Upload
others
View
5
Download
0
Embed Size (px)
Citation preview
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: [email protected]
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; [email protected].
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
References
Alexander, Peter C., �Divorce and the Dischargeability of Debts: Focusing on Women as
Creditors in Bankruptcy,� Catholic University Law Review, Winter 1994, 43 (2), 351�398.
Allen, Douglas W., �Marriage and Divorce: Comment,� American Economic Review, June
1992, 82 (3), 679�685.
Armour, Stephanie, �More families move in together,� USA Today, February 3, 2009,
pp. 1B�2B.
Becker, Gary S., A Treatise on the Family, Cambridge: Harvard University Press, 1981.
Bello, Ottilie, �Bankruptcy and Divorce: The Courts Send a Message to Congress,� Pace
Law Review, Fall 1993, 13 (2), 643�719.
Berkowitz, Jeremy and Richard Hynes, �Bankruptcy Exemptions and the Market for
Mortgage Loans,� Journal of Law and Economics, October 1999, 42 (2), 809�830.
Brinig, Margaret F. and F. H. Buckley, �No-fault laws and at-fault people,� Interna-
tional Review of Law and Economics, September 1998, 18 (3), 325�340.
Clark, Simon, �Law, Property, and Marital Dissolution,� The Economic Journal, March
1999, 109 (454), C41�C54.
Clark-Stewart, Alison and Cornelia Brentano, Divorce: Causes and Consequences,
New Haven: Yale University Press, 2006.
Conger, Rand D., Glen H. Elder Jr., Frederick O. Lorenz, Katherine J. Conger,
Ronald L. Simons, Les B. Whitbeck, Shirley Huck, and Janet Melby, �Link-
ing Economic Hardship to Marital Quality and Instability,� Journal of Marriage and the
Family, August 1990, 52 (3), 643�656.
24
Domowitz, Ian and Robert L. Sartain, �Determinants of the Consumer Bankruptcy
Decision,� The Journal of Finance, February 1999, 54 (1), 403�420.
Edmiston, Kelly D., �A New Perspective on Rising Nonbusiness Bankruptcy Filing Rates:
Analyzing the Regional Factors,� Federal Reserve Bank of Kansas City Economic Review,
Second Quarter 2006, pp. 55�83.
Elias, Stephen, Robin Leonard, and Albin Renauer, How to File for Chapter 7
Bankruptcy, Berkeley: Nolo Press, 2005.
Elrod, Linda D. and Robert G. Spector, �A Review of the Year in Family Law: Parent-
age and Assisted Reproduction Problems Take Center Stage,� Family Law Quarterly, Win-
ter 2006, 39 (4), 879�924.
Fay, Scott, Erik Hurst, and Michelle J. White, �The Household Bankruptcy Decision,�
American Economic Review, June 2002, 92 (3), 706�718.
Friedberg, Leora, �Did Unilateral Divorce Raise Divorce Rates? Evidence from Panel
Data,� American Economic Review, June 1998, 88 (3), 608�627.
Gropp, Reint, John Karl Scholz, and Michelle J. White, �Personal Bankruptcy and
Credit Supply and Demand,� Quarterly Journal of Economics, February 1997, 112 (1),
217�251.
Gruber, Jonathan, �Is Making Divorce Easier Bad for Children? The Long-Run Implica-
tions of Unilateral Divorce,� Journal of Labor Economics, October 2004, 22 (4), 799�833.
Hynes, Richard M., Anup Malani, and Eric A. Posner, �The Political Economy of
Property Exemption Laws,� Journal of Law and Economics, April 2004, 47 (1), 19�43.
Jacoby, Melissa, Teresa Sullivan, and Elizabeth Warren, �Medical Problems and
Bankruptcy Filings,� No. 008, 2000. Harvard Law School Public Law and Legal Theory
Working Paper Series.
25
Johnson, Meredith, �At the Intersection of Bankruptcy and Divorce: Property Division
Debts under the Bankruptcy Reform Act of 1994,� Columbia Law Review, January 1997,
97 (1), 91�132.
Kotliko�, Laurence J. and Avia Spivak, �The Family as an Incomplete Annuities Mar-
ket,� The Journal of Political Economy, April 1981, 89 (2), 372�391.
Leland, John, �Breaking Up Is Harder to Do After Housing Fall,� New York Times, De-
cember 30, 2008, p. A1.
Lin, Emily Y. and Michelle J. White, �Bankruptcy and the Market for Mortgage and
Home Improvement Loans,� Journal of Urban Economics, July 2001, 50 (1), 138�162.
Livshits, Igor, James MacGee, and Michele Tertilt, �Consumer Bankruptcy: A Fresh
Start,� American Economic Review, March 2007, 97 (1), 402�418.
Lloyd, Carol, �Breaking up is harder to do: The housing bust's in�uence on
divorce,� San Francisco Chronicle, April 4, 2008. Retrieved from SFGate.com,
http://www.sfgate.com/cgi-bin/article.cgi?f=/g/a/2008/04/04/carollloyd.DTL.
Manser, Marilyn and Murray Brown, �Marriage and Household Decision-Making: A
Bargaining Analysis,� International Economic Review, February 1980, 21 (1), 31�44.
Margolick, David, �Can Bankruptcy Reduce The Price of a Divorce?,� New York Times,
March 2, 1991, p. A1.
Matouschek, Niko and Imran Rasul, �The Economics of the Marriage Contract: Theo-
ries and Evidence,� Journal of Law and Economics, February 2008, 51 (1), 59�110.
McElroy, Marjorie B. and Mary Jean Horney, �Nash-Bargained Household Decisions:
Toward a Generalization of the Theory of Demand,� International Economic Review, June
1981, 22 (2), 333�349.
26
Nixon, Lucia A., �E�ect of Child Support Enforcement on Marital Dissolution,� Journal
of Human Resources, Winter 1997, 32 (1), 159�181.
Peters, H. Elizabeth, �Marriage and Divorce: Informational Constraints and Private Con-
tracting,� American Economic Review, June 1986, 76 (3), 437�454.
, �Marriage and Divorce: Reply,� American Economic Review, June 1992, 82 (3), 686�693.
Posner, Eric A., �The Political Economy of the Bankruptcy Reform Act of 1978,� Michigan
Law Review, October 1997, 96 (1), 47�126.
Romans, Christine, �Economy prolongs some marriages,
ends others,� March 3, 2009. Retrieved from CNN.com,
http://www.cnn.com/2009/LIVING/03/03/divorce.economy/index.html.
Ruser, John, Adrienne Pilot, and Charles Nelson, �Alternative Measures of Household
Income: BEA Personal Income, CPSMoney Income, and Beyond,� 2004. Federal Economic
Statistics Advisory Committee Report.
Sullivan, Teresa, Elizabeth Warren, and Jay Lawrence Westbrook, The Fragile
Middle Class, New Haven: Yale University Press, 2000.
Sweezy, Kate and Jill Tiefenthaler, �Do State-Level Variables A�ect Divorce Rates?,�
Review of Social Economy, Spring 1996, 54 (1), 47�65.
Tabb, Charles Jordan, �The History of the Bankruptcy Laws in the United States,�
American Bankruptcy Institute Law Review, Summer 1995, 3, 5�51.
Topolnicki, Denise M. and Elizabeth M. Macdonald, �The Bankruptcy Bonanza!,�
Money, August 1993, pp. 82�94.
White, Michelle, �Personal Bankruptcy under the 1978 Bankruptcy Code: An Economic
Analysis,� Indiana Law Journal, 1987, 63 (1), 1�53.
27
Wolfers, Justin, �Did Unilateral Divorce Laws Raise Divorce Rates? A Reconciliation and
New Results,� American Economic Review, December 2006, 96 (5), 1802�1820.
28
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