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The intergenerational e�ects of welfare participation∗
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Yannay Shanan†
February 10, 2020
I estimate the intergenerational e�ects of welfare participation by studying the impli-
cations of an Israeli welfare reform that increased welfare bene�ts generosity and eased eli-
gibility requirements for single mothers during the 1990s. Using a di�erence-in-di�erences
framework, I compare the outcomes of children of single mothers relative to those of mar-
ried mothers before and after the reform was introduced. I show that the reform led to
higher welfare participation rates and a decline in labor supply among single mothers and
that these changes had spillover e�ects on their children's long-term economic outcomes.
Children's future welfare participation rates signi�cantly increased following the reform,
but so did their labor force participation rates and labor earnings, implying an increase in
parental investment as a result of higher discretionary income. Moreover, I �nd that the
e�ect on welfare participation rates was not permanent and was driven by girls, while the
impact on employment rates and earnings was long-lasting and largely driven by boys.
(JEL I38 J62 H53)
Keywords: Intergenerational transmission, Welfare participation, Single mothers
∗This paper is based on a chapter of my PhD dissertation. I am deeply indebted to my advisorAnalia Schlosser for her valuable guidance and support in this project. I would also like to thank MosheHazan and David Weiss for helpful comments. I thank the National Insurance Institute (NII) in Israelfor allowing restricted access to data in the NII protected research lab.†The Eitan Berglas School of Economics, Tel Aviv University, POB 39040, Ramat Aviv, Tel Aviv,
69978, Israel. Email: [email protected]
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1 Introduction
Estimating the intergenerational e�ects of welfare participation is essential in order to
understand the reasons behind persistent welfare dependence and to design e�ective la-
bor market policies. The positive correlation in welfare participation across generations
is well-documented in the literature (see Page (2004) for a review of estimates for the
US). It is unclear however how much of that correlation is causal and how much is simply
driven by the intergenerational correlation in income and other correlated genetic or en-
vironmental characteristics that are correlated between parents and their children (Black
et al., 2011). If, for example, children of welfare recipients learn to become dependent
on the welfare system from their parents, irrespective of any other correlated attributes,
even a temporary increase in welfare participation rates may increase the number of
future welfare recipients.
The potential mechanisms that may explain a casual intergenerational transmission of
welfare participation include: (a) lessened distaste for welfare among children, such that
stigma may no longer work as a deterrent, (b) acquisition by the children of �rst-hand
knowledge on how the system works and what needs to be done to obtain welfare bene�ts,
and (c) reduced informal access to job opportunities via parental social networks and
limited development of proper work etiquette and other non-cognitive skills as a result of
low parental labor force attachment. On the other hand, there also forces that may work
in the opposite direction: (a) the decrease in parental work hours associated with welfare
participation means more time at home, which may lead to greater parental investment
of time in their children; and (b) the potential increase in discretionary income may be
translated into additional investment in the children's education, health, and soft skills.
In this study, I estimate the intergenerational e�ects of parental welfare participation
on children's long-term outcomes by exploiting the e�ects of an Israeli welfare reform that
expanded bene�ts and eased eligibility requirements for single mothers in 1992.1 This is
the �rst study to examine the intergenerational e�ects of welfare participation in Israel,
and one of only a handful of papers that have examined this question using a quasi-
experimental design. Using a comprehensive dataset that draws on several sources of
administrative records, I compare the di�erences in long-term outcomes between children
of single mothers who grew up in the pre-reform period and children of single mothers who
were adolescents when the reform took place, relative to the di�erences between children
of the same cohorts born to married mothers, in a di�erence-in-di�erences framework.
I �nd that single mothers increased their reliance on welfare bene�ts and decreased
their labor supply following the reform and that these changes had spillover e�ects on their
children's future economic outcomes. The next generation's welfare participation rates
1While the reform was also relevant for single fathers, I focus solely on mothers, since less than 10%of single-parent households in Israel are headed by men.
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increased signi�cantly, but so did their likelihood to ever be employed and their life-time
labor earnings. I �nd these seemingly contradicting average e�ects mask heterogeneity
across children and over the life cycle. The e�ect on welfare participation rates is driven
by an increase in girls' welfare participation rates while the e�ects on employment rates
and labor earnings are largely driven by boys. Furthermore, the increase is welfare is
temporary and occurs only during early adulthood (i.e between the ages 21-30). Later
on, when the children are 31-40 there is no di�erence in welfare participation between
children who were and were not exposed to the reform at childhood. The e�ect on labor
earnings, on the other hand, does not fade away with time, suggesting that the increase
in parental resources led to long lasting gains for the next generation.
The increase in children's future welfare participation rates is consistent with the
children learning how the system works and\or adopting a less negative attitude towards
welfare and the stigma associated with it. The increase in earnings is consistent with an
increase in household discretionary income and\or an increase in time spent at home that
prompted an increase in parental investment. These �ndings highlight the importance
of considering spillover e�ects in social assistance programs, as well as the need to con-
sider the long-term e�ects of parental welfare participation beyond that of future welfare
participation rates.
The paper is related to the growing literature on government transfers and their
short to medium-term e�ects on children (Chen et al., 2015, Chetty et al., 2011, Dahl
and Lochner, 2012, Dahl and Lochner, 2017, Løken et al., 2018, Milligan and Stabile,
2011). These studies tend to �nd that additional parental resources in the form of the
Earned Income Tax Credit (EITC) and other social assistance programs lead to increased
investment in children which in return results in improved outcomes, such as test scores
and health outcomes.
It is also related to a not-so-current literature on the intergenerational transmission
of welfare participation, based mainly on observational studies, which provide mixed
evidence on the e�ect of welfare participation on children's future welfare participation
rates (Antel, 1992, Beaulieu et al., 2005, Gottschalk, 1990, Gottschalk, 1996, Levine and
Zimmerman, 1996, Pepper, 2000).
Of particular relevance to my work, two recent quasi-experimental studies which ex-
amine the intergenerational e�ects of parental Disability Insurance (DI) in Norway and
the Netherlands on children's future receipt of DI. Dahl et al. (2014) exploit random
variation in judges' leniency among Norwegian DI applicants whose cases were initially
denied and �nd that when a parent is granted DI in the appeal stage their adult child's
probability of becoming a DI recipient increases. Dahl and Gielen (2018) use a regression
discontinuity design based on a reform in the Netherlands that tightened DI criteria and
�nd that children of parents who whose application for DI was rejected were less likely
to receive DI themselves. It is important to note that DI is slightly di�erent than tra-
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ditional welfare programs. In contrast to means-tested welfare, DI is often an absorbing
state, meaning that once an individual starts receiving this form of bene�t, he hardly
ever returns to self-support. In addition, DI eligibility depends on having a physical or
mental impairment and is usually not a function of household income.
This study is also closely related to a recent working paper by Hartley et al. (2017)
which exploits cross-state variation created by the 1990s reforms of the Aid to Families
with Dependent Children (AFDC) program in the US in order to estimate the inter-
generational transmission of welfare participation and other economic outcomes. They
�nd that maternal welfare participation leads to a 28-percentage-point increase in the
likelihood to ever be on welfare, an 18-percentage-point increase in the likelihood to have
no earnings, and a 70-percentage-point increase in the likelihood to have lower human
capital attainment. Their study di�ers from the current one in several ways. First, their
analysis draws on survey data from The Panel Study of Income Dynamics (PSID) which,
unlike data based on administrative records, may su�er from misreporting of welfare
participation.2 The authors are aware of this issue and tackle the misclassi�cation of wel-
fare recipiency using various methods. Second, their identi�cation strategy relies on the
assumptions that state-level trends in welfare generosity are not correlated with future
welfare participation rates of children and that there is no endogenous migration across
states.
This paper contributes in several ways to the modest empirical literature on the inter-
generational e�ects of welfare participation. First, I exploit a nationwide policy change
that generates quasi-experimental variation in welfare participation in order to estimate
the e�ects of parental welfare participation. Second, I focus on the intergenerational ef-
fects of means-tested welfare as opposed to those of disability insurance programs. Third,
I employ a comprehensive administrative micro-data set that allows me to estimate the
impact of the reform for both mothers and their children, using children outcomes from
adolescence to age 40. Finally, I do not focus solely on the intergenerational transmission
of welfare participation, but rather I estimate the long-term e�ects of parental welfare
participation on various economic outcomes, thus highlighting the potential bene�ts of
social assistance programs in the long run.
The remainder of the paper proceeds as follows: In the next section, I provide the
institutional background by describing the key features of the Israeli welfare system and
the 1992 Single Parent Law which made welfare highly appealing for single-parent families
during the 1990s. Section 3 describes the data. Section 4 discusses the identi�cation
strategy for estimating both the direct impact of the reform on single mothers and the
long-term e�ects on their children. Section 5 presents the estimates of the e�ect of
the reform on the contemporaneous labor market outcomes of single mothers, including
2It has been documented that under-reporting of welfare participation is common in major householdsurveys such as the PSID and that misreporting seems to have increased over time (Meyer et al., 2009).
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results of a variety of robustness checks. Section 6 presents the estimates of the e�ect
of the reform on the long-term economic outcomes of children of single mothers who
were exposed to the reform as adolescents and explores the mechanism underlying the
estimated e�ects, as well as providing estimates from a series of falsi�cation tests. Section
7 concludes.
2 Background
2.1 The Israeli welfare system
The main social assistance program in Israel is called Income Support (or �Guaranteed
Income� in Hebrew), which provides aid to families with little or no income.3 It is a
classic means-tested bene�t that most closely resembles the Temporary Assistance for
Needy Families (TANF) program in the U.S or the Income Support program in the
U.K. Unlike TANF, it is not limited to parents with dependent children and does not
have any time limits. Eligible households receive monthly cash assistance, as well as a
variety of valuable in-kind bene�ts, including rent and mortgage assistance, a reduction in
municipal property taxes, access to public housing, and subsidies for a range of household
expenditures such as public transportation, a phone line, and electricity.
Eligibility is based on income, assets, and an employment test. Those who are able
to work and are not exempt according to the criteria are also expected to actively search
for a job and to report weekly at the employment service. Failure to comply can result
in loss of bene�ts. The monthly allowance is a function of income, age, marital status,
and number of dependent children. For instance, in 1991 (one year before the reform),
the maximal allowance for a single mother under 55 with one child was about 2,250 NIS
(2010 prices), which constituted 37.5% of the average monthly wage at the time.
The intergenerational correlation in welfare participation in Israel has not previously
been estimated, to the best of my knowledge. Using the administrative data described
in section 3, I calculate it for the entire population of Israeli women born between 1974
and 1986. For the sake of comparability, I adopt the same de�nitions and observational
time windows as Page (2004) who estimates the intergenerational correlation in welfare
participation between mothers and daughters in the US, among daughters born between
the years 1951 and 1966. She uses a fairly broad de�nition of welfare participation,
de�ning it as the receipt of either Aid to Families with Dependent Children (AFDC)
bene�ts, General Assistance, Food Stamps, or Supplemental Security Income. In the
Israeli context, welfare participation is simply de�ned as the receipt of Income Support
bene�ts. In the US, daughters of welfare recipient mothers aged 14-16 are 2.45 times more
likely to be on welfare by age 27 than daughters of mothers who were not on welfare at
3Throughout the paper, I refer to Income Support simply as welfare.
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the time. I �nd the equivalent relative risk statistic4 for the 1974-1986 cohorts in Israel
to be 3.15.
2.2 The reform
In April 1992, the Israeli parliament enacted the Single Parent Law which provided
assistance to single parents in addition to that already provided by the 1982 Income
Support Law. The law increased the welfare bene�ts for single parents and also raised
the maximal permitted monthly income for single-parent households, allowing them to
earn a relatively large salary without losing their eligibility. At the time, there were
approximately 65,000 single, widowed or divorced mothers, consisting 10% of all mothers.
Of those, 13% were welfare recipients. By 1995, welfare participation rates among non-
married mothers had risen to as high as 30%.
Until 1992, single or divorced mothers with one child who were eligible for welfare
received a maximal monthly allowance equal to 30% of the average wage for the �rst
two years and an allowance equal to 37.5% of the average wage for each subsequent year.
Having more than one child added another 5% of the average wage. As of 1992, single
mothers with one child received a maximal allowance of 42.5% of the average wage starting
from the very �rst year. Single mothers with two or more children received 47.5%.5 In
addition, the law provided single mothers with supplementary in-kind bene�ts in the form
of income tax credits and subsidization of daycare and school fees. In 1994 and 1995,
legislative changes eased the terms of eligibility by expanding the de�nition of a single
parent to additional groups,6 and increased the allowance by another 2.5% of the average
wage for each of the �rst two children. Meanwhile, the bene�ts for married mothers
remained unchanged.
Figure 1 shows the payment schedule for a single-parent household with one child
before and after the 1992 reform. The X axis is household earnings as a share of the
average wage, while the Y axis is combined household income composed of both earnings
and welfare allowances as a share of the average wage. Before 1992, eligible single mothers
with no earnings received a cash bene�t equal to 30% of the average wage. All earnings
above that but below 17% of the average wage were ignored and did not reduce the size
of the transfer. Above that threshold, higher earnings resulted in a gradual deduction
of bene�ts, up to 47% of the average wage, where the bene�ts become zero. In 1992,
the entire curve shifted upward, and the 47% cap was replaced by a continuous and
gradual deduction in bene�ts up the point they were entirely o�set, when household
4De�ned as P (welfare=1|parental welfare=1)P (welfare=1|parental welfare=0) .
5The 1992 legislation essentially equated the welfare bene�ts of divorced and single mothers to thoseof widows who already enjoyed higher monthly welfare bene�ts.
6Including women who are separated and women whose husbands refuse to divorce them accordingto Jewish religious law.
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earnings reached 80% of the average wage. In 1994, the curve shifted upward by an
additional 2.5% of the average wage. The labor supply response to these changes is
determined by pre-reform earnings. Eligible women earning under the 47% threshold
before the reform faced a negative income e�ect and no substitution e�ect, since their
total income increased but the marginal value of an additional earned Shekel remained
unchanged. Those who earned above that but less than 80% of the average wage faced
both a negative income e�ect and a negative substitution e�ect, since their total income
increased and the marginal value of an additional earned Shekel decreased. Women who
earned just above the 80% cut-o� might have also been incentivized to reduce their labor
supply in order to be eligible for the in-kind bene�ts at the cost of slightly lower overall
income. The only possibly positive impact on labor supply would be at the old earnings
limit (i.e. the 47% threshold). These women may have wanted to increase their work
hours in order to locate themselves between the 47% and 80% levels, an option that
was not previously available to them. Overall, we would expect single women to reduce
their labor supply and increase their welfare participation rates as a result of the reform.
The e�ect of the legislation on the labor supply of single mothers during the 1990s was
previously estimated by Flug and Kasir (2006) and Frish and Zussman (2008). Both
found that the reform led to a reduction in single mothers' employment rates but di�er
somewhat on the magnitude of the e�ect.7
3 Data
The data consist of administrative records provided by the National Insurance Institute
of Israel (NII) and includes all mothers of children under 18 and their children for the
period 1988-1995. The data were collected by the NII from various sources (including
the Israeli Population Registry and the Tax Authority) and then merged and analyzed at
a secure research lab at the NII headquarters in Jerusalem. The data includes monthly
welfare bene�ts, (salaried) employment and wages for the years 1988-2017,8 year of birth,
gender, and complete marital status history. All nominal prices are converted to real
2010 prices using the CPI price index. In addition, the data include information on
any enrollment in an Israeli university or college during the period 2001-2017. Most
importantly, the data makes it possible to link mothers to their children through personal
identi�ers. Overall, the data provides a comprehensive and reliable picture of employment
and welfare participation across a long period of time for mothers and their o�spring.
Regrettably, the data does not contain information on work hours, income not from
7Flug and Kasir (2006) estimate a six-percentage-point decrease in employment by 1995, while Frishand Zussman (2008) estimate a three-percentage-point decrease.
81988 is the only year for which welfare participation is known but there is no information on theamount of welfare bene�ts received.
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welfare transfers or (reported) wages, nor educational outcomes other than post-secondary
education enrollment.
For reasons explained below, I restrict the sample to include only non-ultra-Orthodox
Jewish native-born mothers aged 21 to 54. The �rst sample, consisting of 2,729,896
observations, is constructed as a repeated cross-section of all mothers who meet these
criteria for the years 1988-1995. The sample is used to analyze the contemporaneous
e�ect of the reform on the welfare dependency and labor supply of single mothers. Table
1 provides summary statistics. We see that one year before the reform the average (native-
born Jewish) single mother already relies more heavily on welfare and works less than the
average married mother. They are also a little older and have signi�cantly fewer children.
All incomes and bene�ts are in thousands of 2010 NIS.
The second sample, which is used to estimate the intergenerational e�ects of the
reform, consists of children who were born to non-ultra-Orthodox Jewish native-born
women in the years 1973 and 1976. Children who passed away before reaching the age
of 40, or were born to mothers who were not Israeli residents when the child was 15 or
passed away before the child reached 18 are excluded. This sample consists of 53,414
children born to 48,692 unique mothers. Table 2 provides summary statistics for the
mothers of these two cohorts of children, for the year when the children were 15 years of
age.
4 Identi�cation strategy
A di�erence-in-di�erences framework is used to estimate both the contemporaneous and
intergenerational e�ects of the increase in welfare generosity. This method is meant to
di�erence out unmeasured confounders by eliminating bias from any group-invariant or
time-invariant factors. In this setting, married mothers who were not a�ected by the
reform serve as a comparison group. During the investigated sample period, there were
hardly any Arab or ultra-Orthodox Jewish single mothers. Furthermore, single mothers
were much more likely to be foreign-born, and to an even greater extent following the
wave of immigration from the Former Soviet Union in the early 1990s. In order to com-
pare similar groups of mothers, I restrict the entire analysis to non-ultra-Orthodox Jewish
native-born mothers. Note that the di�erence-in-di�erences model permits di�erences in
average levels of characteristics and outcomes between the treatment group and the com-
parison group, such that the di�erences on average between single mothers and married
mothers, as they appear in table 1, are of little concern. Instead, the model relies on
a common trend assumption which posits that any important unmeasured variables are
either time-invariant group attributes or time-varying factors that are group-invariant.
I start by estimating the following regression for the population of native-born Jewish
mothers during the period 1988-1995:
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Yit = α + β ∗ Singleit ∗ Postt + γ ∗ Singleit + δt +XitΦ+ εit (1)
where Yit is the outcome of woman i in year t ; Singleit is equal to one if woman i
is not married in year t and zero otherwise; Postt is equal to one if t ≥ 1992, and
zero otherwise; δt are year �xed e�ects; Xit is a quadratic in age; and εit are random
error terms. β captures the di�erence in outcomes between married and non-married
mothers, before and after the reform. The model assumes that if not for the reform the
di�erences between the two groups would have remained the same and that marital status
is endogenously determined by the reform. I address the validity of these two assumptions
in section 5. In order to examine the di�erence-in-di�erences in the years leading up to
and during the reform I also estimate the following event study speci�cation:
Yit = α + β ∗∑
Y eart ∗ Singleit + γ ∗ Singleit + δt +XitΦ+ εit (2)
where∑Y eart ∗ Singleit is a vector of interactions of Singleit with year dummies, such
that the omitted year is 1991 (i.e. the year preceding the reform). This speci�cation not
only enables the examination of the e�ect over time but also tests for the validity of the
common trend assumption.
In the interest of estimating the long-term e�ect on children, I focus on a sub-sample
of mothers to children born in either 1973 or 1976. For each of these cohorts, there is full
information on mothers' marital, welfare and labor market status when the children were
15. Those born in 1973 were 15 years of age in 1988, which is the �rst year of complete
data on maternal employment and welfare participation is available. These children were
not exposed to the reform during childhood, since they were already 19 when it started.
Children born in 1976, however, were 15 in 1991 and were exposed to the reform from
age 16 to 18. The following equation is used to estimate the intergenerational e�ects:
Yic = α + β ∗ Treatic ∗ Postc + γ ∗ Treatic +XicΦ+ εic (3)
where Yic is the outcome of child i in birth cohort c ; Treatic is a binary variable taking
the value one if the child's mother is not married when the child is 15 and zero otherwise;
Postt takes the value zero if the child was born in 1973, and the value one if the child
was born in 1976; Xit is a vector of children's and mothers' pre-reform characteristics:
child's gender, birth order, number of siblings and mother's quadratic in age, employment
and welfare status (these characteristics are all measured when the child is 15 and are
therefore pre-determined and una�ected by the reform); and εic are random error terms. β
is the coe�cient of interest, which captures the di�erence in outcomes between children
of both single and married mothers who were and were not exposed to the reform as
adolescents. An analogous speci�cation is used to estimate the e�ect of the reform on
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maternal outcomes, where Yic is the outcome of the mother of child i in birth cohort c
during the subsequent three years (i.e. when the child is between the ages of 16 and 18).
5 Contemporaneous e�ects
The labor market response of single mothers to the reform is key to understanding how the
reform may have a�ected the children. In this section, I estimate the contemporaneous
e�ect on single mothers' labor supply, welfare participation and incomes using the sample
of all non-ultra-Orthodox Jewish native-born mothers aged 21 to 54 between 1988 and
1995. In section 6, I also estimate these e�ects for a sub-sample of mothers which is used
to estimate the long-term impact on children.
5.1 Main results
Table 3 reports the di�erence-in-di�erences estimator for single mothers' labor market
outcomes. The mean of the dependent variable for single mothers during the pre-period
is reported alongside each outcome in order to assess the magnitude of the e�ects. The
reform led to a 4.7-percentage-point (34%) increase in welfare participation rates and
a 1.1 percentage-point (2%) decrease in employment rates. Annual welfare earnings
increased by 2,170 NIS (140%), re�ecting the increase in both welfare participation and
the generosity of bene�ts for single mothers. Annual earnings decreased by 1,850 NIS
(5%) on average. While there is no available data on hours worked, a simple back-of-the-
envelope calculation suggests that the decrease in earnings cannot be driven solely by the
decrease in employment. If we assume that those who stopped working had earned the
average wage among single working mothers, the decrease in employment can account for
no more than 42% of the decrease in annual earnings. This suggests that a substantial
part of the labor supply response was at the intensive margin.
Available data from the annual income surveys conducted by the Israeli Central Bu-
reau of Statistics provides additional suggestive evidence that single mothers decreased
their work hours following the 1992 reform. In 1990-1991, employed native-born Jewish
single mothers worked, on average, 37.1 (s.d=10.5) hours per week. In 1992-1993, their
average weekly working hours was 34.6 (s.d=11.7). At the same time, there was a only a
negligible change in working hours among married mothers.9
Looking at the e�ect on the probability of mothers being on welfare while being
employed, being on welfare without being employed and being employed without relying
on welfare reveals that the increase in welfare participation is associated with mothers
who combine work and welfare and that the reduction in employment while not being on
9The average number of weekly working hours among employed native-born Jewish married mothersincreased from 32.5 (s.d=10.5) to 33.0 (s.d=10.7) during those years.
10
welfare is larger than the general reduction in employment. This is expected since single
mothers could now earn up to 80% of the average wage while still being eligible for welfare.
The e�ect on the annual combined income is positive but insigni�cant. However, there is
a reason to believe that single mothers' discretionary income increased substantially as a
result of the valuable in-kind bene�ts provided to welfare-eligible single mothers following
the reform (on top of those provided both to eligible and non-eligible single mothers). A
conservative estimate for the value of these in-kind bene�ts provided to welfare-eligible
single mothers is 25,000 NIS a year.10 Furthermore, the reduction in work hours likely
reduced childcare expenditures since these women could now spend more time at home.
Figure 2 plots the main outcomes with 95% con�dence intervals strati�ed by the age
of the mother. We can see that the impact of the reform decreases with age and that
the e�ect on employment disappears for women over 40. This �nding may be a result
of individuals being less responsive to an increase in welfare generosity in later stages
of their careers. Figure 3 presents the estimates from the event-study speci�cation with
95% con�dence intervals for the following outcomes: welfare recipiency, annual welfare
bene�ts, employment, and annual labor earnings. It is evident that prior to the reform,
the di�erences in outcomes between single and married mothers were fairly constant and
only in 1992 does a clear gap emerge. The absence of major di�erences between the groups
during the pre-period supports the parallel-trends assumption underlying the model and
provides no evidence of any anticipatory behavior.
5.2 Robustness checks
The lack of di�erential pre-trends supports the validity of the identi�cation strategy in
the sense that native-born non-ultra-Orthodox Jewish mothers appear to have been on
parallel paths in terms of labor market behavior during the years leading to the reform.
However, the estimates may still be biased if the reform induced mothers to become or
stay single. I perform a series of tests to examine whether marital status was endogenously
a�ected. First, I plot the share of single mothers over time (see �gure 4). It is apparent
that the share of single mothers increased continuously during this period; however, there
appears to be no change in the slope after 1991. Second, in order to check whether or
not the results are driven by a change in the composition of single mothers over time, I
re-estimate equation (1) using a �xed sample of mothers composed of all mothers who
10This is calculated by estimating the average subsidy for each of the in-kind bene�ts provided towelfare-eligible single mothers using various secondary sources form di�erent government agencies overthe years, while taking into account the di�erence in real prices over time. For this purpose I assume thatthe average single mother rents an average sized-apartment and pays the average municipal property taxrate. When calculating the value of the child care and school fees subsidies, I assume the average singlemother has no more than a single child under the age of 11 and there is a 33% chance that that child isunder three years of age. I ignore the value of public housing that only a relatively small share of welfarerecipients get to enjoy.
11
appear both in the pre-period and the post-period while holding their marital status
�xed according to their pre-reform years. While some of the e�ects di�er in magnitude,
they are qualitatively similar: the reform led to a signi�cant increase in welfare bene�ts
received and a decrease in labor earnings (see appendix table A1). Finally, I estimate the
probability of marrying among those who were single mothers in 1988 and the probability
of divorce among those who were married mothers in 1988, using the following linear
probability model:
Yit = α + β ∗ Postt + γ ∗ Y eart +XicΦ+ εit (4)
The results in appendix table A2 show that conditional on age and a linear time
trend, single mothers are as likely to marry before the reform as after and that married
mothers are not more likely to divorce in the post-reform period. These �ndings are also
consistent with evidence from program evaluations in the US which show insigni�cant
e�ects of welfare reforms on family composition (Mo�tt, 2008).
6 Intergenerational e�ects
I now turn to estimating the intergenerational impact on children's welfare participation
and other economic outcomes. Since this analysis is performed on a subset of native-born
Jewish mothers, I start by examining the e�ect on of the reform on these mothers when
the children were between the ages 16-18. This is important in order to ensure that
these mothers were also responsive to the reform and that the response occurred when
their children were presumably still at home. The results are shown in table 5. Single
mothers of children born in 1976, as opposed to those of children born in 1973, were more
likely to receive welfare bene�ts but were as likely to work when their child was between
the ages of 16 and 18. The lack of labor supply response at the extensive margin is a
reasonable result if we recall that these are mothers of relatively older children. As �gure
2 showed, there appears to be no change in employment among mothers over 40 following
the reform. Nonetheless, we do see a marginally signi�cant decrease in earnings which is
likely to be the result of a reduction in work hours. The e�ect on the combined income
from welfare transfers and earnings is negative but insigni�cant. Recall however that
this measure does not capture the increase in in-kind bene�ts nor the possible increase
in leisure. Together with the fact that this reform could not possibly have made single
women worse o� by construction in the short run, then discretionary income most likely
went up.
12
6.1 Main results
Table 5 reports the long-term impact on children of single mothers who were exposed to
the reform between the ages of 16 and 18, measured at age 40. The increase in maternal
welfare participation rates had a large e�ect on the probability of ever being on welfare
(a 27% increase) and a corresponding increase in the number of cumulative months on
welfare and in the cumulative welfare bene�ts received throughout the period (49% and
67%, respectively). These e�ects imply either a change in norms or a transmission of
welfare-related information from parent to child. The fact that there is no change in
maternal employment at the extensive margin suggests that the increase in children's
future welfare participation is unlikely due lower parental labor force attachment.
Interestingly, the share of children who were never employed (or at least worked
without their employment being reported to the tax authorities) nor ever received welfare
bene�ts, decreased substantially. This decline actually suggests that the increase in
welfare participation did not come at the expanse of being employed and self-su�cient.
Rather, it came instead of being out of the formal labor force and out the the welfare
system. Not being employed by age 40 while never receiving any welfare aid is associated
with inferior outcomes in other areas. As the reported mean outcome indicates, 10%
of the children of single mothers born in 1973 are in that status. The data shows that
they are twice as likely to receive DI bene�ts at some point, fewer than 2% of have any
post-secondary education, only 30% ever marry, and less than 10% have any children.
There is also a notable increase in employment and life-time labor earnings. The
increase in labor earnings is larger than the increase in the number of months worked in
terms of percentage increase, suggesting that the reform led to an improvement in their
children's wages. This e�ect might be explained by the potential increase in parental
resources. An increase in discretionary income or an increase in time spent at home
may have been translated into increased investment in the children. A possible channel
for this to occur is an increase in educational attainment. We see some increase in the
probability of ever being enrolled in post-secondary education among those exposed to the
reform, but the e�ect is insigni�cant. This of course does not rule out an improvement in
educational outcomes earlier in life. Unfortunately, the data does not include information
on secondary education.
6.2 Falsi�cation tests and additional robustness checks
I perform several falsi�cation exercises to verify that the di�erence in children's outcomes
is not driven by something other than the additional welfare bene�ts provided to single
mothers during the 1990s. First, I check whether there is an impact on children who
grew up in high-income households, which should not be a�ected by the reform since
eligibility for welfare is means-tested. I group all children into 20,0000 NIS-sized pre-
13
reform earnings bins, where pre-reform earnings are calculated by dividing the total
annual household labor earnings when the child is 15 by the number of parents in order
to take into account the inherent di�erence in earnings between households with one
potential earner and those with two. I then re-estimate the "�rst stage" regressions
separately for each bin. The results are plotted in �gure 5 and show that the e�ect
on maternal outcomes disappears for household (normalized) earnings of 100,000 NIS or
more a year. Table 6 shows the children's outcomes in households with earnings below
the cuto� and in those with earnings above the cuto�. The results indicate that the
main e�ects are driven solely by the relatively low-income households and that children
in high-income households were not a�ected.
Next, I re-estimate the main speci�cations using children born in 1970 instead of 1976.
Those born in 1970 were already 22 when the reform was introduced and therefore there
should be no di�erence in outcomes between them and the 1973 cohort. I treat 1973 as
the treatment cohort and assume that the reform took place three years before it actually
did by setting the post dummy to one for the 1973 cohort and zero for the 1970 cohort.
Because there is no information on maternal employment and welfare participation before
1988, I do not control for them as in the main speci�cation and do not estimate the
equivalent "�rst stage" regressions. The results presented in appendix table A3 suggest
no signi�cant di�erences in children's long run outcomes.
An additional concern may be that some idiosyncratic di�erences between children
born in 1973 and 1976 to single or married mothers are driving the results. I make use of
the fact there were additional a�ected and una�ected cohorts in order to test this directly.
Speci�cally, I re-estimate the long run e�ects using children born between 1970 and 1977.
For each of these eight cohorts I can observe annual outcomes from age 21 to 40 and
have full information on their mothers' marital status when they were 15. This sample
consists of 192,804 children. Those born between 1970 and 1973 were not exposed to the
reform during childhood, while those born between 1974 and 1977 were exposed to the
reform for one to four years before reaching 19. I estimate an event study speci�cation
of the following form:
Yic = α + β ∗∑
Cohortc ∗ Treatic + γ ∗ Singleit + δc +XicΦ+ εic (5)
where Yic is the outcome of child i in birth cohort c ; Treatic is a binary variable taking
the value one if the child's mother is not married when the child is 15 and zero otherwise;∑Cohortc ∗Treatic is a vector of interactions of Treatic with cohort dummies, such that
the omitted cohort is 1973 (i.e. the last cohort not a�ected by the reform); Xic is a vector
of children's and mothers' characteristics measured when the child is 15: child's gender,
birth order, number of siblings and mother's quadratic in age.11; and εic are random
11In the absence of data on employment and welfare participation prior to 1988, maternal employment
14
error terms. The estimates are presented graphically in �gure 6. The results indicate
that there is no di�erence in the di�erences in long-term outcomes between children of
single mothers and married mothers among cohorts who were not exposed to the reform.
Furthermore, the long-term e�ects are rather similar across the exposed cohorts, implying
that the main results would have been qualitatively identical using one of the preceding
or following cohorts to 1976 instead of the 1976 cohort. In addition, the similarity in the
magnitude of the e�ects across the exposed cohorts suggests that longer exposure time
is not associated with better outcomes.
6.3 Heterogeneity and life-cycle e�ects
In an e�ort to shed light on the mechanisms underlying the main �ndings, I examine
the e�ects of the increase in single mothers' welfare bene�ts on the outcomes of di�erent
groups and over the life-cycle. I start by stratifying the sample by the median pre-reform
household earnings among the households with pre-reform annual normalized earnings of
under 100,000 NIS. Table 7 reports the e�ect on children's long-term outcomes separately
for children who grew up in households with very low earnings (under 27,000 NIS per
year) and those who grew up in households with higher but still relatively low earnings
(between 27,000 and 100,000 NIS per year). Appendix table A4 reports the corresponding
e�ects on the mothers when the children were between the ages 16-18. Maternal welfare
participation increased in both types of households, but the e�ects on children di�ered
signi�cantly. The results indicate that the entire e�ect on children's subsequent welfare
participation and the probability of ever being employed is driven by children who grew up
in very poor households. The e�ect on children's cumulative labor earnings, however, is
very similar across the two groups. This suggests that the intergenerational transmission
of welfare participation occurs only at the very bottom of the income distribution and
that there is no such transmission among higher-income welfare-eligible households. The
fact that the e�ect on earnings is present among both groups suggests that the potential
increase in discretionary income has improved the outcomes of children, irrespective of
whether or not they are also induced to rely on welfare in the future.
Table 8 reports the e�ect on children's long-term outcomes strati�ed by gender. The
results indicate that the increase in welfare participation is driven by girls, while the e�ect
on employment and labor earnings is largely driven by boys. Girls of single-mothers who
grew up in the post-reform period increase their duration of welfare use by 60%, and
accumulate 73% more welfare bene�ts between the ages 21-40 than girls of single mothers
who were born three years earlier. The equivalent e�ects on boys are insigni�cant. A
possible explanation may be that mothers serve as a more important role model for
daughters than for sons. Looking at the impact on employment and earnings, the e�ects
and welfare participation are not controlled for in this exercise.
15
are at best marginally signi�cant for girls while highly signi�cant for boys. In addition,
boys see larger gains both in absolute terms and relative to girls. This may be explained by
a larger investment in boys following an increase in parental resources or simply that boys
gain more from the same increase in resources. The latter possibility is consistent with
previous evidence suggesting that boys in single-parent families tend to receive less and
lower-quality parental inputs and have larger returns to parental investment than girls
(Bertrand and Pan, 2013), and that family disadvantage in general disproportionately
a�ects the behavioral and academic outcomes of boys relative to girls (Figlio et al.,
2019).
Finally, I examine the e�ects over the life cycle by separately estimating the e�ect on
outcomes between the ages of 21 and 30 and outcomes between the age of 31 and 40.
The results are shown in table 9 and reveal that the increase in welfare participation is
temporary while the gains in terms of employment and earnings are highly persistent.
Children who grew up in the post-reform period spent roughly twice as much time on
welfare in their twenties compared to children born in 1973. There was, however, no
increase in welfare participation rates when these children were in their thirties. Em-
ployment rates and cumulative labor earnings, on the other hand, increased by 5-6% and
10% respectively for both periods. This result highlights the fact that children's gains in
terms of future employment rates and earnings were long-lasting.
7 Summary
In this study, I estimate the long-term impacts of maternal welfare recipiency by ex-
amining the e�ects of an Israeli welfare reform designed to improve the situation of
single-parent households. Using a comprehensive administrative dataset, I �nd that the
reform led single mothers to work less and rely more on welfare in comparison to married
mothers. As a consequence, their children were signi�cantly more likely to rely on wel-
fare bene�ts themselves as young adults. The increase in welfare participation was not
permanent and was driven by girls, and especially among those who grew up in very low-
income households. At the same time, the increase in parental resources raised children's
employment and future labor earnings, an increase that was not limited to children who
grew up in very low-income households nor only to the period of young adulthood.
The �ndings imply that there is a causal transmission of welfare participation across
generations, which is driven by an information transmission mechanism and\or a social
norms mechanism. At the same time, the positive e�ects on employment and earnings
suggest that focusing only on welfare transmission may be misleading, and especially
so when only short observational windows are available, since the increase in welfare
generosity is found to also contribute to the long-term economic outcomes of the next
generation.
16
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18
Figure 1: The welfare payments schedule
Pre reform
010
2030
4050
6070
8090
100
Earn
ings
+ T
rans
fers
0 10 20 30 40 50 60 70 80 90 100Earnings
1992-1994
010
2030
4050
6070
8090
100
Earn
ings
+ T
rans
fers
0 10 20 30 40 50 60 70 80 90 100Earnings
Post 1994
010
2030
4050
6070
8090
100
Earn
ings
+ T
rans
fers
0 10 20 30 40 50 60 70 80 90 100Earnings
Notes: The bene�ts schedule for a single-parent family with one dependent child in terms of percent ofthe average wage.
19
Figure 2: Age heterogeneity
-.2-.1
5-.1
-.05
0.0
5.1
.15
.2
21 24 27 30 33 36 39 42 45 48 51 54age
Welfare
-.2-.1
5-.1
-.05
0.0
5.1
.15
.2
21 24 27 30 33 36 39 42 45 48 51 54age
Employment-1
0-5
05
1015
21 24 27 30 33 36 39 42 45 48 51 54age
Welfare benefits
-10
-50
510
15
21 24 27 30 33 36 39 42 45 48 51 54age
Earnings
Notes: The �gure shows the di�ernce-in-di�ernce estimates by age with a 95 percent con�dence interval.
20
Figure 3: Mothers outcomes
-.1-.0
8-.0
6-.0
4-.0
20
.02
.04
.06
.08
.1W
elfa
re p
artic
ipat
ion
1988 1989 1990 1991 1992 1993 1994 1995Year
-4-2
02
4W
elfa
re b
enef
its (i
n th
ousa
nds
of N
IS)
1989 1990 1991 1992 1993 1994 1995Year
-.1-.0
8-.0
6-.0
4-.0
20
.02
.04
.06
.08
.1E
mpl
oym
ent
1988 1989 1990 1991 1992 1993 1994 1995Year
-4-2
02
4E
arni
ngs
(in th
ousa
nds
of N
IS)
1988 1989 1990 1991 1992 1993 1994 1995Year
Notes: The �gure shows the estimated di�erences in outcomes over time between single mothers andmarried mothers with a 95 percent con�dence interval, in comparison to 1991, which is the omitted year.
21
Figure 4: Share of single mothers over time
0.0
5.1
.15
.2
1988 1989 1990 1991 1992 1993 1994 1995year
Notes: The Sample consists of all native-born Israeli mothers ages 21-54 of children under 18, between1988 and 1995.
Figure 5: Maternal outcomes when child is 18-18 by pre-reform houshold earnings
-.2-.1
5-.1
-.05
0.0
5.1
.15
.2W
elfa
re p
artic
ipat
ion
0 50 100 150 200 250 300Pre-reform household earnings
-.2-.1
5-.1
-.05
0.0
5.1
.15
.2E
mpl
oym
ent
0 50 100 150 200 250 300Pre-reform household earnings
Notes: The �gure shows the di�erence-in-di�erence estimates on maternal outcomes when the child isbetween 16 and 18 years of age, among mothers of children born in 1973 or 1976, by pre-reformearnings in 20K NIS (2010 prices) bins, with a 95 percent con�dence interval. Pre-reform earnings arede�ned as total annual household earnings divided by the number of parents when the child is 15.
22
Figure 6: Children outcomes - 1970-1977 cohorts
-2-1
01
23
Cum
ulat
ive
mon
ths
on w
elfa
re
1970 1971 1972 1973 1974 1975 1976 1977Year of birth
-50
510
Cum
ulat
ive
wel
fare
ben
efits
1970 1971 1972 1973 1974 1975 1976 1977Year of birth
-15
-10
-50
510
1520
Cum
ulat
ive
mon
ths
empl
oyed
1970 1971 1972 1973 1974 1975 1976 1977Year of birth
-200
-100
010
020
030
0C
umul
ativ
e ea
rnin
gs
1970 1971 1972 1973 1974 1975 1976 1977Year of birth
Notes: The �gure shows the estimated di�erences in outcomes across cohorts of children of singlemothers and married mothers with a 95 percent con�dence interval, in comparison to the di�ernce inoutcomes among the 1976 cohort, which is the last cohort unexposed to the reform during childhood.
23
Table 1: Mothers' characteristics in 1991
Marriedmothers
Single mothers
Age 34.625 36.696(6.920) (7.033)
Dependent children 2.301 1.674(1.096) (0.879)
Employed 0.640 0.563(0.480) (0.496)
Annual earnings36.374 36.497(49.427) (58.180)
Annual earningsconditional on working
60.015 67.742(56.542) (68.391)
Spousal earnings 102.017 -(138.459) -
On welfare 0.018 0.143(0.133) (0.350)
Annual welfare bene�ts 0.317 2.052(2.897) (5.958)
Observations 307,708 27,956
Notes: : Characteristics are measured in 1991, one year before thereform. Earnings and bene�ts are in thousand NIS (2010 prices).
24
Table 2: Pre-reform characteristics of mothers to children born in1973 and 1976
Marriedmothers
Single mothers
Age 40.013 40.111(4.052) (4.419)
Dependent children 1.945 1.196(1.182) (1.04)
Employed 0.610 0.602(0.488) (0.489)
Annual earnings 38.938 40.825(56.242) (57.842)
Spousal earnings 119.639 -(162.682) -
On welfare 0.010 0.096(0.101) (0.295)
Annual welfare bene�ts 0.140 0.733(1.940) (3.610)
Observations 48,608 4,806
Notes: Characteristics are measured the year the child turned 15.Earnings and bene�ts are in thousand NIS (2010 prices). Standarddeviations are reported in parentheses. :
25
Table 3: Mothers outcomes
Control mean E�ect
Welfare 0.137 0.047***(0.002)
Annual welfare bene�ts 1.538 2.168***(0.030)
Employment 0.546 -0.011***(0.002)
Annual earnings 35.251 -1.856***(0.254)
Annual combined income 36.789 0.312(0.252)
Decomposition of labor supply response:
Welfare with employment 0.076 0.025***(0.001)
Welfare w\o employment 0.062 0.022***(0.001)
Employment w\o welfare 0.484 -0.034***(0.002)
Observations 2,729,896
Notes: : The table reports the di�erence-in-di�erence estimates onmothers outcomes. The control mean is the mean of the dependentvariable for single mothers during the pre-reform period. Earningsand bene�ts are in thousand NIS (2010 prices). Standard errors clus-tered at the individual level in parentheses. * p<0.10, ** p<0.05,*** p<0.01
26
Table 4: Impact on mothers when child is 16-18
Control mean E�ect
Welfare 0.092 0.063***(0.008)
Months on welfare 1.640 1.771***(0.169)
Cumulative welfare bene�ts 2.870 3.603***(0.359)
Employment 0.669 0.003(0.010)
Months employed 18.686 -0.373(0.328)
Cumulative Earnings 132.244 -9.364*(4.784)
Cumulative combinedincome
135.102 -5.711(4.778)
Observations 53,414
Notes: : The table reports the di�erence-in-di�erence estimates formaternal outcomes when the child is between 16 and 18 years ofage, among mothers of children born in 1973 or 1976. The controlmean is the mean of the dependent variable for single mothers of chil-dren born in 1973. Earnings and bene�ts are in thousand NIS (2010prices). Heteroskedasticity-Robust Standard errors in parentheses. *p<0.10, ** p<0.05, *** p<0.01
27
Table 5: Children long-term outcomes
Control mean E�ect
Ever on welfare 0.092 0.025***(0.009)
Months on welfare 2.437 1.201***(0.455)
Cumulative welfare bene�ts 3.964 2.668***(0.880)
Never employed or onwelfare
0.108 -0.038***
(0.008)Months employed 125.866 7.482***
(2.201)Cumulative earnings 1,112.693 114.832***
(36.418)Cumulative combinedincome
1,116.657 114.832***(36.418)
Ever on disability 0.050 0.002(0.007)
Post-secondary education 0.436 0.014(0.014)
Observations 53.414
Notes: : The table reports the di�erence-in-di�erence estimates forchildren's outcomes by age 40. The control mean is the mean ofthe dependent variable for children of single mothers who were bornin 1973. Earnings and bene�ts are in thousand NIS (2010 prices).Heteroskedasticity-Robust Standard errors in parentheses. * p<0.10,** p<0.05, *** p<0.01
Table 6: Children long-term outcomes by pre-reform household earnings
Earning < 100K Earning > 100K
Control mean E�ect Control mean E�ect
Months on welfare 2.707 1.235** 0.433 0.152(0.523) (0.316)
Cumulative welfare bene�ts 4.405 2.754*** 0.693 0.721(1.008) (0.850)
Months employed 122.554 7.752*** 150.412 1.911(2.423) (5.245)
Cumulative Earnings 1050.830 118.496*** 1571.257 8.551(38.025) (116.593)
Observations 36,855 16,559
Notes: : The table reports the di�erence-in-di�erence estimates for children's long-term outcomes bypre-reform earnings. Pre-reform earnings are de�ned as total annual household earnings divided by thenumber of parents when the child is 15. The control mean is the mean of the dependent variable for chil-dren of single mothers who were born in 1973. Earnings and bene�ts are in thousand NIS (2010 prices).Heteroskedasticity-Robust Standard errors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
28
Table 7: Children long-term outcomes by pre-reform household earnings - cont.
Earnings < 28K 28K < Earning <100K
Control mean E�ect Control mean E�ect
Months on welfare 2.965 1.737** 2.214 0.319(0.717) (0.739)
Cumulative welfare bene�ts 4.677 4.028*** 3.887 0.581(1.380) (1.425)
Months employed 111.083 8.429*** 144.483 5.371(3.255) (3.540)
Cumulative Earnings 925.807 90.120** 1289.825 133.857*(44.838) (68.430)
Observations 18,428 18,427
Notes: : The table reports the di�erence-in-di�erence estimates for children's long-term outcomes by pre-reform earnings in 20K NIS bins. Pre-reform earnings are de�ned as total annual household earnings di-vided by the number of parents when the child is 15. The control mean is the mean of the dependentvariable for children single mothers who were born in 1973. Earnings and bene�ts are in thousand NIS(2010 prices). Heteroskedasticity-Robust Standard errors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
Table 8: Children long-term outcomes strati�ed by gender
Girls Boys
Control mean E�ect Control mean E�ect
Months on welfare 3.114 1.854** 1.814 0.476(0.787) (0.479)
Cumulative welfare bene�ts 5.660 4.123** 2.402 1.031(1.641) (0.709)
Months employed 131.027 6.014* 121.112 8.874***(3.097) (3.124)
Cumulative Earnings 973.566 49.204 1240.834 161.055***(40.361) (59.112)
Observations 25,721 27,693
Notes: : The table reports the di�erence-in-di�erence estimates for children's long-term outcomes by gen-der. The control mean is the mean of the dependent variable for children of single mothers who were bornin 1973. Earnings and bene�ts are in thousand NIS (2010 prices). Heteroskedasticity-Robust Standarderrors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
29
Table 9: Impact across the life cycle
Between 21 and 30 Between 31 and 40
Control mean E�ect Control mean E�ect
Months on welfare 0.878 0.942*** 1.559 0.259(0.228) (0.295)
Cumulative welfare bene�ts 1.728 1.771*** 2.236 0.897*(0.505) (0.511)
Months employed 56.119 2.715*** 69.746 4.767***(0.967) (1.488)
Cumulative Earnings 346.148 34.956*** 766.545 79.876***(9.629) (30.193)
Observations 53,414 53,414
Notes: : The table reports the di�erence-in-di�erence estimates for children's outcomes during their twen-ties and thirties. The control mean is the mean of the dependent variable for children single mothers whowere born in 1973. Earnings and bene�ts are in thousand NIS (2010 prices). Heteroskedasticity-RobustStandard errors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
30
8 Appendix
Table A1: Mothers outcomes - �xed sample
Control mean E�ect
Welfare 0.137 0.037***(0.002)
Annual welfare bene�ts 1.539 1.874***(0.031)
Employment 0.546 0.008***(0.003)
Annual earnings 35.251 -1.150***(0.264)
Annual combined income 36.790 0.789***(0.261)
Decomposition of labor supply response:
Welfare with employment 0.062 0.019***(0.001)
Welfare w\o employment 0.076 0.018***(0.001)
Employment w\o welfare 0.484 -0.011***(0.002)
Observations 2,551,569
Notes: : The table reports the di�erence-in-di�erence estimates onmothers outcomes when the sample is restricted to all mothers tochildren under 18 who appear both in the pre-period and the post-period, while holding thier marital status �xed according to the lastpre-reform year. The control mean is the mean of the dependentvariable for single mothers during the pre-reform period. Earningsand bene�ts are in thousand NIS (2010 prices). Standard errors clus-tered at the individual level in parentheses. * p<0.10, ** p<0.05,*** p<0.01
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Table A2: Selection into single-motherhood
Divorce Marry
Post -0.001** -0.004(0.0005) (0.003)
Linear time trend ! !
Age controls ! !
Observations 1,933,432 122,904
Notes: : Column 1 reports the estimated probability a marriedmother in 1988 will divorce in the years following the reform. Col-umn 2 reports the estimated probability a single mother in 1988 willmarry in the years following the reform. Standard errors clustered atthe individual level in parentheses. * p<0.10, ** p<0.05, *** p<0.01
Table A3: Children long-term outcomes - 1970 vs. 1973 cohorts
Control mean E�ect
Months on welfare 2.492 -0.419(0.494)
Cumulative welfare bene�ts 4.588 -1.349(0.995)
Months employed 115.690 3.944(2.794)
Cumulative earnings 1103.727 67.285(46.561)
Observations 37,801
Notes: : The table reports the di�erence-in-di�erence estimates forchildren's outcomes by age 40 among children born in 1970 and 1973.The control mean is the mean of the dependent variable for childrenof single mothers who were born in 1970. Earnings and bene�ts arein thousand NIS (2010 prices). Heteroskedasticity-Robust Standarderrors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
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Table A4: Impact on mothers when child is 16-18 by pre-reform household earnings
Earnings < 28K 28K < Earning <100K
Control mean E�ect Control mean E�ect
Welfare 0.140 0.073*** 0.034 0.056***(0.012) (0.011)
Months on welfare 2.539 2.317*** 0.556 1.184***(0.267) (0.234)
Welfare bene�ts 4.546 5.249*** 0.747 1.992***(0.572) (0.413)
Employment 0.438 -0.017 0.980 0.014*(0.018) (0.008)
Months employed 9.163 -1.081** 30.939 0.914*(0.477) (0.479)
Earnings 43.195 -11.298*** 184.222 3.285(3.312) (5.030)
Combined income 47.740 -6.048* 184.970 5.277(3.328) (4.984)
Observations 18,428 18,427
Notes: : The table reports the di�erence-in-di�erence estimates for maternal outcomes when the child isbetween 16 and 18 years of age, among mothers of children born in 1973 or 1976, by pre-reform earnings(2010 prices). Pre-reform earnings are de�ned as total annual household earnings divided by the number ofparents when the child is 15. The control mean is the mean of the dependent variable for single mothers ofchildren born in 1973. Earnings and bene�ts are in thousand NIS (2010 prices). Heteroskedasticity-RobustStandard errors in parentheses. * p<0.10, ** p<0.05, *** p<0.01
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