Divorce Rates and Bankruptcy Exemption Levels in the

U.S.∗

Je�rey Traczynski†

October 6, 2010

University of Wisconsin-Madison

Department of Economics

O�ce: 7310 Social Science

O�ce Phone: (608) 263-2327

E-mail: traczynski@wisc.edu

Abstract

While raising bankruptcy exemption levels protects consumers from negative asset

shocks, an unintended consequence of this policy that remains unexplored is the e�ect on

divorce rates. This e�ect arises from reducing bene�ts of marital risk sharing. I establish

conditions under which increases in exemptions lead to more divorce and investigate

using data from the Vital Statistics of the United States. I estimate that exemption

increases between 1989 and 2005 resulted in over 200,000 additional divorces during

this period and that the e�ect on divorce rates in 2005 is comparable in magnitude to

the introduction of unilateral divorce laws from Wolfers (2006).

∗I would like to thank Karl Scholz, Jane Cooley, Chris Taber, Ben Cowen, Shannon Mok, Mai Seki,Kamil Sicinski, Caleb White, Alex Yuskavage, and participants in the UW-Madison Public Finance researchseminar for their useful comments and suggestions. All remaining errors are my own.†Department of Economics, University of Wisconsin-Madison, Madison, WI 53706; traczynski@wisc.edu.

1 Introduction

In 2007, there were approximately 1.1 million divorces and 850,000 bankruptcy �lings in the

U.S.1 Researchers have devoted signi�cant attention to the negative e�ects of divorce on the

�nancial and psychological health of both adults and children (Clark-Stewart and Brentano,

2006) and to how bankruptcy declarations under di�erent sets of laws a�ect consumer welfare

(Livshits et al., 2007). Generous bankruptcy laws can be helpful in shielding individuals from

negative income or asset price shocks, just as marriage allows individuals to share risk by

pooling resources. Since both institutions o�er some of the same bene�ts in the form of

insurance, changes in bankruptcy laws may have unforeseen consequences for divorce rates.

This tradeo� has yet to be explored in either the divorce or bankruptcy literature.

In personal bankruptcy, individuals may keep any assets that �t under predetermined

exemption levels. These exemptions allow individuals to retain some property to begin the

�fresh start� that U.S. bankruptcy law o�ers. By guaranteeing a minimum amount of assets

that debtors may keep, exemptions are an important part of bankruptcy's protections for

individuals against misfortune. The size and allowable categories of exemptions have been

central to the debate surrounding every federal bankruptcy law revision since 1978. Previous

papers have exploited variation in exemptions both across states and over time to determine

the various ways in which exemptions a�ect consumers' ability to obtain loans and the

interest rates that banks o�er (White, 1987; Gropp et al., 1997; Lin and White, 2001).

This paper quanti�es the speci�c e�ect of bankruptcy exemption levels on divorce rates.

I use a simple model of divorce in which individuals only obtain a divorce if there is no

possible outcome from bargaining within marriage that makes both partners better o�. This

model suggests a way to test whether variation in bankruptcy exemption levels a�ects state

divorce rates. Intuitively, bankruptcy o�ers more insurance against negative income and

1Total number of divorces from author's calculations. Divorce calculation includes the assumption thatstates that do not report total divorces have a divorce rate equal to the national rate. See Section 4 andTable 2 for more information on divorce data availability. The total number of bankruptcies was obtainedfrom http://www.uscourts.gov/Press_Releases/2009/BankruptcyFilingsDec2008.cfm.

1

asset shocks when exemption levels increase, so I use exemption levels as a measure of the

insurance o�ered by bankruptcy.

I investigate empirically whether increases in exemption levels lead to increases in the

divorce rate using state level panel data from the Vital Statistics of the United States, which

covers the vast majority of divorces in the U.S. over the period 1989-2005, along with data on

state bankruptcy exemption levels. The empirical analysis exploits variation within states

over time. I �nd evidence that increases in the total exemption amount available to debtors

are associated with increases in the divorce rate. I estimate that over the sample period,

increases in exemption levels resulted in over 200,000 additional divorces and that, by 2005,

the magnitude of the increase in divorce rates is comparable to the short run e�ect of

introducing unilateral divorce as estimated by Wolfers (2006). Overall, the e�ect of changes

in bankruptcy exemption levels on the divorce rate is a large unintended consequence.

2 Previous Literature and Background

2.1 Divorce

Various studies attempt to pin down economic factors in�uencing the divorce rate. Con-

ger et al. (1990) demonstrate that couples with self-reported economic hardships are more

likely to divorce, while Sweezy and Tiefenthaler (1996) show that a number of state level

variables, such as mandatory separation requirements before divorce and AFDC and food

stamp payments, have no signi�cant e�ect on the divorce rate. Neither paper focuses on the

role of social insurance in marital dissolution: this paper is the �rst to examine the e�ect of

changes in bankruptcy exemption levels on divorce rates.

Much of the debate on determinants of divorce rates has centered on the e�ects of uni-

lateral divorce laws. Becker (1981) uses the Coase Theorem to argue that unilateral divorce

should have no e�ect on divorce rates, while Clark (1999) presents a model in which intro-

ducing unilateral divorce may have a positive, negative, or no e�ect on divorce. Both models

2

analyze how resource allocation within marriage and property division in divorce change in

response to changes in divorce laws, altering an individual's �nancial position and optimal

divorce decision. In empirical work, Peters (1986, 1992) �nds that the introduction of uni-

lateral divorce in a state does not a�ect the probability that a woman in that state gets

divorced, while Allen (1992) and Gruber (2004) �nd the opposite. Using the same state level

data employed in this paper, Friedberg (1998, p. 626) claims that the widespread adoption of

unilateral divorce laws between 1968 and 1988 accounts for 17% of the overall increase in the

divorce rate during this period. However, Wolfers (2006, p. 1814) revisits Friedberg's results

and �nds that �unilateral divorce laws explain only a very small fraction of the dramatic rise

in divorce.�

Previous work analyzes the correlation between being divorced and declaring bankruptcy.

In a 1991 survey described by Sullivan et al. (2000), 22.1% of bankruptcy �lers reported

divorce as a major cause of their decision to �le, and a 1999 survey discussed in Jacoby

et al. (2000) ranks divorce as the third leading cause of bankruptcy �lings behind job loss

and medical expenses. Since a single person is more susceptible to income shocks from

job loss and unforeseen medical bills than a married person, these surveys may understate

the full e�ect of divorce on bankruptcy decisions. Nonetheless, they establish the general

pattern that a single person has a higher probability of declaring bankruptcy than one who

remains married. Domowitz and Sartain (1999, p. 413) estimate that a single individual is

approximately twice as likely as a married individual to �le for bankruptcy, while Fay et al.

(2002) shows that individuals are more likely to �le for bankruptcy if they become divorced

in the previous year and Edmiston (2006, p. 78) claims that a one percentage point higher

share of the population being divorced would lead to 7.8 additional bankruptcies per 1000

households each year. These papers focus on the choice to declare bankruptcy after divorce,

while this study uses changes in the asset protection o�ered by bankruptcy law to examine

the decision to divorce.

The recent decline of U.S. home prices provides an example of an asset shock that married

3

couples can withstand better than singles. Stories about couples who �can't leave their

current situation because the �nancial costs are too great and because it's too di�cult to

sell their house� (Romans, 2009) and end up �reluctantly staying together until the housing

market turns around� (Armour, 2009, p. 2B) have appeared in national media such as

CNN and USA Today.2 These stories o�er anecdotal evidence of couples facing �nancial

uncertainty choosing not to divorce. In this paper, I show how changes in bankruptcy

exemption levels can a�ect the risk sharing motive for marriage. Since di�erent divorce laws

provide di�erent rules for property division, I test whether changes in bankruptcy exemptions

have di�erent e�ects on divorce across legal regimes.

2.2 Bankruptcy

In the past thirty years, the United States has had three major reforms to its bankruptcy

laws: the Bankruptcy Reform Act of 1978 (BRA78), the Bankruptcy Reform Act of 1994

(BRA94), and the Bankruptcy Abuse Prevention and Consumer Protection Act of 2005

(BAPCPA). BRA78 was a comprehensive reform that replaced the Bankruptcy Act of 1898.

Unlike all previous bankruptcy law changes, BRA78 was not a response to an economic

recession, but rather designed to �meet the demands of present technical, �nancial, and

commercial activities� (Tabb, 1995, p. 33). BRA78 established a uniform national set of

exemptions while allowing states to opt out and set their own exemption levels if desired.

Every state set their own exemptions by 1987, though several allowed debtors the option of

using the federal exemptions.3 White (1987) shows that higher state exemption levels are

associated with a high number of bankruptcy �lings.

Many di�erent categories of exemptions are available to debtors in bankruptcy. The

largest in dollar terms is the homestead exemption, which allows debtors to prevent cred-

itors from seizing part of their housing equity during bankruptcy. Smaller exemptions are

2See also Lloyd (2008) in the San Francisco Chronicle and Leland (2008) in the New York Times.3Posner (1997) o�ers an excellent discussion of the political background leading up to the passage of

BRA78.

4

also available for a wide variety of personal property, ranging from cars and furniture to

ceremonial guns and wedding rings. Some states list maximum exemption amounts for each

type of good separately, while others o�er �wildcard� exemptions that allow debtors to retain

any personal property of their choice up to a speci�ed dollar amount.

Both BRA94 and BAPCPA were intended to alter the relationship between bankruptcy

and divorce settlements; in fact, these alterations were a goal of BRA94.4 Under BRA78,

a debtor's obligations to a former spouse were divided into two categories: support debts

and property division debts. Support debts include alimony, maintenance payments, and

child support payments, while property division debts result from the decision of a judge to

divide the equity that the married couple held in a large item, usually the marital home.

In a typical divorce, a judge awards the home to one partner and orders that the other

spouse be paid half the value of the home, so whomever gets the house also receives a large

debt. In bankruptcy, property division debts were considered dischargeable, while support

debts were not. Since bankruptcy courts applied this distinction strictly when determining

dischargeability, a former spouse unhappy with the divorce settlement had the opportunity

to use bankruptcy courts to try to obtain a more favorable division of property and debts.5

Such litigation was su�ciently prevalent and, occasionally, high-pro�le that judges be-

gan to push Congress for bankruptcy reform.6 One case, Farrey v. Sanderfoot, reached

the Supreme Court in 1991 and prompted a dissenting lower court judge to remark that

the debtor was using bankruptcy laws �to steal from his former wife.�7 BRA94 instructed

bankruptcy courts not to discharge any debts resulting from a divorce settlement that �would

result in a bene�t to the debtor that outweighs the detrimental consequences to a spouse,

4Johnson (1997) contains a detailed discussion of changes under BRA94 and their motivations.5Alexander (1994, p. 360): �The advantage for the debtor arises because Congress clearly intended, in

drafting the Code, to provide debtors with a fresh �nancial start.�6Bello (1993, p. 645). Bello (1993) also provides contemporary analysis of the legal environment sur-

rounding marital support cases.7Farrey v. Sanderfoot (In re Sanderfoot), 899 F. 2d 598, 607 (7th Cir. 1990) (Posner, J., dissenting),

revised, 500 U.S. 291 (1991). This case was almost certainly the most publicized of all divorce cases involvingbankruptcy - see Margolick (1991) for a contemporary newspaper article.

5

former spouse, or child of the debtor.�8 BRA94 also roughly doubled the previous federal

exemption levels and prompted a number of states that had opted out of the federal statute

to increase their own exemption levels in response. BAPCPA declared that under Chapter

7 bankruptcy, any debt �incurred by the debtor in the course of a divorce or separation

or in connection with a separation agreement, divorce decree or other order of a court of

record� would not be considered dischargeable and limited a debtor's ability to engage in pre-

bankruptcy planning by empowering judges to reverse any asset transfers between exemption

categories made shortly before the bankruptcy �ling.9

3 Model

I develop a simple one-period framework, building on the work of Kotliko� and Spivak (1981),

to describe the e�ects of changes in bankruptcy exemption levels on the utility of being

divorced relative to remaining married.10 All individuals begin married.11 Utility functions

are strictly increasing and strictly concave in both consumption, ci, and match quality with

spouse θi, i = (h,w). Match quality is a random variable with known distribution. At

the beginning of the period, each partner receives a realization of match quality and decides

whether to declare divorce. After making this decision, individuals learn Yi, the present value

of their assets and income in that period. A household consists of either two married partners

or one divorced individual. All married households carry debt with present value 2T and may

declare bankruptcy at no cost, keeping any assets below exemption level βE, where E is the

exemption level available to a single individual and 1 ≤ β ≤ 2 is a multiplier representing

the fact that the exemptions available to married households are higher in many states.

Following the empirical results of Gropp et al. (1997) and Lin and White (2001), I assume

811 U.S.C. �523(a)(15)(A), 1994.911 U.S.C. �523(a)(15)(5), 2005.10There are also many bargaining models of marriage, with the Nash bargaining models of Manser and

Brown (1980) and McElroy and Horney (1981) being among the most prominent. Using a Nash bargainingmodel does not change the intuition of the results presented here.

11This assumption ignores the potential e�ect of exemption levels on the probability of getting married. Iaddress this issue in Section 6 below.

6

that the amount of debt that a household can carry is a function of the state exemption

level, so T = T (βE). For simplicity, I assume that there are only two possible asset levels,

so Yi ∈{Y L, Y H

}, where βE < Y L < Y H . I further assume that a divorced household will

declare bankruptcy if it receives a low asset value while a married household will only declare

bankruptcy if both spouses receive low asset values, so 2Y L − 2T < βE < Y L + Y H − 2T . I

denote the probability of receiving each asset value by pj, j = (H,L).

Within marriage, I assume that asset transfers are costless. This implies that the couple

will always consume on the Pareto frontier of possible utility values for each partner, with the

exact point determined by relative bargaining power of the spouses within the marriage. This

also means that divorce will only occur when there is no point on the frontier that o�ers both

partners a utility level higher than they would receive in divorce. Thus, a married household

solves

maxch,cw

Uh (ch, θh) + αUw (cw, θw) s.t. ch + cw ≤ Yh + Yw − 2T

if it does not declare bankruptcy and

maxch,cw

Uh (ch, θh) + αUw (cw, θw) s.t. ch + cw ≤ βE

otherwise, where α > 0 represents the relative bargaining power of the two married indi-

viduals. Let(cNBh , cNBw

)denote the solution if the couple does not declare bankruptcy and(

cBh , cBw

)denote the solution if they do.

If either spouse chooses divorce, then each former partner solves the trivial problem

maxci

Ui (ci, 1) s.t. ci ≤ Yi − T

if the individual does not declare bankruptcy and

maxci

Ui (ci, 1) s.t. ci ≤ E

7

otherwise.12 Here, it is clearly optimal for the individual to consume Yi − T if not declaring

bankruptcy and E otherwise.13

Now consider the e�ect of an increase in E, the bankruptcy exemption level, on the utility

values of divorce and marriage. For i, the value of being married with a given match quality

θi is given by

mi = p2L · Ui

(cBi , θi

)+

(1− p2

L

)· Ui

(cNBi , θi

)(1)

while the value of being divorced is given by

di = pL · Ui (E, 1) + (1− pL) · Ui (Yi − T, 1) . (2)

E appears in both explicitly in di and implicitly in the value of marriage through its e�ect

on(cBh , c

Bw

). For concreteness, I assume that each individual's utility function is a standard

Cobb-Douglas function of the form

Ui (ci, θi) = cγii θ1−γii

with 0 < γi < 1. Under this assumption, taking the derivative of Equations 1 and 2 with

respect to E yields

∂mi

∂E= p2

L ·(γi

(cBi

)γi−1θ1−γii

)· β · ∂c

Bi

∂E(3)

and

∂di∂E

= pL ·(γiE

γi−1)− (1− pL) ·

(γi(Yi − T )γi−1

)· β · ∂T

∂E, (4)

which we can compare to determine how changes in bankruptcy exemption levels a�ect

divorce rates. Speci�cally, if ∂di∂E

> ∂mi∂E

, then the value of divorce increases relative to the

value of marriage whenever E increases. With less joint surplus available for a married

12I normalize the distribution of the match quality parameter such that θi > 0 and for single individuals,θi is set to 1.

13I ignore here any issues raised by property division laws by assuming that each individual keeps onlyhis or her own assets after getting divorced and that all debts of the married households are joint. I addressthis concern in Section 6 below.

8

couple to divide, there are now a greater range of θi values that will cause the partners to

prefer divorce. In the context discussed above, a low asset value realization can be thought

of as a fall in home prices and the risk-sharing motive for remaining married is rendered less

important by the higher guaranteed level of consumption o�ered by the higher exemption

level. Determining which derivative is larger and whether marriages become more fragile

when bankruptcy exemption levels increase is thus an empirical question.

Comparing Equations 3 and 4, ∂di∂E

> ∂mi∂E

implies

pL ·(γiE

γi−1)− (1− pL) ·

(γi(Yi − T )γi−1

)· β · ∂T

∂E> p2

L ·(γi

(cBi

)γi−1θ1−γii

)· β · ∂c

Bi

∂E

or (1

E

)1−γi> pL · β ·

∂cBi∂E

(θicBi

)1−γi+

((1− pL) · β

pL · (Yi − T )1−γi

)· ∂T∂E

. (5)

From previous empirical estimates, I assume that ∂T∂E

< 0.14 It can be shown that 0 <∂cBi∂E

<

1.15 In Equation 5, all three terms grow smaller when E increases, taking into account

changes in the equilibrium values of cBi , T ,∂cBi∂E

, and ∂T∂E. Equation 5 is also consistent with

the e�ects of the housing market decline under the interpretation of pL as the probability of

getting a low house price. As pL increases, it is more likely that ∂di∂E

< ∂mi∂E

, so couples are

more likely to stay together as E increases.

If the inequality in Equation 5 is satis�ed, then the value of divorce increases relative to

the value of staying married in response to an increase in exemption levels. This inequality

can be tested in the data by determining if divorce rates increase when exemption levels rise.

This simple model captures the most salient elements of the interaction between uncertain

asset values, exemptions, and the divorce decision.

14See Lin and White (2001, p. 155) and Gropp et al. (1997, p. 220) for empirical evidence supporting thisassumption.

15See Appendix for a discussion of the properties of∂cB

i

∂E .

9

4 Empirical Model and Data

The reduced-form model I estimate is given by

divrateit = α + δ · TotalExemptionit + κ · (TotalExemptionit)2

+ π ·Xit + µt + ηi + ζi · Trend+ ξi · Trend2 + εit

(6)

Here, divrate is the number of divorces in state i during year t per 1000 married persons.

TotalExemption is the sum of the real homestead and nonhome exemptions available to mar-

ried couples under the assumption that doubling of exemptions is allowed unless speci�cally

prohibited measured in units of $10,000 and adjusted to constant 2007 dollars using CPI

data from the Bureau of Labor Statistics. Xit represents other control variables discussed

below. µt is a year �xed e�ect and ηi is a state �xed e�ect, while ζi · Trend and ξi · Trend2

are state-speci�c linear and quadratic time trends. This regression form is similar to those

used by Friedberg (1998) and Wolfers (2006) to analyze the impact of unilateral divorce laws

on the divorce rate and by Berkowitz and Hynes (1999) to study the e�ect of bankruptcy

exemptions on mortgage rates.

Under this empirical model, identi�cation of the e�ect of the bankruptcy exemption

levels comes from variation within individual states over time. If there were no changes

in bankruptcy exemption levels, the vector of exemption levels would be perfectly collinear

with the state �xed e�ects. Thus, the state �xed e�ects control for di�erences between

states in exemption levels, while the exemption variables control for the e�ects of changes

within states in exemption levels. The state �xed e�ects also control for any state speci�c

characteristics that may a�ect divorce rates and are constant over time, including separation

and residency requirements, while year �xed e�ects capture any unobserved national changes

in divorce propensity.16

16To check the appropriateness of using this estimation strategy for this analysis, I also estimate a versionof Equation 6 including leads and lags of bankruptcy exemption levels to determine if state divorce ratesrespond to future exemption levels. I �nd that only past and contemporaneous exemption levels signi�cantlya�ect the divorce rate, which I interpret as evidence in favor of this research design. Results are available

10

I choose the sample period, 1989 to 2005, to avoid complications created by changes

in bankruptcy laws that raise potential issues of policy endogeneity or alter the nature of

bankruptcy. Hynes et al. (2004) claim that a state's decision to opt out of the federal exemp-

tion levels may be correlated with a number of other variables such as the state bankruptcy

�ling rate and the generosity of the state's assistance programs for the poor, which is po-

tentially problematic for this study given the relationship between income transfers to the

poor and the willingness of individuals to leave a risk sharing arrangement such as marriage.

I begin my sample in 1989 since no state opts out after this time. Similarly, if there are

substantial changes in how debtors may use exemptions in the bankruptcy process, then the

level of insurance o�ered by exemptions is not well captured by their dollar amount. The

2005 BAPCPA reform made pre-bankruptcy planning more di�cult, so consumers are less

able to move assets between exemption categories after this time. Since this change makes

the assumption of fungibility of assets less viable, I end the sample for analysis in 2005.17

Another potential concern is that divorce rates and bankruptcy exemption levels may be

jointly determined by a third factor, such as the rate of bankruptcy �lings in the state, or that

policymakers change exemption levels in response to changes in the state divorce rate. Hynes

et al. (2004, p. 31) examine potential determinants of state exemption levels from 1975-96,

including the state divorce rate and bankruptcy �ling rate, and �nd that the �only robust

predictor of exemption levels ... was historic levels of exemptions.�18 Fay et al. (2002, p.

709) argue that exemption levels should be treated as exogenous with respect to bankruptcy

�ling rates because �states change their exemption levels only rarely - mainly to correct

nominal exemption levels for in�ation,� and this intuition also applies for why exemption

levels can be considered exogenous with respect to divorce rates. Over 1989-2005, there were

upon request.17See Edmiston (2006, pp. 60-61) for a summary of relevant changes brought about by BAPCPA. This

sample period does contain BRA94, which altered the types of debts that result from divorce that may bedischarged in bankruptcy. As this was a national law change, much of the e�ect of the reform will be capturedby year �xed e�ects included in the regression. I examine one potential source of cross-state variation in thee�ect of this provision of BRA94 on state bankruptcy laws in Section 6 below.

18I con�rm in my sample that the state divorce rate, including lagged values, is not a statistically signi�cantpredictor of the state's exemption level. Results are available upon request.

11

a total of 112 changes in the nominal total value of state exemption levels, an average of 6.6

changes per year. Since it does not appear that exemption levels are determined directly

by the divorce rate nor through a third factor such as bankruptcy �lings per capita, I treat

exemption levels as exogenous with respect to the divorce rate.

I obtained data on state bankruptcy exemption levels from state statutes for the years

1989-2005. The homestead exemption is simply the amount listed in each year, which is

unlimited in some states. I ignore any conditions on lot size or location for the homestead.

I construct the total nonhome exemption level by adding together all allowable exemptions

for cars, personal possessions, tools of trade, bank deposits, and wildcard exemptions. I omit

any explicit exemption amounts for clothing or household goods, as well as any insurance,

burial plot, or pension exemptions. These items are not subject to value limits in many

states or have speci�c bene�ciary requirements and therefore do not have easily quanti�able

changes in value over time. This formulation of the nonhome exemption is comparable to

that used by Gropp et al. (1997) and Berkowitz and Hynes (1999). Table 1 lists both the

homestead and nonhome exemption levels for singles in nominal terms for all states in 1989,

1995, and 2005. These years mark the beginning and end of the sample as well as the year

immediately following the passage of BRA94. In most states in this sample the homestead

and nonhome exemption levels rise over time, both in nominal and real terms.

Since the model presented above relies on the in�uence of bankruptcy exemptions on

divorce values, I calculate the available exemptions in three di�erent ways. Many states

have laws explicitly allowing married couples to exempt a larger amount than a single person,

while others have expressly disallowed married couples to double the available exemptions.

In other cases, there is no speci�c court ruling determining whether a couple may double an

exemption. I therefore calculate the exemption level in a state under the assumption that

all exemptions may be doubled unless explicitly disallowed.19 Table 1 lists whether a state

19As a sensitivity check, I also calculate exemption levels under the assumption that only exemptionsexplicitly allowed may be doubled and exemptions levels as they apply to single individuals, ignoring anypossible doubling. In the context of the theoretical model, these alternative calculations allow for changesin β. The results below are qualitatively unchanged under these alternative measures of exemption levels,

12

allowed married couples a larger homestead exemption than singles in 2005.

I then create the total exemption level by combining the total amounts of the homestead

and nonhome exemptions under the assumption that assets are generally fungible across the

two categories. Many states, including Georgia, Nebraska, New Mexico, New York, North

Dakota, Vermont, and Virginia, o�er substantial cash or wildcard exemptions that can be

used in place of a homestead exemption, or follow the federal exemptions in allowing a

debtor to claim any unused portion of the homestead exemption as a wildcard, possibly up

to some predetermined dollar limit.20 Even in states without such explicit rules, Bello (1993,

p. 655) supports the legal validity of the assumption that exemptions may be combined,

claiming that �pre-petition planning that enables a debtor to convert property into the type

of property that will be exempt . . . is generally allowed.�

I obtained data on the number of divorces within a state from the Vital Statistics of

the United States and the percentage of the population over age 15 currently married from

March Current Population Survey supplements for years 1989-2005 for all �fty states and the

District of Columbia. While using individual level data that contains divorce information

might be preferable due to the availability of other characteristics thought to be correlated

with divorce rates such as education, age at marriage, or presence of children, I use state level

data to avoid potential problems including the possibility that getting divorced is correlated

with dropping out of the panel and the small sample sizes in most available panel data sets.

Figure 1 shows the federal divorce rate during this period.

Due to irregularities in state reports, not all states have an o�cial divorce rate in all

years. Table 2 lists the average divorce rates per 1000 married persons for each state from

1989-2005. I also list the years for which divorce data are available for each state.21 In

and in no case is the point estimate of the e�ect of exemption levels outside the 95% con�dence intervalof any other point estimate. The results in Table 3 use the highest e�ective value of β and are the mostconservative estimates of the e�ect of changes in bankruptcy exemption levels on divorce rates.

20See Elias et al. (2005) for an exhaustive list of state exemptions and the conditions under which theymay be combined.

21If data do not exist for a state for some years within an interval, I report the average divorce rate of theyears for which data exist.

13

general, the divorce rate is falling over this sample period in the vast majority of states, as

is the national average.

Several states have unlimited household exemptions. Since the majority of the states

with unlimited household exemptions are adjacent to the Mississippi River, I present results

for samples using only states with de�ned homestead exemptions and the full sample of all

states. I do this to control for possible policy endogeneity, as individuals have a limited ability

to select more favorable divorce or bankruptcy laws by satisfying residency requirements.22

Since it is easier to satisfy such requirements in neighboring states, the choice of each state

to continue to o�er an unlimited exemption may be related to the decision of neighboring

states. I assign a nominal value of $500,000 for the homestead exemption in states with

unlimited exemptions, following the precedent established by Berkowitz and Hynes (1999).23

In states that allow their residents to choose between using state exemptions and federal

exemptions, I assume that individuals choose the federal exemptions if the total value of the

federal exemptions (homestead plus nonhome) exceeds that of the state exemptions.

To control for contemporaneous economic conditions in each state, I include real personal

income as obtained from the Bureau of Economic Analysis, along with the state unemploy-

ment rate from the BLS and the real median state house price from the Census Housing

Tables and Federal Housing Financial Agency's all-transactions index.24 I also include the

state homeownership rate, reported by the Census Bureau, as a measure of economic wealth

and because exemption levels are generally higher for homeowners. I then add each state's

population percent black, percent Hispanic, and percent between ages 15 and 64 as demo-

22For divorce laws, Nevada's very low residency requirements and high divorce rate suggests that indi-viduals do exhibit some selection across states. The main results presented are robust to the exclusion ofNevada. For bankruptcy laws, see Topolnicki and Macdonald (1993) for examples of individuals attemptingto use di�erent bankruptcy laws to skirt repaying creditors.

23Berkowitz and Hynes (1999) try several di�erent values for the homestead exemption in unlimited states,including $1,000,000, $750,000, $500,000, and $250,000. Using these alternative values does not substantivelychange the results presented here. I present results using a dummy variable for unlimited exemption statesin Section 6. Also note that when determining the exemption level available to married couple, I do notdouble the nominal amount in unlimited homestead states.

24The BEA de�nes personal income as �income received by persons from participation in production, fromgovernment and business transfer payments, and from government interest.� (Ruser et al., 2004, p. 1)

14

graphic controls. To capture state laws that may a�ect divorce rates, I add dummy variables

for whether a state has child custody guidelines, whether judges are permitted to consider

marital fault when determining property division or maintenance payments, and whether a

state allows covenant marriage. Each of these variables is taken from the year-end review

of family law changes in Family Law Quarterly and has been considered as a possible de-

terminant of divorce rates in the literature.25 I include other measures of each state's social

insurance programs designed to assist individuals with low income. I use the real maximum

AFDC/TANF payment as reported in the Green Book of the U.S. House Committee on Ways

and Means and real maximum earned income tax credit calculated from NBER TAXSIM

and the Brookings Institution Tax Policy Center. All control variables measured in dollars

are adjusted for in�ation to constant 2007 dollars using the CPI.

5 Empirical Results

Table 3 reports the estimation results from Equation 6 using the exemption levels avail-

able to married couples assuming that all exemptions available to singles may be doubled

unless speci�cally disallowed. Speci�cations (1)-(4) use a sample of states with de�ned

homestead exemptions while (5)-(8) are estimated over the full panel of states. All esti-

mates are performed using weighted least squares, where each observation is weighted by

the contemporaneous married population of the state. The weights ensure that the changes

in bankruptcy exemptions that a�ect the most people are given the most importance in the

estimation. Since changes in state bankruptcy exemption levels can be prompted by changes

at the federal level, estimates from the weighted speci�cation may be more relevant to fed-

eral policymakers. In both the de�ned homestead and full samples, the coe�cients on the

bankruptcy exemptions decline in absolute value as additional controls are added, but no

25For more information on the laws, see Elrod and Spector (2006) and previous years. Nixon (1997) exploresthe link between divorce and child support payments, which are determined in part by custody decisions.Brinig and Buckley (1998) look at the e�ect of no-fault settlements on divorce rates, while Matouschek andRasul (2008) discuss covenant marriages as part of a marriage contract.

15

set of estimates of the exemption coe�cients is outside the con�dence interval of any other

set of estimates. Thus, the estimates are robust to the inclusion of a variety of additional

control variables. In all speci�cations, the coe�cient on the total exemption value is positive

and the coe�cient on its square is negative, indicating that increases in exemption levels

lead to increases in the divorce rate with a diminishing marginal e�ect.

Table 4 provides context for the magnitude of these estimates. Weighting each state's

exemption level by its married population, the mean total exemption level for states with a

de�ned homestead exemption was $53,949.60 in 1989 and $79,993.07 in 2005, while the mean

for all states was $106,632.90 in 1989 and $160,014.10 in 2005 when unlimited exemptions

are set to $500,000. Using the estimates from column (4) of Table 3, these results indicate

that if a state with a de�ned homestead exemption increased its total exemption level from

$53,949.60 to $79,993.07 between 1989 and 2005, this would lead to an increase of 0.093 in

the 2005 divorce rate. Since the national divorce rate in 2005 was 6.38 divorces per 1000

married persons, the mean increase in bankruptcy exemption levels is associated with a 1.46%

increase in the mean divorce rate. Turning attention to the full sample of states and using

the coe�cients from column (8) of Table 3, I estimate that increasing exemption levels from

$106,632.90 to $160,014.10 between 1989 and 2005 leads to an increase of 0.151 in the 2005

divorce rate, a 2.37% increase. This e�ect size is approximately the same as Wolfers (2006,

p. 1814) attributes to unilateral divorce laws a few years after their introduction. While the

e�ect of the introduction of unilateral divorce laws on the divorce rate has decreased over

time, continual increases in bankruptcy exemption levels make the estimates of the e�ect size

larger as time passes. Over the sample period, I estimate that more than 200,000 additional

divorces have occurred in the U.S. as an unintended consequence of increases in bankruptcy

exemptions. As I have used the speci�cations with the smallest estimated coe�cients, this

should be interpreted as a conservative estimate. Overall, the results presented in Table 3

and subsequent analysis show that increases in bankruptcy exemption levels over the sample

period led to increases in the divorce rate and a sizable number of additional divorces.

16

6 Sensitivity Analysis

As noted above, the model o�ers other testable implications and ignores several potentially

important aspects of the divorce decision, including the in�uence of di�erent property divi-

sion laws and the e�ect that exemptions have on the initial choice to marry. These omitted

features may bias the estimates of the impact of exemption levels. To investigate these

alternatives, I use the general regression form

divrateit = α + δ · TotalExemptionit + κ · (TotalExemptionit)2

+ ψ · TotalExemptionit ·Dummyi + φ · (TotalExemptionit)2 ·Dummyi

+ π ·Xit + µt + νt ·Dummyi + ηi + ζi · Trend+ ξi · Trend2 + εit

(7)

where Dummyi is a dummy variable equal to 1 if state i has a particular characteristic

of interest. Speci�cally, I use dummies for states with community property laws, unlim-

ited homestead exemptions, and unilateral divorce laws. Under this speci�cation, ψ and

φ estimate the di�erence between the e�ect of a given variables in states with a speci�c

characteristic and states without. I determine whether these di�erences are signi�cant using

F-tests on the hypothesis that the bankruptcy interaction coe�cients ψ and φ are jointly 0.

I report all results for all sensitivity checks in Table 5.

6.1 Community Property Laws

In the context of bankruptcy, one particularly important aspect of a divorce is the division

of marital assets. In a state with community property laws, all assets and debts acquired

during a marriage are considered to be equally the property of both spouses. In addition,

most property that individuals owned prior to the marriage is considered community property

after a speci�c period. Upon divorce, community property states are more likely to award

spouses equal shares of marital assets than other states. This a�ects the value of divorce

for both partners directly by changing their income levels and indirectly by making income

17

levels more or less certain. If a judge has great latitude in dividing the equity in the marital

home, then both spouses face additional risk in declaring divorce as they are less able to

predict their asset levels after divorce.

Community property laws have a theoretically ambiguous e�ect on the divorce rate, as

changing the division of marital assets will increase the utility of divorce for one partner while

lowering it for the other relative to another legal regime. However, the division of property

after divorce is more predictable, as judges have less discretion when deciding settlements.

As in the theoretical model, the newly single individuals still face �uctuations in asset values

after divorce. With a more accurate estimate of their asset levels after divorce, individuals can

better assess how much insurance is o�ered by exemption levels. For example, an individual

who knows he will receive so many assets that declaring bankruptcy is a remote possibility

will not be o�ered much insurance by exemption levels, while someone who knows he will

receive few assets may �nd exemption levels very relevant to his divorce decision. In this

sense, exemption levels are a cleaner measure of available social insurance when uncertainty

has been removed from marital asset division.

Community property states also provide an opportunity to evaluate the importance of the

divorce debt reforms included in BRA94. As discussed above, BRA94 restricted the ability

of debtors to use bankruptcy to discharge debts acquired in divorce settlements. Since the

division of debts between spouses is di�erent in community property states than in others,

this provision of BRA94 may have a di�erent e�ect in community property states. If so,

there should be a signi�cant coe�cient on the interaction between the year dummy and the

community property dummy in the years after BRA94.

Table 1 lists the states with community property laws. I de�ne CPi to be 1 if a state

has community property laws and present results in columns (1) and (2) of Table 5. In

states with de�ned exemption levels, the hypothesis that the coe�cients on the interactions

are jointly zero is rejected and the marginal e�ects of changes in bankruptcy exemptions on

divorce rates are larger in community property states than states without such laws. In the

18

sample of all states, I �nd no di�erential e�ect in community property states, and the point

estimates on the interaction variables falls by roughly half. Across both regressions, of all

the interactions between year dummies and the community property dummy, only the 1992

interaction is statistically signi�cant.

As there appears to be no di�erential e�ect in the year dummies after passage of BRA94,

I interpret this as evidence that the divorce debt provisions of this law did not meaningfully

alter the relationship between divorce rates and bankruptcy exemptions levels. The signi�-

cance of the interaction between community property laws and bankruptcy exemption levels

shows that individuals do �nd exemption levels relevant after an even division of all marital

assets, so the results in this paper are not driven by the relevance of state exemption levels to

divorce settlements that leave one spouse with a very small asset level. I interpret the lack of

statistical signi�cance and smaller estimated interaction e�ects when including the unlimited

exemption states as evidence that in unlimited exemption states, the additional insurance

provided by the equality of distribution is small compared to the large size of the available

exemptions. For de�ned exemption states, the larger estimated e�ect of exemption levels on

divorce rates in community property states is evidence that exemption levels a�ect divorce

rates by changing the amount of available social insurance, a substitute for the insurance

o�ered by marriage.

6.2 Unlimited Homestead Exemptions

In the main results, states with unlimited homestead exemptions were assigned a nominal

value for the homestead exemption, and changing this value does not qualitatively alter the

main results. However, the main results also suggest that the marginal e�ect of changes in

exemption levels on divorce rates falls as the exemption level rises, and this result may be

sensitive to the nominal value assigned to unlimited homestead states. To test whether the

marginal e�ects of increases in exemption levels are di�erent in states with unlimited home-

stead exemption than in states with de�ned homestead exemptions, I de�ne UnlimHomei

19

to be 1 if a state has an unlimited homestead exemption and present results in column (3) of

Table 5. This approach also has the bene�t of producing estimates of the e�ect of increases

in exemption levels on divorce rates that do not require a speci�c choice of the nominal

exemption value assigned to unlimited homestead states.

With the additional controls included, the coe�cients on the interacted exemption vari-

ables are not signi�cant. This indicates that changes in bankruptcy exemption levels do not

have a signi�cantly di�erent e�ect on the divorce rate in states with unlimited homestead

exemptions. This result also provides evidence that the estimated diminishing marginal ef-

fect of exemption level increases on the divorce rate is not sensitive to the nominal value

assigned to unlimited homestead exemptions.

6.3 Unilateral Divorce

The model in Section 3 assumes that the decision to divorce can be made unilaterally by

either spouse, a potentially restrictive feature of the model. To investigate whether the

presence of unilateral divorce laws alters the extent to which bankruptcy exemption levels

a�ect divorce rates, I use Unilati, a dummy equal to 1 if divorce in a state �requires the

consent of only one spouse and is granted on no-fault grounds� based on the coding in

Friedberg (1998, p. 613).26 Columns (4) and (5) of Table 5 contain the results. Under both

speci�cations, I reject the hypothesis that the marginal e�ect of an increase in exemption

levels is di�erent in unilateral divorce states. This result shows that the estimated e�ect of

bankruptcy exemption levels on divorce rates is not a�ected by a prominent characteristic

of state divorce laws.

26While there is some debate as to how to properly de�ne �unilateral� divorce, Wolfers (2006) shows thatestimates of the e�ect of unilateral divorce laws on the divorce rate are similar across seven di�erent codingschemes.

20

6.4 Marriage Rates

One hypothesis not captured by the model is that if individuals are su�ciently forward

looking to consider the possibility of declaring bankruptcy after divorce, these same individ-

uals would likely consider this possibility before entering marriage. This yields a potential

alternative explanation for the results presented here: if increases in exemption levels are

associated with increases in the marriage rate, then these marginal marriages might be espe-

cially vulnerable to match quality shocks, resulting in a higher divorce rate. If this alternative

is true, then the data should show that an increase in marriage rates is associated with an

increase in bankruptcy exemption levels. I test this hypothesis using marriage data from the

Vital Statistics of the United States over 1989-2005 by replacing the divorce rate per 1000

married persons with the marriage rate per 1000 total population as the dependent variable

in Equation 6 and present results in columns (6) and (7) of Table 5. In both speci�cations,

the coe�cients on the bankruptcy exemption levels fail to be signi�cantly di�erent from zero

at any conventional level.27 Thus, I �nd no evidence that changes in exemption levels are

associated with changes in marriage rates and thereby a�ect the divorce rate.

7 Conclusion

This paper investigates the e�ect of bankruptcy exemption levels on divorce rates, exploring

the relative level of insurance o�ered by marriage as a determinant of divorce not previ-

ously considered in the literature. I present a simple theoretical model of divorce decisions

and uncertain asset values in which marriage allows individuals to share risk. My empirical

strategy identi�es the e�ect of increases in exemption levels on divorce rates by exploiting

exogenous within-state variation in exemption levels over time. Under the assumption that

debtors can plan before bankruptcy to take full advantage of both homestead and nonhome

27As marriages are measured by state of occurrence, not state of residence of those getting married, I alsorun these regressions excluding Nevada, which has an extremely high marriage rate due to lax marriage laws.The results in Table 5 are robust to the exclusion of this outlier.

21

exemptions by moving assets between the two categories and that married debtors can dou-

ble any exemption level they are not speci�cally prohibited from doing so by law, I �nd that

increases in exemption levels are associated with increases in divorce rates and estimate that

exemption increases across the U.S. between 1989 and 2005 led to approximately 200,000

additional divorces over this period. I perform a variety of sensitivity tests on this speci�ca-

tion to address aspects of the divorce decision neglected by the theoretical model, showing

the di�erential e�ects that changes in bankruptcy exemptions can have under various di-

vorce laws and that the divorce debt provisions of BRA94 did not meaningfully a�ect the

estimated relationship. I also show that bankruptcy exemptions do not a�ect the divorce

rate by altering the marriage rate.

Ultimately, I �nd that the cumulative e�ect of the increases in exemption levels over this

sample is to raise the divorce rate per 1000 married persons approximately 0.151 percentage

points, a 2.37% increase over its 2005 level, with the magnitude growing larger each year.

This e�ect size is similar to that of other changes in divorce laws thought to be more imme-

diately relevant to divorce rates, such as the introduction of unilateral divorce. As such, the

increase in the divorce rate is a sizable unintended consequence of the changes in bankruptcy

exemption levels.

22

Appendix

Recall that(cBh , c

Bw

)is the solution to the problem

maxch,cw

Uh (ch, θh) + αUw (cw, θw) s.t. ch + cw ≤ βE.

Without loss of generality, I restrict attention to agent h. Under the assumption of Cobb-

Douglas utility, I obtain the unconstrained maximization problem

maxch

cγhh θ1−γhh + α (βE − ch)γw θ1−γw

w

with corresponding �rst order condition

γhcγh−1h θ1−γh

h = αγw (βE − ch)γw−1 θ1−γww .

The value of ch that solves this above equation is cBh . I rearrange to obtain

(cBh

)γh−1=

[αγwθ

1−γww

γhθ1−γhh

] (βE − cBh

)γw−1.

The �rst term on the right hand side is a constant where all terms are positive, so let

φ = αγwθ1−γww

γhθ1−γhh

. I now take the derivative of the above with respect to E to obtain

(∂cBh∂E

)γh−2

1− ∂cBh∂E

=φ (γw − 1)

(βE − cBh

)γw−2

γh − 1.

Since 0 < γw, γh < 1 and βE > cBh , the right hand side of this last equation is positive. As

such, the left hand side of this equation can only be positive when 0 <∂cBh∂E

< 1.

23

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Table 1: Bankruptcy Exemptions by State

1989 1995 2005State Homestead Nonhome Homestead Nonhome Homestead Nonhome Higher Homestead Federal

Alabama 5000 3000 5000 3000 5000 3000 Yes NoAlaska 54000 5800 54000 5800 67500 9000 No NoArizona 50000 4150 100000 4150 150000 7650 No NoArkansas unlimited 2150 unlimited 2150 unlimited 2150 No YesCalifornia 30000 3700 50000 3700 50000 8375 Yes NoColorado 20000 2800 30000 2800 45000 13600 Yes No

Connecticut 0 1500 75000 2500 75000 2500 Yes YesDelaware 0 5000 0 5000 0 5000 Yes No

District of Columbia 0 1050 0 1050 unlimited 5400 � YesFlorida unlimited 1000 unlimited 2000 unlimited 2000 � NoGeorgia 5000 1900 5000 1900 10000 5600 Yes NoHawaii 20000 1000 20000 1000 20000 2575 Yes YesIdaho 25000 1500 50000 2500 50000 5300 No NoIllinois 7500 3950 7500 3950 7500 3950 Yes NoIndiana 7500 2500 7500 2500 7500 2500 Yes NoIowa unlimited 15100 unlimited 15100 unlimited 15600 No NoKansas unlimited 27500 unlimited 27500 unlimited 27500 No NoKentucky 5000 3800 5000 3800 5000 3800 � NoLouisiana 15000 0 15000 0 25000 7500 No NoMaine 7500 2600 12500 7900 35000 10400 Yes No

Maryland 0 5500 0 8000 0 16000 � NoMassachusetts 100000 2800 100000 2350 500000 2300 No YesMichigan 3500 1000 3500 1000 3500 1000 � YesMinnesota unlimited 7000 unlimited 11200 200000 13300 No YesMississippi 30000 10000 75000 10000 75000 10000 � NoMissouri 8000 2900 8000 3400 15000 6600 No NoMontana 40000 4700 40000 4700 100000 6000 � NoNebraska 10000 1500 10000 1500 12500 2400 No NoNevada 90000 5500 95000 6000 200000 19500 No No

New Hampshire 5000 2600 30000 2600 100000 10400 � YesNew Jersey 0 1000 0 1000 0 1000 � YesNew Mexico 20000 6000 30000 6000 30000 6000 Yes YesNew York 10000 3000 10000 3000 10000 3600 Yes No

North Carolina 7500 1500 10000 2250 10000 2750 Yes NoNorth Dakota 80000 3700 80000 3700 80000 3700 No No

Ohio 5000 2550 5000 2550 5000 2550 � NoOklahoma unlimited 8000 unlimited 8000 unlimited 8000 No NoOregon 15000 7350 25000 12600 25000 12600 Yes No

Pennsylvania 0 300 0 300 0 300 � YesRhode Island 0 550 0 550 200000 11250 No YesSouth Carolina 5000 1950 5000 1950 5000 1950 Yes NoSouth Dakota unlimited 2000 unlimited 2000 unlimited 4000 No NoTennessee 5000 4750 5000 4750 5000 5900 Yes NoTexas unlimited 15000 unlimited 30000 unlimited 30000 No YesUtah 8000 3000 8000 3000 20000 6000 Yes No

Vermont 30000 8600 30000 8600 75000 8600 Yes YesVirginia 5000 0 5000 12000 5000 12000 Yes No

Washington 30000 4700 30000 8500 40000 9500 No YesWest Virginia 7500 2350 7500 2350 25000 4700 Yes NoWisconsin 40000 1000 40000 9700 40000 9700 No YesWyoming 10000 2000 10000 4000 10000 4400 Yes No

Federal 7500 2350 15000 4700 18450 5775 Yes �

Exemption amounts from state statutes. �Higher Homestead� indicates whether married couples are speci�cally given by law a higher homesteadexemption than single individuals. �Federal� indicates whether a state allows its residents to choose to use the federal exemption levels in place ofthe state levels. Data on the homestead exemption for married couples from Elias et al. (2005) and previous editions.

29

Table 2: Divorce Data by State

Divorce Laws Avg. Divorces/1000 Married PersonsState Data Available Unilateral Community Property 1989-1994 1995-2000 2001-2005

Alabama 1989-2005 Yes No 10.63 9.83 9.02Alaska 1989-2005 Yes Yes 9.68 7.89 7.43Arizona 1989-2005 Yes Yes 10.57 9.68 7.46Arkansas 1989-2005 No No 11.80 10.45 10.37California 1989 Yes Yes 8.03 � �Colorado 1989-1994, 2001-2005 Yes No 9.36 � 7.56

Connecticut 1989-2005 Yes No 5.73 4.70 5.43Delaware 1989-2005 No No 8.12 8.09 6.74

District of Columbia 1989-2005 No No 13.31 9.04 6.89Florida 1989-2005 Yes No 10.51 9.58 8.76Georgia 1989-2003 Yes No 9.72 8.04 5.08Hawaii 1989-2002 Yes No 7.72 7.12 7.34Idaho 1989-2005 Yes Yes 9.66 8.92 8.32Illinois 1989-2005 No No 6.81 5.96 5.12Indiana 1989 Yes No 10.21 � �Iowa 1989-2005 Yes No 6.54 5.61 4.72Kansas 1989-2005 Yes No 8.22 7.01 5.59Kentucky 1989-2005 Yes No 9.56 9.07 8.27Louisiana 1989, 2002-2003 No No 4.03 � 6.10Maine 1989-2005 Yes No 7.41 6.89 7.64

Maryland 1989-2005 No No 6.07 5.62 5.73Massachusetts 1989-2005 Yes No 4.89 4.71 4.43Michigan 1989-2005 Yes No 7.38 7.01 6.49Minnesota 1989-2004 Yes No 6.26 5.25 5.00Mississippi 1989-2005 No No 9.45 9.36 8.70Missouri 1989-2005 No No 8.28 8.28 6.80Montana 1989-2005 Yes No 8.24 6.54 7.05Nebraska 1989-2005 Yes No 6.33 6.22 5.73Nevada 1989-1990, 1994-2005 Yes Yes 18.34 16.05 12.31

New Hampshire 1989-2005 Yes No 7.71 8.59 6.89New Jersey 1989-2005 No No 5.65 5.58 5.52New Mexico 1989-2005 Yes Yes 9.67 9.77 8.74New York 1989-2005 No No 6.03 6.28 6.12

North Carolina 1989-2005 No No 8.51 8.11 7.49North Dakota 1989-2005 Yes No 5.64 5.47 4.99

Ohio 1989-2005 No No 7.88 7.28 6.68Oklahoma 1989-2000, 2004-2005 Yes No 11.36 9.19 8.74Oregon 1989-2005 Yes No 8.71 8.06 7.65

Pennsylvania 1989-2005 No No 5.64 5.61 5.10Rhode Island 1989-2005 Yes No 6.09 6.02 5.76South Carolina 1989-2005 No No 7.10 6.64 5.93South Dakota 1989-2005 Yes No 6.39 6.37 5.30Tennessee 1989-2005 No No 11.36 10.20 8.54Texas 1989-1995, 1997-2005 Yes Yes 9.45 7.45 6.40Utah 1989-2005 No No 7.56 6.93 6.60

Vermont 1989-2005 No No 7.98 9.11 7.07Virginia 1989-2005 No No 7.45 7.55 7.03

Washington 1989-2005 Yes Yes 9.45 8.67 7.71West Virginia 1989-2005 No No 8.74 8.52 8.75Wisconsin 1989-2005 No Yes 6.10 5.72 5.35Wyoming 1989-2005 Yes No 10.46 9.86 8.71

Unilateral divorce coding is due to Friedberg (1998). Divorce rates are averages over the period listed for all years where data are available.

30

Table 3: Main ResultsDependent Variable is Divorce Rate per 1000 Married Population

Indep. Variable (1) (2) (3) (4) (5) (6) (7) (8)

Total Exemption 0.0611* 0.0605** 0.0496** 0.0457** 0.0466* 0.0491** 0.0473** 0.0433**(0.0311) (0.0259) (0.0199) (0.0210) (0.0267) (0.0243) (0.0183) (0.0205)

Total Exemption2 -0.00117*** -0.000935*** -0.000843*** -0.000751** -0.000560 -0.000528 -0.000620** -0.000565**(0.000393) (0.000320) (0.000274) (0.000301) (0.000497) (0.000419) (0.000246) (0.000251)

Real Personal Income 0.851 0.733 0.556 0.388 0.350 0.114(0.910) (0.853) (0.886) (0.777) (0.726) (0.758)

Unemployment Rate -0.0981 -0.0626 -0.0679 -0.125** -0.0963 -0.0999(0.0676) (0.0656) (0.0665) (0.0614) (0.0628) (0.0618)

Real Median House Price -0.0320 -0.0424 -0.0293 -0.0611 -0.0607 -0.0517(0.0627) (0.0539) (0.0496) (0.0600) (0.0547) (0.0499)

Homeownership Rate -4.677** -4.721** -3.977* -5.043** -5.152*** -4.658**(2.230) (1.981) (2.027) (2.112) (1.904) (1.924)

Percent Hispanic -42.21** -43.34*** -39.58** -40.28**(15.90) (15.96) (16.44) (16.50)

Percent Black 13.62 15.63 -18.62 -17.42(20.85) (20.04) (23.55) (23.32)

Percent Age 15-64 4.095* 4.362* 4.242* 4.445**(2.367) (2.291) (2.183) (2.146)

Child Custody Guidelines -0.204 -0.199 -0.412 -0.400(0.252) (0.251) (0.267) (0.259)

No Fault Maintenance -0.0889 -0.0748 0.0520 0.0621(0.104) (0.0829) (0.121) (0.107)

Covenant Marriage -0.840*** -0.819*** -0.468 -0.433(0.149) (0.166) (0.294) (0.292)

Real Max AFDC/TANF Payment 0.000730 0.00180(0.00114) (0.00134)

Real Max EITC Payment 0.000575 0.000606(0.000362) (0.000380)

Constant 7.359*** 8.559** 6.803 5.903 7.642*** 10.88*** 13.51** 12.49**(0.178) (3.388) (5.721) (5.394) (0.268) (2.853) (5.465) (5.246)

Homestead Exemption De�ned De�ned De�ned De�ned All All All AllNumber of obs. 674 674 674 674 801 801 801 801State Dummies Yes Yes Yes Yes Yes Yes Yes Yes

State Linear and Quadratic Trends Yes Yes Yes Yes Yes Yes Yes YesR2 0.943 0.945 0.947 0.948 0.948 0.950 0.952 0.953

Regression Type WLS WLS WLS WLS WLS WLS WLS WLS

Standard errors in parentheses. All standard errors are Huber-White robust estimates, clustered at the state level. *, **, *** denotes signi�canceat the 10%, 5%, or 1% level, respectively.

31

Table 4: Divorce Rate Increases

De�ned Exemption States Estimates from 3.4 All States Estimates from 3.8Year Mean Total Exemption Divorce Rate Increase Additional Divorces Mean Total Exemption Divorce Rate Increase Additional Divorces1989 $53,949.60 � � $106,632.90 � �1990 $52,835.42 -0.004 -614 $108,476.90 0.006 8391991 $61,453.46 0.028 4,066 $117,942.90 0.035 5,0661992 $60,280.62 0.024 3,451 $121,120.70 0.044 6,4731993 $62,070.73 0.030 4,492 $125,547.40 0.057 8,5381994 $60,750.44 0.025 3,801 $126,260.00 0.059 8,9151995 $66,864.43 0.047 7,180 $133,773.80 0.081 12,2421996 $71,351.54 0.063 9,569 $133,313.10 0.079 12,0241997 $70,562.06 0.060 9,152 $134,395.00 0.082 12,4891998 $70,053.75 0.059 9,004 $134,993.60 0.084 12,9201999 $68,837.61 0.054 8,402 $135,628.50 0.086 13,2832000 $64,027.52 0.037 5,935 $134,668.80 0.083 13,2962001 $67,460.20 0.049 7,941 $140,736.80 0.100 16,0682002 $74,641.51 0.075 12,189 $147,699.50 0.119 19,4172003 $76,076.46 0.080 13,059 $150,487.90 0.126 20,7242004 $74,979.52 0.076 12,494 $151,402.60 0.129 21,2112005 $79,993.07 0.093 15,478 $160,014.10 0.151 25,133Total 125,599 208,637

3.4 refers to Table 3, speci�cation (4). Mean total exemption is the married population-weighted mean of the real value of total exemptionsavailable in each year.

32

Table5:

SensitivityAnalysis

DependentVariable:

Div.Rate

Div.Rate

Div.Rate

Div.Rate

Div.Rate

Marr.Rate

Marr.Rate

Indep.Variable

(1)

(2)

(3)

(4)

(5)

(6)

(7)

TotalExem

ption

0.0336

0.0301

0.0461**

0.0958**

0.111***

0.0196

0.0230

(0.0300)

(0.0293)

(0.0208)

(0.0384)

(0.0362)

(0.0262)

(0.0270)

TotalExem

ption

2-0.000662

-0.000534

-0.000748**

-0.00197

-0.00246***

0.000139

0.0000431

(0.000479)

(0.000417)

(0.000294)

(0.00188)

(0.000808)

(0.000524)

(0.000405)

TotalExem

ption*CPi

0.174

0.122

(0.217)

(0.0999)

TotalExem

ption

2*CPi

-0.00300

-0.00167

(0.00767)

(0.00202)

TotalExem

ption*UnlimHom

e i-0.328

(0.434)

TotalExem

ption

2*UnlimHom

e i0.00407

(0.00568)

TotalExem

ption*Unilat i

-0.0700

-0.0830*

(0.0466)

(0.0460)

TotalExem

ption

2*Unilat i

0.00148

0.00208**

(0.00194)

(0.000913)

Constant

5.394

6.085

6.846

8.527

13.76***

2.894

13.24

(5.946)

(5.973)

(5.986)

(5.351)

(5.050)

(10.71)

(8.122)

Hom

estead

Exem

ption

De�ned

All

All

De�ned

All

De�ned

All

Number

ofobs.

674

801

801

674

801

735

863

F-test,BankruptcyInteractions

4.04**

1.04

0.33

1.24

2.01

��

State

Dummies

Yes

Yes

Yes

Yes

Yes

Yes

Yes

State

LinearandQuadraticTrends

Yes

Yes

Yes

Yes

Yes

Yes

Yes

AdditionalControls

Yes

Yes

Yes

Yes

Yes

Yes

Yes

R2

0.952

0.958

0.957

0.951

0.956

0.992

0.991

RegressionType

WLS

WLS

WLS

WLS

WLS

WLS

WLS

Standard

errors

inparentheses.

Allstandard

errors

are

Huber-Whiterobust

estim

ates,clusteredatthestate

level.

*,**,***denotessigni�canceatthe10%,5%,or1%

level,respectively.Dependent

variable

�Div.Rate�is

thedivorcerate

per1000marriedpersons,

while�M

arr.Rate�is

themarriagerate

per1000totalpopulation.F-testsreport

thetest

statistic

ofthejointhypothesisthatthe

coe�cients

onallinteractionvariablesare

0.

33

Figure 1: Federal Divorce Rate

3.5

44.

55

5.5

66.

57

7.5

88.

5D

ivor

ce R

ate

1989 1991 1993 1995 1997 1999 2001 2003 2005Year

Divorce Rate per 1000 total populationDivorce Rate per 1000 married persons

Federal Divorce Rate

34