The intergenerational e�ects of welfare participation∗

Click here for most recent version

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: yannaysh@mail.tau.ac.il

1

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.

2

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-

3

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).

4

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.

5

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.

6

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.

7

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:

8

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

9

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

References

Antel, J. J. (1992): �The intergenerational transfer of welfare dependency: Some sta-

tistical evidence,� The Review of Economics and Statistics, 467�473.

Beaulieu, N., J.-Y. Duclos, B. Fortin, and M. Rouleau (2005): �Intergener-

ational reliance on social assistance: Evidence from Canada,� Journal of Population

Economics, 18, 539�562.

Bertrand, M. and J. Pan (2013): �The trouble with boys: Social in�uences and the

gender gap in disruptive behavior,� American Economic Journal: Applied Economics,

5, 32�64.

Black, S. E., P. J. Devereux, et al. (2011): �Recent Developments in Intergenera-

tional Mobility,� Handbook of Labor Economics, 4, 1487�1541.

Chen, K., L. Osberg, and S. Phipps (2015): �Inter-generational e�ects of disability

bene�ts: evidence from Canadian social assistance programs,� Journal of Population

Economics, 28, 873�910.

Chetty, R., J. N. Friedman, and J. Rockoff (2011): �New evidence on the long-

term impacts of tax credits,� IRS Statistics of Income White Paper.

Dahl, G. B. and A. C. Gielen (2018): �Intergenerational spillovers in disability in-

surance,� Tech. rep., National Bureau of Economic Research.

Dahl, G. B., A. R. Kostøl, and M. Mogstad (2014): �Family welfare cultures,�

The Quarterly Journal of Economics, 129, 1711�1752.

Dahl, G. B. and L. Lochner (2012): �The impact of family income on child achieve-

ment: Evidence from the earned income tax credit,� American Economic Review, 102,

1927�56.

��� (2017): �The impact of family income on child achievement: Evidence from the

earned income tax credit: Reply,� American Economic Review, 107, 629�31.

Figlio, D., K. Karbownik, J. Roth, M. Wasserman, et al. (2019): �Family

disadvantage and the gender gap in behavioral and educational outcomes,� American

Economic Journal: Applied Economics, 11, 338�81.

Flug, K. and N. Kasir (2006): �The single parent law, labor supply and poverty,�

Israel Economic Review, 4, 59�110.

Frish, R. and N. Zussman (2008): �The e�ect of transfer payments on the labor supply

of single mothers,� The Journal of Socio-Economics, 37, 627�643.

17

Gottschalk, P. (1990): �AFDC participation across generations,� American Economic

Review, 80, 367�71.

��� (1996): �Is the correlation in welfare participation across generations spurious?�

Journal of Public Economics, 63, 1�25.

Hartley, R. P., C. Lamarche, J. P. Ziliak, et al. (2017): �Welfare Reform and

the Intergenerational Transmission of Dependence,� Tech. rep., Institute for the Study

of Labor (IZA).

Levine, P. B. and D. J. Zimmerman (1996): The Intergenerational Correlation in

AFDC Participation: Welfare Trap or Poverty Trap?, University of Wisconsin-Madison,

Institute for Research on Poverty.

Løken, K. V., K. E. Lommerud, and K. H. Reiso (2018): �Single mothers and their

children: Evaluating a work-encouraging welfare reform,� Journal of Public Economics,

167, 1�20.

Meyer, B. D., W. K. Mok, and J. X. Sullivan (2009): �The under-reporting of

transfers in household surveys: its nature and consequences,� Tech. rep., National

Bureau of Economic Research.

Milligan, K. and M. Stabile (2011): �Do child tax bene�ts a�ect the well-being

of children? Evidence from Canadian child bene�t expansions,� American Economic

Journal: Economic Policy, 3, 175�205.

Moffitt, R. (2008): �Welfare reform: The US experience,� Tech. rep., Working

paper//IFAU-Institute for Labour Market Policy Evaluation.

Page, M. E. (2004): �10 New evidence on the inter generational correlation in welfare

participation,� Generational income mobility in North America and Europe, 226.

Pepper, J. V. (2000): �The intergenerational transmission of welfare receipt: A non-

parametric bounds analysis,� Review of Economics and Statistics, 82, 472�488.

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

31

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

32

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

33