2018, What drives racial and ethnic differences in high-cost 

mortgages? The role of high-risk lenders, Review of Financial Studies 31, 175– 205. 

Becker, G. S., 1957, The Economics of Discrimination (University of Chicago Press, Chicago). 

Begley, T. A., and A. K. Purnanandam, 2017, Color and credit: Race, regulation, and the quality of 

financial services, Working paper, University of Michigan. 

Bertrand, M., and S. Mullainathan, 2004, Are Emily and Greg more employable than Lakisha and Jamal? 

A field experiment on labor market discrimination, American Economic Review 94, 991–1013. 

Black, H., R. L. Schweitzer, and L. Mandell, 1978, Wage discrimination: Reduced form and structural 

estimates, American Economic Review, Papers and Proceedings 68, 186–191. 

Black, H. A., and R. Schweitzer, 1977, A canonical analysis of mortgage lending terms: Testing for 

lending discrimination at a commercial bank, Urban Studies 22, 13–19. 

Blinder, A. S., 1973, Wage discrimination: Reduced form and structural estimates, Journal of Human 

Resources 8, 436–455. 

Buchak, G., G. Matvos, T. Piskorski, and A. Seru, 2017, Fintech, regulatory arbitrage, and the rise of 

shadow banks, Working Paper 23288, NBER. 

Card, D., A. R. Cardoso, and P. Kline, 2016, Bargaining, sorting, and the gender wage gap: Quantifying 

the impact of firms on the relative pay of women, Quarterly Journal of Economics 131, 633–686. 

Elenev, V., T. Landvoigt, and S. van Nieuwerburgh, 2016, Phasing out the GSEs, Journal of Monetary 

Economics 81, 11–132. 

31
Engel, K. C., and P. A. McCoy, 2011, The Subprime Virus: Reckless Credit, Regulatory Failure, and 

Next Steps (Oxford University Press, New York). 

Federal Housing Finance Administration, 2009, Fannie Mae and Freddie Mac single-family guarantee 

fees in 2007 and 2008. 

Federal Housing Finance Administration, 2010, Fannie Mae and Freddie Mac single-family guarantee 

fees in 2008 and 2009. 

Federal Housing Finance Administration, 2011, Fannie Mae and Freddie Mac single-family guarantee 

fees in 2009 and 2010. 

Federal Housing Finance Administration, 2012, Fannie Mae and Freddie Mac single-family guarantee 

fees in 2010 and 2011. 

Federal Housing Finance Administration, 2013, Fannie Mae and Freddie Mac single-family guarantee 

fees in 2012. 

Fortin, N., T. Lemieux, and S. Firpo, 2011, Decomposition methods in economics, in O. Ashenfelter, and 

D. Card, eds., Handbook of Labor Economics, volume 4A, chapter 1 (Elsevier, Amsterdam). 

Fulks, M., 2017, Living with the robust causality requirement for disparate impact under the Fair Housing 

Act, Baker Donelson. 

Fuster, A., P. Goldsmith-Pinkham, T. Ramadorai, and A. Walther, 2018, Predictably unequal? the effects 

of machine learning on credit markets, Working paper, Federal Reserve Bank of New York. 

Fuster, A., L. Goodman, D. Lucca, L. Madar, L. Molloy, and P. Willen, 2013, The rising gap between 

primary and secondary mortgage rates, Federal Reserve Bank of New York Economic Policy Review 19. 

Fuster, A., and P. Willen, 2010, $1.25 trillion is still real money: Some facts about the effects of the 

Federal Reserve’s mortgage market investments, Public Policy Discussion Paper 10-4, Federal Reserve 

Bank of Boston. 

Gano, A., 2017, Disparate impact and mortgage lending: A beginner’s guide, University of Colorado Law 

Review 88, 1109–1166. 

Ghent, A. C., R. Hernandez-Murillo, and M. T. Owyang, 2014, Differences in subprime loan pricing 

across races and neighborhoods, Regional Science and Urban Economics 48, 199–215. 

Glassman, A. M., and S. Verna, 2016, Disparate impact one year after Inclusive Communities, Journal of 

Affordable Housing & Community Development Law 25, 12–24. 

Gronau, Reuben, 1998, A useful interpretation of R-squared in binary choice models, Princeton 

University Working Paper. 

Gruenstein Bocian, D., K. Ernst, and W. Li, 2008, Race, ethnicity and subprime home loan pricing, 

Journal of Urban Economics 60, 110–124. 

32
Hanson, A., Z. Hawley, H. Martin, and B. Liu, 2016, Discrimination in mortgage lending: Evidence from 

a correspondence experiment, Journal of Urban Economics 92, 48–65. 

Hastie, T., R. Tibshirani, and J. Friedman, 2009, The Elements of Statistical Learning: Data Mining, 

Inference, and Prediction, second edition (Springer, New York).

Haughwout, A., C. Mayer, and J. Tracy, 2009, Subprime mortgage pricing: The impact of race, ethnicity, 

and gender on the cost of borrowing, Staff Report 368, Federal Reserve Bank of New York. 

Heckman, J., 1998, Detecting discrimination, Journal of Economic Perspectives 12, 101–116. 

Heckman, J., and P. Siegelman, 1993, The Urban Institute audit studies: Their methods and findings, in 

M. E. Fix, and R. J. Struyk, eds., Clear and Convincing Evidence: Measurement of Discrimination in 

America, 187–258 (Urban Institute Press, Washington, DC). 

James, G., D. Witten, T. Hastie, and R. Tibshirani, 2015, An Introduction to Statistical Learning with 

Applications in R (Springer, New York). 

Kaye, D., 1982, Statistical evidence of discrimination, Journal of the American Statistical Association 77, 

773–783. 

Kerr, W. R., 2008, Ethnic scientific communities and international technology diffusion, Review of 

Economics and Statistics 90, 518–537. 

Kerr, W. R., and W. F. Lincoln, 2010, The supply side of innovation: H-1B visa reforms and U.S. ethnic 

invention, Journal of Labor Economics 28, 473–508. 

Kline, P., 2011, Oaxaca-Blinder as a reweighting estimator, American Economic Review: Papers & 

Proceedings 101, 532–537. 

Maddala, G., and R. P. Trost, 1982, On measuring discrimination in loan markets, Housing Finance 

Review 1, 693–709. 

Munnell, A. H., L. Browne, J. McEneaney, and G. Tootel, 1996, Mortgage lending in Boston: 

Interpreting HMDA data, American Economic Review 86, 25–54. 

Oaxaca, R., 1973, Male-female wage differentials in urban labor markets, International Economic Review 

14, 693–709. 

Oaxaca, R. L., and M. R. Ransom, 1999, Identification in detailed wage decompositions, Review of 

Economics and Statistics 81, 154–157. 

Phelps, E. S., 1972, Money, public expenditure and labor supply, Journal of Economic Theory 5, 69–78. 

Rachlis, M., and A. Yezer, 1993, Serious flaws in statistical tests for discrimination in mort- gage 

markets, Journal of Housing Research 4, 315–336. 

Reid, C.K., D. Bocian, W. Li, and R. G. Quercia, 2017, Revisiting the subprime crisis: The dual mortgage 

market and mortgage defaults by race and ethnicity, Journal of Urban Affairs, 39, 469-487. 

33
Ross, S., 2002, Paired testing and the 2000 Housing Discrimination Survey, in W. Foster, A. F. Mitchell, 

and S. E. Fienberg, eds., Measuring Housing Discrimination in a National Study: Report of a Workshop 

(National Academies Press, Washington DC). 

Ross, S. L., M. A. Turner, E. Godfrey, and R. R. Smith, 2008, Mortgage lending in Chicago and Los 

Angeles: A paired testing study of the pre-application process, Journal of Urban Economics 63, 902–919. 

Sandler, A. L., and J. Biran, 1995, The improper use of statistics in mortgage lending discrimination 

actions, in A. Yezer, ed., Fair Lending Analysis (American Bankers Association, Washington, DC). 

Shafer, R., and H. Ladd, 1981, Discrimination in Mortgage Lending (MIT Press, Cambridge, MA). 

Sloczynski, T., 2015, The Oaxaca-Blinder unexplained component as a treatment effects estimator, 

Oxford Bulletin of Economics and Statistics 77, 588–604. 

Smith, R., and M. Delair, 1999, New evidence from lender testing: Discrimination at the pre- application 

stage, in M. Turner, and F. Skidmore, eds., Mortgage Lending Discrimination: A Review of Existing 

Evidence (Urban Institute Press, Washington, DC). 

Vickery, J. I., and J. Wright, 2013, TBA trading and liquidity in the agency MBS market, Economic 

Policy Review 19. 

34
Figure 1: Put-Backs for Issuances 2000 - 2010

Source: Goodman, Laurie S. and Jun Zhu, 2013. “Reps and Warrants: Lessons from the 

GSEs Experience”. Urban Institute: Housing Policy Center White Paper.

Presented is a copy of Table 2 from the aforementioned Urban Institute White Paper (permission 

granted in copyright.). The Figure shows the dollars put back on Freddie Mac loans by issuance 

vintage. The take-away of the figure for our purposes are twofold. First is the small value of putbacks after 2008. Second is the low volume of put-backs in the first 30 days of loan life, when 

issues remain credit-risk exposed to these loans.

35
Figure 2: An Example of the GSE Grid 

Presented is the LLPA (Loan-Level Price Adjustment) Grid of Fannie Mae for 2011. The figures is from the Fannie Mae Selling Guide, 

dated 12/23/2010. (MCMs, now retired, refers to “My Community Mortgages”, a program of subsidized loans for low-income target 

areas.) The LLPA Grid has a parallel grid at Freddie Mac called the Credit Fees in Price chart. These grids provide the additional g-fee 

(guarantee fee) that lenders must pay the GSE for guaranteeing the mortgage, varying by LTV and credit score. In practice, these lumpsum fees are translated to flows concepts to be added to the interest rate passed on to borrowers to pay for credit risk.

36
Panel A: Raw Interest Rates by Ethnicity 

Panel B: Excess Interest Rates over Month-year GSE Grid Rate by Ethnicity 

Figure 3: Interest Rate Histograms by Ethnicity: The Role of the GSE Grid 

Presented are two histograms of interest rates originated 30-year fixed-rate mortgages, 2008-2015, 

that are processed in the GSE system. Panel A shows the raw data histogram of interest rates. Panel 

B de-means the histogram to the GSE grid. The histograms are plotted for Latinx and AfricanAmericans and then for everyone else. 

0 2 4 6 8 10 12 14 16 Percent

0.02 0.04 0.06 0.08 0.10 0.12 Mortgage Interest Rate

Everyone Else

Latinx/African-American

0 5 10 15 20 Percent

-0.04 -0.02 0.00 0.02 0.04 0.06 0.08 Mortgage Interest Rate

Everyone Else

Latinx/African-American

37
Panel A: Purchase Mortgage 

Panel B: Refinance Mortgage 

Figure 4: Histogram of GSE Purchase and Refinance Mortgages across Loan-to-Value

Panels A and B present the distribution of purchase and refinance mortgages by loan-to-value 

ratios. The sample is 2008-2015, 30-year fixed mortgages, securitized through the GSE system.

38
Figure 5: Interest Rate Discrimination Estimates by Year 

Plotted are the estimates for each two-year block of time of the full fixed effects model (akin to 

Table 3, column 3) for interest rate discrimination. The years are the years of issuance. The 

estimates are divided by purchase loans (left-hand side) and refinance loans (right-hand side). The 

sample is all loans for 30-year fixed-rate mortgages, securitized through the GSE system, 2008-

2015. The estimation regresses interest rates on the GSE grid fixed effects, month-year fixed 

effects, county fixed effects, and lender fixed effects. All estimates presented, except the refinance 

results for 2014-2015, are significant with p-values well below 0.01. The estimates are converted 

to basis points (1 basis point =0.1%) for ease of conveyance. 

0 1 2 3 4 5 6 7 8

2008 to 2009 2010 to 2011 2012 to 2013 2014 to 2015 2008 to 2009 2010 to 2011 2012 to 2013 2014 to 2015

Purchase Refinance 

Graphs by loan

39
Table 1: Summary Statistics

Panel A: Rejected Applications (N = 8,648,282) 

Mean

Standard 

Deviation Minimum Median Maximum

Interest Rate % -- -- -- -- --

Loan Amount $,000 $197.9 $102.3 $30.0 $180.0 $428.0

Applicant Income $,000 $100.6 $115.9 $19.0 $80.0 $9,863.0

Credit Score (census tract) 739.0 29.1 0.0 743.5 825.0

Loan-to-Value (census tract) 0.814 0.097 0.146 0.800 2.147

FinTech 0.038 0.191

Online Application Lender 0.503 0.500

Top 25 Lender 0.670 0.470

Latinx-/African-American 0.153 0.360

Purchase=1; Refinance=0 0.737 0.440

Panel B: Accepted Applications (N = 6,865,349)

Mean

Standard 

Deviation Minimum Median Maximum

Interest Rate % 4.34% 0.63% 2.00% 4.38% 12.50%

Loan Amount $,000 $214.4 $114.8 $30.0 $191.0 $729.0

Applicant Income $,000 $108.4 $103.1 $19.0 $89.0 $9,983.0

Credit Score 754.7 38.7 630.1 759.9 850.0

Loan-to-Value 0.723 0.155 0.300 0.738 1.300

FinTech 0.043 0.204

Online Application Lender 0.507 0.500

Top 25 Lender 0.520 0.500

Latinx-/African-American 0.118 0.323

Purchase=1; Refinance=0 0.670 0.470

Presented are summary statistics for the complete dataset of HMDA applications merged with ATTOM,

McDash, and Equifax. Interest rate, LTV, and Credit Score are from the McDash-Equifax merge. Loan 

Amount , Applicant Income and Latinx-/African-American are from HMDA. Top 25 Lender are calculated

from volume of loans by lender. FinTech is a platform identifier from Buchak et al (2017). Online

Application Lender is from our manual coding in 2018 of lenders who have a full mortgage application

online. 

40
Table 2: Interest Rate Discrimination - GSE Grid Results

Panel A: Estimates

(1) (2) (3) (4)

Latinx-/African-American 0.000903*** 0.000792*** 0.000299*** 0.000356***

[0.000101] [3.09e-05] [7.96e-05] [2.96e-05]

Observations 1,480,186 1,480,186 2,068,453 2,068,271

R-squared 0.003 0.729 0.000 0.693

Month-Year FE NY NY

GSE Grid FE N Y N Y

Panel B: Economic Magnitude Calculation

Market Size of Housing Debt (Federal Reserve of New York) ($B) 12,188.7

African-American/Latinx as a Percent of Participants in Mortgage Market 0.133

Extra Interest per Basis Point of Discrimination ($M) 160.89

Estimates from Panel A: 

Extra Interest Rate (bps): Purchase Mortgages 7.92

Extra Interest Rate (bps): Refinance Mortgages 3.56

Share of refinance loans in stock of float 0.75

Weighted average extra interest rate 4.65

Aggregate Extra Interest Paid per Year by Latinx-/African-American $748.15

Dependent Variable: Mortgage Interest Rate

Purchases

Panel A reports discrimination results within the GSE Grid for purchase and refinance mortgages. The

dependent variable is the interest rate on GSE 30-year fixed mortgages originated. Columns (1) and (3)

report an estimation with just the minority ethnicity variable, equivalent to raw differences in means.

Column (2) and (4) report the main "full credit risk" model, which includes GSE grid fixed effects and

month-year fixed effects. Standard errors are clustered at the fund level. ***, **, and * indicate significance

at the 1%, 5%, and 10% conventional levels. Panel B presents economic magnitude calculations,

aggregating the findings in Panel A to the mortgage market outstanding as of the beginning of 2017. The

minority representation in the mortgage market is from our data in Table 1. It is conservative relative to the

Survey of Consumer Finance (where minorities hold 17% of debt). The aggregate housing debt if from the

Federal Reserve Bank of New York.

Refinances

41
Table 3: Interest Rate Discrimination - Robustness to Costs of Lender and Geography

Panel A: Purchases

(1) (2) (3) (4)

Latinx-/African-American 0.000792*** 0.000701*** 0.000552*** 0.000525***

[3.09e-05] [2.37e-05] [2.76e-05] [3.33e-05]

Observations 1,480,186 1,480,170 1,478,947 1,453,606

R-squared 0.729 0.733 0.747 0.758

Month-Year FE YYYY

GSE Grid FE YYYY

County FE NYYY

Lender FE NNYY

County#Lender FE NNNY

Panel B: Refinances

(1) (2) (3) (4)

Latinx-/African-American 0.000356*** 0.000365*** 0.000223*** 0.000203***

[2.96e-05] [2.84e-05] [2.21e-05] [1.79e-05]

Observations 2,068,271 2,068,262 2,067,160 2,038,761

R-squared 0.693 0.695 0.711 0.72

Month-Year FE YYYY

GSE Grid FE YYYY

County FE NYYY

Lender FE NNYY

County#Lender FE NNNY

Dependent Variable: Mortgage Interest Rate

Estimation: OLS with Varying Fixed Effects

Dependent Variable: Mortgage Interest Rate

Estimation: OLS with Varying Fixed Effects

This table mitigates the concern that our Table 2 estimates are picking up differential costs of delivering a

mortgage by lender or by geography. Panels A and B report estimates for rate discrimination for purchase and

refinance mortgages, respectively. Column (1) repeats the OLS estimate of Table 2, regressing interest rates on

the GSE grid dummy variables, month-year effects, and an indicator for whether the borrower is ethnically

Latinx or African-American. This is the main credit risk model. Column (2) adds county fixed effects; column

(3) adds lender fixed effects; and column (4) adds lender cross with county fixed effects. Standard errors in

brackets are clustered at the lender level. ***, **, and * indicate significance at the 1%, 5%, and 10%

conventional levels.

42
Table 4: Interest Rate Discimination - Robustness

Dependent Var:

(1) (2) (3) (4)

Robustness 

Concern: Positive Points Negative Points

Residual Risk via 

Servicing or MBS 

Holding

Ethnicity Designation

Sub-Sample: 0.795 < LTV 

<0.801 

LTV not near grid cell 

maximum Small Lenders

Only use HMDAClassified Ethnicity 

Observations

Reasoning: At budget 

contraint

Borrower will not face 

higher interest for 

slightly larger loan to 

cover closing costs

Unlikely to service the 

loans or hold as MBS 

on balance sheet

Eliminate software 

errors in classification

Panel A: Purchases

0.000944*** 0.000945*** 0.000868*** 0.000814***

[4.21e-05] [4.07e-05] [3.49e-05] [3.23e-05]

Observations 327,355 334,677 833,579 1,356,670

R-squared 0.739 0.725 0.730 0.729

Month-Year FE YY Y Y

GSE Grid FE Y Y Y Y

Panel B: Refinances

0.000294*** 0.000332*** 0.000385*** 0.000384***

[5.58e-05] [4.32e-05] [2.24e-05] [3.25e-05]

Observations 202,897 377,928 850,559 1,844,965

R-squared 0.73 0.704 0.732 0.694

Month-Year FE YY Y Y

GSE Grid FE Y Y Y Y

Latinx-/AfricanAmerican

Latinx-/AfricanAmerican

This table addresses a series of robustness concerns concerning interpreting the interest rates results as

discrimination. All specification use the formulation of the main credit risk model of Table 2, columns (2) and

(4), which include GSE grid dummies and month-year effects. In Panel A and B, we present results for purchase

and refinance mortgages respectively. Column (1) address the possibility that our loading on the Latinx/AfricanAmerican variable is due to positive point taking by majority ethnicities. We limit the sample to mortgages with

LTVs precisely at 0.8, where borrowers on average are more likely to be fully utilizing their cash for down

payment, leaving no cash to pay positive points. Columns (2) addresses the possibility that our results are driven

by minority borrower taking negative points (a rebate) to pay closing costs . Borrowers who have LTVs between

(0.70, 0.74) and (0.75, 0.795) are not at the edge of the GSE grid bucket in LTV. These borrowers could pay

slightly less in down payment without incurring an interest rate increase, implying taking a negative point rebate

for closing costs is likely to be suboptimal. Column (3) subsamples to only non top-25 volume lenders,

addressing robustness of our results to the concern that large lenders who retain the MBS and servicing rights

after securitizing through the GSEs. Small lenders do not service GSE loans. The concern is that servicing costs

increase with default risk. Column (4) drops borrowers who do not designate their ethnicity directly in HMDA.

Standard errors clustered at the lender level are in brackets. ***, **, and * indicate significance at the 1%, 5%,

and 10% conventional levels. mean (0.795 to 0.801).

Mortgage Interest Rate

43
Table 5: Interest Rate Discimination - FinTech Results

FinTech FinTech + 

Algorithm FinTech FinTech + 

Algorithm

(1) (2) (3) (4)

Latinx-/African-American 0.000534*** 0.000680*** 0.000200** 0.000249***

[4.36e-05] [5.50e-05] [6.34e-05] [3.08e-05]

Observations 41,832 354,680 110,870 348,833

R-squared 0.731 0.607 0.707 0.65

Year FE YY YY

GSE Grid FE Y Y Y Y

This table replicates our main credit risk specification, but only for subsamples of FinTech or algorithm-based

lenders. All columns reproduce the specification of column (2) of Table 2, regressing interest rates on the GSE

grid dummy variables, month-year effects, and an indicator for whether the borrower is ethnically Latinx or

African-American. The sample in columns (1) and (3) are the list of FinTech platforms from Buchak et al

(2017). The sample in columns (2) and (4) are the lenders who, as of summer 2018, have a full mortgage

application online, applied to lenders 2013-2015. A full mortgage application means that all pulling of data

from the applicant (tax forms, bank information, credit report, etc.) are online and the borrowing is provided an

accept/reject decision as well as a rate. The only residual underwriting that remains for the lender concerns the

house (appraisal, lien check, etc.). Columns (1) and (2) report purchase mortgage estimates, and columns (3)

and (4), refinance mortgages. Standard errors clustered at the lender level are in brackets. ***, **, and *

indicate significance at the 1%, 5%, and 10% conventional levels.

Dependent Variable: Mortgage Interest Rate

Purchase Refinance

44
Table 6: Application Rejection Discrimination 

Panel A: Purchase Mortgages

(1) (2) (3) (4) (5) (6)

Latinx/African-American 0.0509** 0.0444** 0.0316** 0.0934*** 0.0858*** 0.0639***

[0.0206] [0.0191] [0.0130] [0.00646] [0.00563] [0.00266]

Observations 4,539,812 4,539,812 4,539,812 2,007,520 2,007,520 2,007,520

R-squared 0.033 0.037 0.304 0.058 0.064 0.329

Application Variable Splines Y Y Y Y Y Y

Year FE Y Y Y Y Y Y

Census Tract Proxies: Splines of 

LTV, Credit Score, Debt -toIncome (DTI), Total DebtNYY NYY

County FE N N Y N N Y

Lender F.E N N Y N N Y

Panel B: Refinance Mortgages

(1) (2) (3) (4) (5) (6)

Latinx/African-American 0.0514*** 0.0434*** 0.0313** 0.0982*** 0.0823*** 0.0553***

[0.0177] [0.0153] [0.0140] [0.00764] [0.00638] [0.00330]

Observations 10,973,819 10,973,819 10,973,819 4,138,514 4,138,514 4,138,514

R-squared 0.027 0.03 0.172 0.066 0.078 0.305

Application Variable Splines Y Y Y Y Y Y

Year FE Y Y Y Y Y Y

Census Tract Proxies: Splines of 

LTV, Credit Score, Debt -toIncome (DTI), Total DebtNYY NYY

County FE N N Y N N Y

Lender F.E N N Y N N Y

The dependent variable is an indicator for an application being rejected by the lender. The sample in columns (1)

to (3) is the full sample, in columns (4) to (6) is FinTech platforms from Buchak et al (2017), and in columns (7)

to (9) is the set of lenders who, as of summer 2018, have a full mortgage application online. Panels A includes all

size lenders, and Panel B subsamples to non-top-25 volume lenders by year. The variables sets included are

indicated beneath each estimation, where application variables (from HMDA) include 21 splines of income and

47 splines of loan amount. Census tract variables include 25th, 50th, and 75th percentiles of LTV and credit

score by census tract from McDash. Also included in the census tract variables is the census tract mean loan

amount. Robust standard errors are in brackets. ***, **, and * indicate significance at the 1%, 5%, and 10%

conventional levels.

Dependent Variable: Application Rejection

All Issuers Small Lenders

Dependent Variable: Application Rejection

All Issuers Small Lenders

45
Table 7: Application Rejection Discrimination - FinTech

Panel A: Purchase Mortgages

(1) (2) (3) (4) (5) (6)

Latinx/African-American 0.023 0.0112 0.0259** 0.0531*** 0.0496*** 0.0344***

[0.0200] [0] [0.00881] [0.0160] [0.0136] [0.0118]

Observations 116,893 116,893 116,893 926,542 926,542 926,542

R-squared 0.082 0.088 0.301 0.039 0.046 0.345

Application Variable Splines YYY YYY

Year FE Y Y Y Y Y Y

Census Tract Proxies: Splines of 

LTV, Credit Score, Debt -toIncome (DTI), Total DebtNYY NYY

County FE N N Y N N Y

Lender F.E N N Y N N Y

Panel B: Refinance Mortgages

(1) (2) (3) (4) (5) (6)

Latinx/African-American 0.0357** 0.0243* 0.0264*** 0.0502*** 0.0471*** 0.0307***

[0.0128] [0.0108] [0.00671] [0.0112] [0.00989] [0.00738]

Observations 508,045 508,045 508,045 2,108,623 2,108,623 2,108,623

R-squared 0.072 0.084 0.143 0.046 0.051 0.193

Application Variable Splines YYY YYY

Year FE Y Y Y Y Y Y

Census Tract Proxies: Splines of 

LTV, Credit Score, Debt -toIncome (DTI), Total DebtNYY NYY

County FE N N Y N N Y

Lender F.E N N Y N N Y

FinTech FinTech + Algorithm

The dependent variable is an indicator for an application being rejected by the lender. The sample in columns (1)

to (3) is the set of stand-alone FinTech platforms from Buchak et al (2017). The sample in columns (4) to (6) is

the combination of the FinTech platforms plus the set of lenders who, as of summer 2018, have a full mortgage

application online, restricted to post 2012.The credit risk model controls consists of three potential sets of

variables, with the inclusion noted beneath the estimation. The first two comprise the full set of variables used in

the black box of the GSE underwriter system that determines GSE acceptability of applications. We do nto know

the functional form of the GSE black box. The first set includes the variables which HMDA records for both

accepts and rejects -- income and loan amount. We use 21 splines of income and 47 splines of loan amount. The

second set includes census tract proxies for the other variables known to be inputs in the GSE balck box -- LTV,

credit score, debt-to-income (DTI), and total debt. We include splines of the medians of these variables by census

tract from McDash (for LTV and credit score) and from the GSE data (for DTI). The third set of variables

includes lender and county fixed effects, as an robustness exercise econometrically. Standard errors clustered at

the lender lever are in brackets. ***, **, and * indicate significance at the 1%, 5%, and 10% conventional levels.

Dependent Variable: Application Rejection

FinTech FinTech + Algorithm

Dependent Variable: Application Rejection

46