DISCUSSION PAPER SERIES
IZA DP No. 15654
Andrew E. Clark
Conchita D’Ambrosio
Anthony Lepinteur
Marriage as Insurance:
Job Protection and Job Insecurity
in France
OCTOBER 2022
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DISCUSSION PAPER SERIES
ISSN: 2365-9793
IZA DP No. 15654
Marriage as Insurance:
Job Protection and Job Insecurity
in France
OCTOBER 2022
Andrew E. Clark
Paris School of Economics, University of Luxembourg and IZA
Conchita D’Ambrosio
University of Luxembourg
Anthony Lepinteur
University of Luxembourg and IZA
ABSTRACT
IZA DP No. 15654 OCTOBER 2022
Marriage as Insurance:
Job Protection and Job Insecurity
in France
*
Job insecurity is one of the risks that workers face on the labour market. As with any risk,
individuals can choose to insure against it, and we here consider marriage as one potential
source of this insurance. The 1999 rise in the French Delalande tax, paid by larger private
firms when they laid off workers aged 50 or over, led to an exogenous rise in job insecurity
for the uncovered (younger workers) in these larger firms. A difference-in-differences
analysis using French panel data reveals that this greater job insecurity for the under-50s
led to a significant rise in their probability of marriage, and especially when the partner had
greater job security, consistent with marriage providing insurance against labour-market
risk.
JEL Classification: I38, J13, J18
Keywords: marriage, insurance, employment protection, difference-in-
differences
Corresponding author:
Anthony Lepinteur
University of Luxembourg
2 Av. de l’Universite
4365 Esch-sur-Alzette
Luxembourg
E-mail: [email protected]
* We are grateful to the Editor and two anonymous referees for making good points that helped us to develop our
analysis. We would also like to thank Petri Böckerman, Arnaud Chevalier, Thomas Dohmen, Markus Gebauer, Kevin
Lang, Marion Leturcq, Giorgia Menta, Anne Solaz and seminar participants at the IZA/HSE seminar for their help and
feedback. We acknowledge financial support from CEPREMAP, the Fonds National de la Recherche Luxembourg
(Grant C18/SC/12677653) and EUR grant ANR-17-EURE-0001. Declaration of interest: none.
1. Introduction
Why do people get married? Weiss (1997) suggests that marriage comes with a number of
economic advantages. The first reflects the benefits of specialisation between spouses (Becker,
1973 and 1981). Second, in a context of imperfect credit markets, marriage may also relax
credit constraints via implicit credit arrangements within households (Borenstein and Courant,
1989) and enhance investment (for example, one partner may work while the other is in
education investing in their human capital). Collective and non-rival goods are also jointly
produced and consumed within partnerships, with common examples being children or
housework (Chiappori, 1992, and Van Klaveren et al., 2008). Stevenson and Wolfers (2007),
in their empirical review of the changing trends in marriage and divorce in the US, highlight
the roles of pre-marital cohabitation (which has risen), specialisation in marriage (now argued
to be less important), the tax implications of partnership, birth control, changes in relative
wages, Divorce Laws, and the marriage “matching function” (via education, the workplace and
the internet).
The benefit from marriage that we will address here, which appears in both Weiss (1997)
and Stevenson and Wolfers (2007), is that of risk-sharing. As noted by Hess (2004) and Shore
(2010), partnerships provide insurance by allowing couples to diversify risk, as long as the
exogenous income shocks of the two partners are not perfectly (positively) correlated. Couples
can in addition adjust their relative labour supply to reduce the impact of shocks.
We here consider labour-market risk, and focus on the role of job loss in shaping marriage
formation. We appeal to an exogenous variation in job insecurity at the individual level
resulting from a French quasi-natural experiment: the 1999 increase in the Delalande tax.
Introduced in 1987, this tax had to be paid by firms that laid off older workers. From January
1999 up to its abolition in 2008 the Delalande tax rose, but only for firms with 50 or more
employees. Using difference-in-differences regressions exploiting this firm-size discontinuity
1
and the perverse effects of the reform on the separation rate of younger workers, we estimate
the causal impact of job insecurity on the probability of marriage.
As in Clark and Lepinteur (2022), our empirical analysis is based on the French component
of the European Household Community Panel (ECHP). We first show that the 1999 increase
in the Delalande tax in large firms produced greater feelings of job insecurity for the under-
50s. We then consider the changes in marital status of these newly-insecure workers, as
compared to a control group of under-50s in smaller firms (for whom the reform did not apply).
We are interested in the way in which this exogenous greater risk affects behaviour. Clark and
Lepinteur (2022) suggest that one reaction to an exogenous change in job insecurity is to reduce
risk exposure, via lower fertility; we instead here ask whether individuals will take out more
insurance against risk, via their marital status.
1
We will restrict our analysis here to younger
workers, as most respondents aged 50 or more in our sample were already married before the
reform, with only little post-reform movement out of marriage.
We conclude that greater job insecurity amongst French workers increased women’s
probability of marriage by four percentage points; there is no effect for men. This is consistent
with evidence on gender differences in preferences (Croson and Gneezy, 2009): women are
generally less willing to take risks in the context of lotteries (Hartog et al., 2002, Holt and
Laury, 2002, and Fehr-Duda et al., 2006) and portfolio selection (Sunden and Surette, 1998,
Finucane et al., 2000, and Charness and Gneezy, 2012). Falk et al. (2018) measure risk aversion
via both a self-report (“In general, how willing are you to take risks?”) and revealed preference
from a series of binary choices between a lottery and a sure return. Women were found to be
more risk-averse than men in almost all of the 76 countries for which they collected data (see
their Figure III). As such, women may react more to the threat of future job loss than do men.
1
The analysis in Clark and Lepinteur (2022) does not use the same sample as we do here. Their fertility analysis
is carried out on the sample of workers who were already married before the reform and continued to be married
after it. On the contrary, we explicitly model moves between marital statuses.
2
The finding of no change in the marriage probability of men with greater job insecurity does
necessarily imply that they would not like to obtain insurance through marriage. However,
being more at risk of layoff might have made them “riskier” partners.
We check that our estimates are robust to a number of possible confounding events,
including European macroeconomic trends and the reduction in the French working week to
35 hours that was announced in 1998 and implemented two years later in 2000; they are also
qualitatively similar using a number of different estimation methods. While we do identify an
insecurity effect on marriage, there is no change in the probability of entering a partnership in
general, or indeed of leaving one: the greater probability of marriage for women then mostly
reflects a shift into marriage from pre-reform cohabitation. The effect of job insecurity on
women’s marriage probability is similar by age, education and wages, but larger for women
who were already mothers before the reform. Last, as predicted by risk-sharing, the probability
of marriage only rises when the partner is employed and did not experience greater job
insecurity due to the layoff-tax rise.
The remainder of the paper is structured as follows. Some existing work on marriage and
insurance is discussed in Section 2, and the institutional background and our empirical
approach to the analysis of individual-level insecurity appear in Section 3. Section 4 then
presents the ECHP data and the estimation sample. Section 5 displays the main results, while
the robustness tests and heterogeneity analyses appear in Section 6. The importance of partner
characteristics in the light of risk-sharing theory is then discussed in Section 7. Last, Section 8
concludes.
2. Marriage as Insurance: Background and Contribution to the Literature
A number of contributions have provided indirect evidence of an insurance role of
marriage. In Rosenzweig and Stark (1989), arranged marriages between families from different
3
villages in South India significantly reduced the variability in food consumption. In addition,
farm households subject to greater income risks were more likely to engage in arranged
marriages at longer distances. In US data, Halla and Scharler (2011) find that the influence of
idiosyncratic output-growth shocks on consumption is smaller in States where the percentage
of married is higher. Bertocchi et al. (2011) consider household investment decisions, and
conclude from their analysis of 14 years of Italian SHIW data that the married invest more in
risky assets (as marriage is considered to be a safe asset). Anderson and Ray (2019) show that
marriage protects against the risk of death, especially for women. In a similar vein, Van den
Berg and Gupta (2015) use individual data from Dutch registers (from 1815 to 2000) and find
a protective effect of marriage against mortality for men.
With respect to the labour market, there is a considerable literature on marriage and the
business cycle. Schaller (2013) analyses 32 years of US State-level panel data: marriage is
shown to be pro-cyclical, with unemployment being associated with less marriage. Lichter et
al. (2006) appeal to individual-level NLSY-79 data, and conclude that the probability of
transition from cohabitation to marriage rises with partner’s education and the partner working;
it is lower for the unemployed. Education and employment are found to have a similar influence
on marriage probability in Chinese data (Yu and Xie, 2015). Early labour-market experiences
also seem to matter. Ekert-Jaffé and Solaz (2001 and 2002) and Landaud (2021) analyse
different French datasets to show that early-career unemployment and temporary jobs reduce
the probability of forming a couple; De La Rica and Iza (2005) come to similar conclusions
using Spanish data. One interpretation is that unemployment provides a negative signal about
the potential partner’s unobserved characteristics. Consistent with this interpretation, in
Charles and Stephens (2004), job loss increases the risk of divorce when resulting from layoff
but not as a result of plant closings: they note that the former may convey information about
the partner’s non-economic suitability as a mate (via, for example, their temperament). Similar
4
results are also found in other disciplines such as Sociology and Demography (Oppenheimer,
1988, Kalmijn, 2011, de Lange et al., 2014, and van Wijk et al., 2021)
Some work has explicitly looked not at events that have already occurred, but rather
measures of the future risk that individuals face on the labour market. Schneider et al. (2018)
discuss the role of economic resources in marriage, which they extend to include wealth and
expected future earnings. Schneider and Reich (2018) continue in the same line, and find that
union membership (as an indicator of economic security) is a predictor of marriage in NLSY-
79 data. Xie et al. (2003) also adopt an individual-level and forward-looking approach to
economic insecurity, and calculate five different measures of current and future “earnings
potential”: they show that all five are positively correlated with the transition from cohabitation
to marriage using US Census and cohort data.
Although the role of forward-looking job insecurity in partnership formation has already
been mentioned in the literature (Schneider et al., 2018, and Bolano and Vignoli, 2021), we are
not aware of any work that has been able to appeal to exogenous variation in job insecurity at
the individual level in this context. Using difference-in-differences regressions based on a
quasi-natural experiment, we here propose to estimate the causal impact of job insecurity on
the probability of marriage. As opposed to past individual unemployment or current
macroeconomic shocks, we are able to identify a plausible exogenous change in job insecurity
following a French labour-market reform that put some workers more at risk than others of
future job loss. Our findings confirm the predictions of risk-sharing theory: greater job
insecurity leads to an increase in the probability of marriage for those who are arguably more
risk-averse (women with children). In other words, marriage can be used as an insurance in
moments of economic insecurity.
5
3. Institutional Background and Empirical Approach
a. The Advantages of Marriage in France
Marriage in France brings a number of financial advantages, some of which are
particularly attractive for workers facing economic insecurity. First, French legislation
considers cohabiting couples as two separate tax units whereas married couples count as a
single tax unit. As shown in Leturcq (2012), the formula used by the French government to
calculate annual income tax in the early 2000s produced a considerably-lower income tax rate
for the married as compared to the cohabiting. These gains from marriage are particularly high
when the income gap between the spouses is large. According to Echevin (2003), around 50%
of married couples in 1999 benefited from a lower yearly income tax bill of at least 1000 Euros.
Buffeteau and Echevin (2004) provide additional simulations of the financial gains from
marriage in France in the early 2000s.
There are in addition a number of marital-property regimes in France. Since 1965, the
standard regime is the “regime de communauté de biens réduite aux acquêts” (Frémeaux and
Leturcq, 2013), in which any property owned by one of the spouses before marriage continued
to be treated as belonging only to that spouse in the case of divorce. On the contrary, all
property, assets and income acquired during the marriage were treated as common property
and were thus subject to division in the case of divorce. Although the share of married couples
opting for alternative regimes with greater individual control over the resources accumulated
during marriage increased from 1975 to 2010, over 82% of newly-wedded couple were still
under the “regime de communauté de biens réduite aux acquêts” that provides the greatest level
of insurance between partners during our period of interest, that is the end of the 1990s
(Frémeaux and Leturcq, 2018). Consequently, we can arguably consider marriage to be an
insurance mechanism in most cases in France during this period.
Article 212 of the French Civil Code states that “les époux se doivent mutuellement […]
secours”. The “devoir de secours” (obligation of assistance) rarely takes the form of a legal
6
and formal transfer of resources during marriage; it is rather supposed to come about naturally
via resource-pooling and solidarity between the two spouses. However, in the case of divorce
or even during the divorce proceedings, the spouse with the most-favourable financial situation
can be asked to transfer resources to the other spouse.
The social-security and health-insurance eligibility conditions for the unemployed with an
employed partner are less restrictive when the relationship is formalised by marriage. In the
case of the death of a partner, marriage also simplifies the inheritance procedures for the
surviving spouse and gives the survivor the right to receive a part of the pension of the dead
spouse (“pension de reversion”).
Finally, married couples must legally support each other even after a divorce. A judge may
decide that a compensatory allowance should be paid to compensate for differences in living
standards between the newly-separated spouses. About 100,000 people received such a benefit
at the end of the 1990s, with a monthly transfer of around 2,000 Francs (306 Euros).
2
The
obligation to provide child support also implies the payment alimony to the spouse with
custody of the children – with the amount of the alimony being determined either by the parents
or a judge. Given that women have lower incomes than men on average, and that children were
in the sole custody of the mother in 80% of divorce cases in the late 1990s in France,
3
the
various pensions paid in the event of divorce can also be seen as a source of marriage-related
insurance for women.
b. The unemployment benefit system in France at the end of the 1990s
The French unemployment insurance scheme provided two benefits at the end of the
1990s: the “allocation unique dégressive” (AUD) et “l’allocation de chômeurs âgés” (ACA).
The former varies by age the time spent in employment. The eligible unemployed need to have
2
See more details in http://www.justice.gouv.fr/art_pix/rapport-prest-compens.pdf.
3
See Table 2 in http://www.justice.gouv.fr/art_pix/stat_Infostat%20132%20def.pdf.
7
spent at least four months in employment out of the last eight. The duration of AUD benefits
ranges from four months to five years. The AUD replacement rate changes over time: it is at
its maximum level during the first months of unemployment and then falls. Only workers who
had worked for overs 160 trimesters were eligible for the ACA for a limited amount of time, at
the full ACA replacement rate. At the end of 2000, average monthly unemployment benefits
were 5,202 Francs, for an average replacement rate of 68%. Note that unemployment benefits
do not depend on marital status. See https://travail-
emploi.gouv.fr/IMG/pdf/publication_pips_200111_n-46-1_indemnisation-chomage-99-
2000.pdf for a complete description of the French unemployment benefit system.
c. The Delalande Tax
The French government introduced the Delalande tax in 1987, with the goal of tackling
rising layoffs among older workers. While there were a number of modifications between its
implementation and its final abolition in 2008, the principle of the experience-rating of the tax
did not change: firms that laid off older workers (where the definition of “old” has changed
over time) were required to pay the Delalande tax to help balance the unemployment-insurance
system. This tax was proportional to the gross wage of the laid-off worker, and was applied to
private-sector permanent-contract workers. From 1987 to 1992, the amount of tax due
following the layoff of a worker aged 55 or over was equal to three months of gross wages.
The tax rules were revised in July 1992, January 1993 and January 1999; in particular, starting
in 1992 this layoff tax was extended to all workers aged 50 or above.
Table 1 reports the different schemes of the Delalande tax over time according to firm size
and the age of the laid-off worker. From 1993 to December 1998, the tax to be paid only
depended on worker age, and was independent of firm size. From January 1999 up to its
abolition in 2008 the Delalande tax rose, but only for firms with 50 or more employees. This
tax represents a non-trivial part of the total separation costs: in Behaghel et al. (2004) the
8
average separation costs for older workers in France ranged from 3 to 11 months of gross
wages, with over half of this cost reflecting the payment of the Delalande tax.
The net effect of the Delalande contribution on employment is theoretically ambiguous.
First, a company looking for a new employee might prefer to avoid the anticipated extra-cost
of a dismissal due to the Delalande tax by hiring a young worker. This perverse effect was
identified and then corrected with the 1992 reform which introduced an exemption from
payment of the tax for all unemployed people over age 50 who find a job (after at least three
months of unemployment). Behaghel et al. (2004) use this exogenous change to show that the
hiring of older workers was indeed reduced before 1992 due to this tax.
Second, a higher Delalande tax should reduce the separation rate of older workers.
Behaghel et al. (2004) also address this issue, and show that while in general the increase in
the Delalande tax over time had only modest effects on this separation rate, the 1999 reform
(which we analyse here) did have a greater effect in protecting older workers against
separations.
Last, the theoretical model in Behaghel (2007) predicts that a higher Delalande tax has
another perverse effect: it should also increase the separation rate of younger workers. The
results in Georgieff and Lepinteur (2018) are consistent with these predictions. In data from
the French component of the ECHP, the perceived job insecurity of younger workers in larger
firms rose due to the perverse effects of the higher Delalande tax; and data from the French
Labour Force Survey reveal an increase in the actual risk of layoff for the same group.
We here aim to establish the causal impact of this exogenous variation in job insecurity
resulting from the 1999 reform on the marriage probability of younger workers. We do so by
exploiting the firm-size discontinuity and the resulting increase in job insecurity for younger
workers in treated firms. The 1999 rise in the Delalande tax provides a natural quasi-experiment
allowing for a difference-in-differences (D-i-D) analysis, where the treatment group is younger
9
workers (under age 50) in large private firms (whose job insecurity changed) and the control
group younger workers in smaller private firms (who were not affected by the reform).
As the 1999 reform of the Delalande tax was announced by the French government one
year beforehand (in early 1998), and the reintroduction of the firm-size discontinuity was also
public knowledge by the end of 1998, employers may have strategically adjusted their labour
demand before the official change in the Law. There is evidence of such anticipation effects in
Georgieff and Lepinteur (2018), as employers in large firms dismissed relatively-more
employees between the announcement and the implementation of the reform. Higher layoffs
are found for both those aged over 50 and younger workers, so that the tax rise brought about
some restructuring in the treated firms.
4
Given this anticipation, job insecurity for the younger
workers in large firms may have risen before the reform’s implementation. We take this
possibility into account by estimating the following D-i-D equation, including both a post-1999
treatment effect and a 1998 announcement effect of the reform:
(1)
This equation is estimated only for workers under the age of 50. Here
is either the
perceived job security of worker i, or a dummy variable for worker i being married in year t.
5
Treat
it
indicates whether a worker belongs to the treatment group: it is equal to one for younger
private-sector workers in large firms (50+ employees) and to zero for younger private-sector
workers in firms with fewer employees. The Post1998
t
dummy captures observations on
individuals after the implementation of the higher Delalande tax (in January 1999), λ
t
are year
fixed effects (so that λ
1998
covers observations in the policy-announcement year of 1998) and
4
One aim of this restructuring may have been to avoid the tax by having fewer than 50 employees, as in Garicano
et al. (2016). We will address this selection into firm size issue below by dropping workers who reported that
their firms’ size changed in any year after the date at which the reform was announced.
5
We also consider in Appendix Table A2 the following complete set of marital outcomes: partnered, partnered
but not married, divorced or separated, and never in a relationship.
10
X
it
is a set of individual characteristics. This latter contains the following variables: age
dummies (in five-year bands), health status, the lagged number of children in the household,
6
the monthly wage (logged), weekly working hours, and occupation and region fixed effects.
We also account for time-invariant heterogeneity by controlling for individual fixed effects µ
i
.
The main effects of λ
1998
and Post1998
t
are entirely subsumed by the year dummies. The
coefficients of interest in the equation above are α
2
and α
3
: these respectively capture the
impacts of the reform’s announcement in 1998 and its implementation from 1999 onwards on
first job security and then marital status.
The risk-sharing hypothesis predicts positive and significant α
2
and α
3
coefficients for
workers in the marriage-probability regression, reflecting their greater risk from higher job
insecurity and the subsequent rise in the demand for insurance. However, it is also possible that
the treatment reduces the probability of marriage, as workers who have become more at risk of
layoff are now seen to be “riskier” partners.
The effects of the announcement and the implementation of the reform are expected to be
of the same sign. As such, we also estimate the simpler D-i-D regression below:
(2)
where Post1997
t
is a dummy variable reflecting observations after the announcement year of
the higher layoff tax (from 1998 onwards). The coefficient β
2
in Equation (2) now captures
what we will consider as the “total effect” of the reform, starting from its announcement. As
such, β
2
corresponds to equation (1) when α
2
= α
3
. Both Equations (1) and (2) will be estimated
6
It is not clear whether children should be included in the list of control variables. Marriage may encourage
parenthood, or alternatively those who already have children may be more likely to marry. Including the lagged
number of children in the household as a control somewhat attenuates concerns about reverse causality. The effect
of job insecurity on marriage turns out to change only little when we do not control for the lagged children variable.
11
using linear techniques (although we will also estimate non-linear regressions in the robustness
checks).
4. The ECHP Data and the Estimation Sample
Our empirical analysis is based on the European Community Household Panel (ECHP).
This is a nationally-representative longitudinal survey of households that covered 15 European
countries (including France) between 1994 and 2001. The sample size in France was on average
12,000 adults per wave. The interviews mainly took place at the end of the year (between
November and December). The ECHP data includes a large set of variables on a variety of
aspects of respondents’ lives, including demographic characteristics, financial situation,
employment, and health.
7
Respondents report their marital status in each ECHP wave.
The 1999 increase in the Delalande tax only applied to private-sector firms with 50 or more
employees. The ECHP records the number of workers in the firm in which the respondent is
employed using the following categories: “None”, “1 to 4”, “5 to 19”, “20 to 49”, “50 to 99”,
“100 to 499” and “500 or more”. This is the variable that we will use to assign younger workers
to the treatment or control groups.
In the first part of our analysis, we show that the higher layoff tax for those aged 50 or
over in larger firms produced greater feelings of job insecurity for individuals aged under 50
working in the same firms; the main analysis will then address the question of whether this
same tax rise affected marriage.
Our measure of job insecurity comes from the following ECHP question:
“How satisfied are you with your present job in terms of job security?”
7
For more details about the dataset, see http://ec.europa.eu/eurostat/web/microdata/european-community-
household-panel.
12
The answers to this satisfaction question were given on a scale from 1 to 6, with 1
corresponding to “Not Satisfied” and 6 to “Fully Satisfied”.
8
Böckerman et al. (2011) show
that this measure of perceived job security reflects objective variations in layoff and hiring
rates. It is also known to be a strong predictor of individual choices on the labour market such
as job quits (Clark, 2001).
9
In Clark and Postel-Vinay (2009), it is also correlated with
unemployment-insurance benefits, permanent contracts and employment-protection
legislation.
Our analysis sample is made up of private-sector workers between the ages of 20 and 49
with permanent contracts, and with valid information on perceived job security, job
characteristics and the sociodemographic variables. As the details of the reform were public
knowledge before its implementation, employers may have deliberately changed their labour
demand. We will thus only analyse workers who report the same firm size from 1997 onwards
in order to address any issues of self-selection into the treatment. By doing so, we lose a little
under six per cent of the baseline sample (the coefficients for the effect of job insecurity on the
probability of being married are actually not materially affected by this restriction). Note that
we do not consider here the workers who were protected by the reform, i.e. workers aged 50+.
Almost 80% of respondents in this group were already married, and there was little subsequent
movement out of marriage (97.5% of those aged 50+ reported the same marital status during
the years before and after the announcement of the reform). The limited within-variation
considerably attenuates the statistical power for this older group.
Our analysis sample consists of 10,371 observations on 2,797 different individuals over
the 1995 to 2001 period (we cannot include the first 1994 ECHP wave, as it does not include
8
Figure A2 in the Appendix plots the distribution of self-reported job security. Around 70% of the respondents
use the values of 4 or 5 on the 1-6 scale. Negative skewness of this type is commonly found for subjective
measures.
9
This quit analysis can also be carried out using the ECHP data: the probability of both the end of the job match,
and of layoffs in particular, at time t are negatively correlated with self-reported job security at time t-1.
13
information on whether the firm is public or private) between the ages of 20 and 49. Appendix
Figure A1 describes the number of observations at each stage of our sample selection.
10
Table
2 lists the descriptive statistics first for the whole analysis sample and then separately by
gender. Just under half (47%) of the sample are in the treated group. With respect to our two
dependent variables, 57% of observations come from individuals who report being married,
and average job security is equal to 4.13. The women in our sample are on average more
educated than their male counterparts but report fewer weekly working hours and lower
monthly wages.
Although we focus on a particular part of the French population, i.e. young private-sector
workers with a permanent contract, the share of married workers in our estimation sample is
similar to that in national statistics (see Figure 1). The share of cohabiting workers in our
sample is slightly higher than the national figure, which likely reflects that we exclude
individuals above age 49. The marriage share in the French adult population fell steadily from
1990 to 2009 in Figure 1, from 56% to 52%. Over the same period, the share of cohabiting
individuals rose markedly by 9.5 percentage points (from 6.7% to 16.2%).
5. Main Results
The effect of the higher Delalande firing tax on the perceived job security of younger
workers appear in Table 3. These results come from the analysis sample described in Table 2.
Columns (1), (3) and (5) of Table 3 display the D-i-D estimates of Equation (1), where we
separately identify the effects of the announcement and implementation of the tax change. The
estimates in columns (2), (4) and (6) then refer to the coefficients in Equation (2), where we
10
The fall in the number of observations due to missing values (from 13,966 to 10,936 observations) is mostly
due to missing firm size. As firm size is essential in defining the treatment status in our regressions, we do not
impute its missing values. Balance tests confirm that the observable characteristics of the observations dropped
due to missing firm size are not significantly different from those with valid information. Around 1,000
observations are also lost because of missing values for weekly working hours. Again, these dropped observations
have characteristics similar to those in the estimation sample.
14
combine the effects of the announcement and implementation of the reform into one “total”
effect. The first two columns of Table 3 show the estimates for the whole sample, while
columns (3) - (4) and (5) - (6) respectively refer to those for women and men.
Column (1) reveals that the perceived job security of younger workers in large firms fell
significantly after the 1999 rise in the layoff tax (as compared to that of younger workers in
smaller firms, where the layoff tax remained unchanged). This fall is seen after both the
announcement and the implementation of the higher layoff tax. The magnitude of the estimates
is in line with the findings in Georgieff and Lepinteur (2018) and Clark and Lepinteur (2022).
11
We carry out pairwise Wald-tests to confirm that there is no statistical difference between the
estimated announcement and implementation impact of the tax change in the first and second
rows. Column (2) then combines these two into one variable reflecting the total effect of the
reform. Unsurprisingly, the estimated coefficient on this variable turns out to be significantly
different from zero, negative and of similar size to the two coefficients in column (1). Our
estimates suggest that the perceived job security of younger workers in treated firms fell by 0.2
points due to the 1999 increase in the Delalande tax: from Table 2, this corresponds to a change
of one-sixth of a standard deviation. We find similar estimates for women and men in the
remaining columns of Panel A. Although the reform effects look larger for men, given the
associated standard errors these gender differences are not significant. We can replicate these
results using a simple binary variable for job security (comparing values 1-4 against 5-6): all
of the treatment estimates continue to be significant, with an effect that is again larger, but not
significantly so, for men.
11
The figures are not exactly the same, as the analysis samples differ across the papers. In Georgieff and Lepinteur
(2018), the ECHP analysis sample pools workers of all ages, and a triple difference-in-differences is used to
identify the estimated effect of the firing tax on workers below 50 and 50 or above (the age threshold). Clark and
Lepinteur (2022) analyse the same age group as we do here, namely the under-50’s, but focus on the fertility
behaviour of those who were already married before the reform and continued to be so afterwards.
15
Panel B of Table 3 assesses the impact of greater job insecurity on the probability of being
married in the same analysis sample.
12
The estimated coefficients for the whole sample in
columns (1) and (2) are positive but not significant at conventional levels. As it has been
suggested that women are on average more risk averse than men (Sunden and Surette, 1998;
Finucane et al., 2000; Hartog et al., 2002; Holt and Laury, 2002; Fehr-Duda et al., 2006;
Croson and Gneezy, 2009; Charness and Gneezy, 2012), we might expect these estimates to
differ by gender. The results in columns (3) and (4) show that women are significantly more
likely to be married following the implementation of the reform. In line with Panel A, the
pairwise Wald-tests again confirm that there is no statistical difference between the anticipation
and implementation effects of the reform. We combine these two effects in column (4), and
find that the rise in the Delalande tax increased the probability of being married for women by
just over four percentage points.
13
This corresponds to 7.5% of the pre-reform share of married
women in the treated group (57%). It is also half of the observed difference in female marriage
rates between the early- and late -20 age groups (see Table A1).
There is no significant marriage effect for the corresponding sample of men in columns
(5) and (6). A first possibility is that prime-age women who worked in permanent private-
sector jobs in the 1990s in France were differentially selected than their male counterparts. The
comparison of the pre-reform characteristics of men and women who do and do not appear in
our estimation sample actually reveals only small differences: there is more positive selection
of women into the estimation sample by education than there is for men, and less selection by
hours of work. However, we control for both of these variables in our estimations (and will, in
addition, show below that there is no differential effect of the reform treatment on marriage by
either education or hours of work).
12
Appendix Table A1 reports the estimated coefficients for all of the control variables.
13
These regressions include a number of sociodemographic controls, as noted at the foot of the table. The inclusion
of these controls does not change the treatment effects. This is not surprising if the assignment to the treatment is
random. The results without these control variables are available upon request.
16
Second, this may reflect that men are on average less risk-averse: the rise in job insecurity
from the 1999 increase in the Delalande tax may not have had much effect on their demand for
insurance. Last, men who looked for insurance through marriage may not have been able to
find it: their greater job insecurity may make them less attractive. This is consistent with Ekert-
Jaffé and Solaz (2001 and 2002) and Landaud (2021), where men without a job or in temporary
employment were less likely to be partnered.
Third, according to Oppenheimer (1988) and subsequent work (see Kalmijn, 2011, for a
detailed review), the job security of men has a greater influence on the marriage decision than
does that of women in societies where men are traditionally considered as the breadwinner. In
our case, the loss in job security caused by the 1999 rise in the Delalande tax may be more
detrimental for men in the sense that their “value” as a spouse is reduced.
One of the requirements for D-i-D estimation to produce causal effects is that of a common
trend in the dependent variable in the control and treatment groups in the absence of the policy
reform. Figure A3 in the Online Appendix provides additional evidence in favour of the
parallel-trend assumption by plotting the time profile of average perceived job security and the
marriage probability for the control and treatment groups. In the spirit of an event-study, Figure
2 thus plots the estimated yearly effects of being in the treatment (as opposed to the control)
group on perceived job security (in Panel A) and the probability of being married (Panel B).
The left-hand side shows the results for the whole sample and the right-hand figure those for
men and women separately.
14
None of the pre-reform announcement estimates (that can be
considered as placebos) in Figure 2 are significantly different from zero, providing evidence in
favour of the common-trend assumption.
15
On the contrary, all the post-reform announcement
14
In these latter figures we have slightly horizontally-shifted the curves for men and women so that the confidence
intervals around the estimates can be seen clearly.
15
In the bottom-right panel of Figure A3, the marriage rate of women in the treatment group is lower than that of
women in the control group before the reform. Table A3 reveals that women in the treatment group are more
educated and work more hours, both of which are associated with less marriage in this age group.
17
estimates in Panel A are significantly different from zero with no significant differences by
sex. The fall in perceived job security is also constant over time, so that there does not seem to
be any adaptation to the 1998 rise in the Delalande tax. The figures in Panel B confirm what
we saw in Table 3: only women reported a greater probability of marriage rose after the
announcement of the reform. As with perceived job security, the treatment effects on marriage
remain stable over time.
16
The causal interpretation of our estimates also relies on there being no endogenous changes
in the sample composition. By initially excluding the workers who reported a change in firm
size after the announcement of the reform, we addressed the issue of self-selection into treated
or control firms. This restriction addresses both the influence of workers changing jobs and
firms adjusting the number of employees. However, we also need to ensure that the reform, via
differences in layoff decisions between small and large firms, did not change the sample
composition. We rule out this possibility in Appendix Tables A2 and A3, where we find no
significant differences in the gaps between the control and treatment groups with respect to
most observable characteristics before and after the reform’s announcement. However, weekly
working hours in the treatment group dropped significantly more (both in the whole sample
and for women) than in the control group. This is unsurprising: in 2000, the standard workweek
in France fell from 39 to 35 hours for workers in firms with 20 employees and more. While we
do not worry about the whole sample (as there is no significant effect of the 1999 rise of the
Delalande tax on the probability to be married for the whole sample), our treatment estimates
for women might then capture the influence of the 2000 shorter workweek. We will rule out
this possibility in the robustness section by dropping workers in firms with fewer than 20
16
These results can be explained by timing effects and changes at extensive margin. In the first case, the decision
to marry has already been taken and the reform has simply brought it forward. In the second case, the reform may
encouraged women who had not initially planned to marry to do so. Our data do not allow us to distinguish
between these two cases. It is however plausible to think that the timing effect would fall over time. This is not
what we see in Panel B of Figure 2 reveals: the stable treatment effect for women suggests that timing does not
play a major role.
18
employees from the estimation sample. Last, it could be argued that Appendix Tables A2 and
A3 do not tackle endogenous changes in unobservable characteristics: we will also address this
issue in the robustness check section.
Last, we ask in Table A4 whether the total impact of the increase in the Delalande Tax led
to other changes in terms of marital status and couple formation. The first column of this table
reproduces the estimated marriage coefficients from columns (2), (4) and (6) of Table 3 for
comparison. We then consider in column (2) the probability of being in a partnership,
irrespective of legal marital status: the estimates here show that the greater job insecurity from
the reform had no effect on being in a couple in general. Combined with the results for
marriage, we thus expect (for women) a negative reform effect on cohabitation, which is indeed
what we find in column (3) of Panel B: women whose job insecurity rose following the change
in the Delalande tax are less likely to be in a non-married relationship. As the ECHP is a
household panel, where all adults are interviewed, and as we here focus on the transition from
cohabitation to marriage, it might be thought that the estimation results for men and women
should be identical (as we are looking at the two sides of a couple). This is not the case in Table
A4. The explanation is that the men and women in this table do not come from the same sample
of cohabitees: fully one half of treated women who we observe switching from cohabitation to
marriage marry men who are not in the sample in Panel C (as the men’s jobs are temporary or
in the public sector, they are not active in the labour force, or they are aged 50 or over).
Last, Columns (4) and (5) of Table A4 look in turn at being divorced/separated or never
having been in a relationship, finding no significant effects for men, women, or the whole
sample.
17
That our results only concern cohabitation-marriage transitions may reflect the rather
17
The marital-status categories in Table A4 are not mutually-exclusive: an individual can be both divorced and
in a new relationship. We find similar results to those in Table A4 when we estimate a multinomial-logit model
using the following mutually-exclusive categories: “Married in a partnership”, “Non-married in a partnership”,
“Divorced or separated with no partner” and “Never in a relationship” (we exclude widows here, as there are too
few observations). These results are available upon request.
19
short run of post-reform observations. Had the ECHP survey continued beyond 2001, we may
well have found other types of transitions too, as moving from being single or divorced to
married likely takes more time than switching from cohabitation to marriage.
6. Robustness Checks and Heterogeneity
The estimates in Table 3 refer to the effect of the 1999 rise in the Delalande tax on first
perceived job security and then marital status for the whole sample. We now turn to a number
of robustness checks, and then ask whether the effects of job insecurity on the probability of
being married are larger for certain types of workers. The analyses here refer to the total impact
of the 1999 rise in the Delalande tax, as in Equation (2).
18
We have also estimated the analogous
effect of all of our robustness and heterogeneity tests for the estimated effect of the reform on
perceived job security; these results appear in Appendix Table A5.
a. Robustness Tests
i. Ruling out confounding events
As noted above, the 2000 working-time reduction was another notable labour-market
reform around the time of the modification in the Delalande tax. In 1998, the French Ministry
of Labour decided to reduce the standard workweek from 39 to 35 hours, but only in larger
firms (those with 20 employees or more). A shorter workweek may induce worries about job
security due to potential effects on firm profitability, affecting our estimated coefficients. We
evaluate this possibility by excluding all workers who were unaffected by the 35-hour week
(those in firms with under 20 employees). As such, all of the workers in the restricted analysis
sample will have been affected by the 35-hour week in the same way. This exclusion drops
around 15% of the original estimation sample. The resulting estimated coefficients in column
(1) of Table 4 are close to those in the baseline (in columns (2), (4) and (6) of Panel B in Table
18
We obtain similar results for both the robustness checks and heterogeneity analyses when we separately estimate
the effects of the announcement and implementation of the reform.
20
3). Note that when we exclude workers in firms with under 20 employees from our estimation
sample, the differences in weekly working hours in Tables A2 and A3 become statistically
insignificant.
As part of our identification relies on changes over time, we should check that any effect
comes from the change in the Delalande tax, and not some wider macro-economic changes.
We rule this out by re-estimating our baseline D-i-D equations using samples of employees
with similar characteristics from neighbouring and arguably-similar countries (where of course
the French layoff tax did not apply). Although the ECHP is relatively well-harmonised across
countries, we have to restrict this comparison to Spain, Italy and Denmark due to data
limitations.
19
The resulting D-i-D estimates for these three countries in columns (2), (3) and (4)
of Table 4 are not significantly different from zero. The impact of job insecurity on the share
of married women does not then seem to reflect macroeconomic trends.
ii. The estimation method
Our main results above come from fixed-effect analyses, comparing the same individual
before and after the labour-market reform. We expect fixed-effects and OLS analyses to
produce different estimates for two reasons. First, the former introduce attenuation bias when
there is measurement error: in absolute terms, the resulting FE estimates would be lower than
their OLS counterparts. Second, unobserved individual time-invariant characteristics that are
correlated with the treatment and are not controlled for in pooled analyses will bias the OLS
estimates. Column (5) of Table 4 lists the treatment effects without individual fixed effects.
These are somewhat larger than the baseline estimates (that include individual fixed effects)
but are not statistically different.
19
The variables in the final waves of the ECHP in Belgium and Germany does not allow us to accurately
distinguish the public from the private sector, or to measure perceived job security.
21
We also ask whether our results are influenced by the way in which we define the
dependent variable. As we use linear-probability models with individual fixed-effects, our
baseline regressions treat the probability of being married as a cardinal variable. However, it
can be argued that non-linear estimation is more suitable for this dummy dependent variable.
Column (6) of Table 4 thus applies a conditional fixed-effect logit model to re-estimate our
main regression. The results remain the same: job insecurity significantly increases the
probability of being married for women only.
iii. Sample composition
Firm size is reported by the respondent and may not be accurate. In this case, individuals
can be mis-allocated to firm-size groups, and hence to the control or treatment groups. This
mis-reporting may be random, in which case we estimate a lower bound of the treatment (as
some of the control group are treated, and some of the treated group are not). A potentially
more-serious problem arises if the mis-allocation is not random, conditional on the control
variables in our regressions. We address mis-reporting using a regression in line with the donut
regression-discontinuity design. We here drop employees who are close to the firm-size
treatment threshold of 50 employees: from the firm-size categories listed in Section 3, this
implies re-estimating the treatment without respondents who report working in firms with “20
to 49 employees” or “50 to 99 employees”. Intuitively, mis-judgement may cause workers to
erroneously report a firm-size category above or below the correct value, but not to jump three
categories (so that they report a firm size of under 20 employees when the real value is over
100 employees, for example).
The estimated treatment coefficients from the baseline analysis (in panel B of Table 3) and
these “donut” regressions (in the last column of Table 4) are of the same size, although the
estimate of the latter for women is no longer significant (due to the smaller sample size when
we drop two firm-size categories, producing larger standard errors). That the estimated
22
coefficient does not change between the two specifications helps to dispel any concerns
regarding any systematic mis-reporting of firm size.
20
The last robustness check addresses the issue of attrition. 16% of both the treatment and
control group left the analysis sample between the announcement of the reform and the last
wave (with no noticeable differences between men and women). Our estimates may be biased
if the implementation of the reform and attrition are not independent, in particular if the
marriage probability of leavers falls once they leave the survey. We rule out this issue in two
ways. First, we can make use of the two attrition weights supplied by ECHP (we do not use
weights in our main regressions). When we do so, our results are little affected.
We can also calculate the “unobserved” D-i-D in the attrition group that would be required
to cancel out our main treatment effect. From Table 3, the difference in the probability of being
married between women in treated and non-treated firms rose by 4 percentage points post-
reform. To cancel this figure out, we would require the analogous figure for the 16% of women
in the attrition group to be a fall of 21 percentage points after the reform (= 0.04*84%/16%).
This figure is more than double the largest age effect on marriage in column (4) of Table A1,
and does not seem plausible.
b. Heterogeneity Analysis
The estimates in Panel B of Table 3 show the average effect of the 1999 increase in the
Delalande tax for employees in large firms. In Table 5 we consider whether these effects might
change across types of workers. We first consider age as this may ambiguously affect the
treatment. First, one could argue that the older workers may have less of an incentive to get
married as they will spend less time on the labour market threatened by the rise in the Delalande
tax than younger workers. But we also know that workers just below the age-50 threshold
20
As in Clark and Lepinteur (2022), we can consider public-sector workers as an alternative control group but
none of our treatment estimates here were significantly different from zero at conventional levels. However, the
marriage probability pre-trends differ significantly between the treatment group and public-sector workers, casting
doubt on the validity of this control group. All of these results are available upon request.
23
perceived a greater risk of job loss (as shown in Georgieff and Lepinteur, 2018) and that risk-
aversion increases with age (Falk et al., 2018). We investigate by interacting the total effect of
the reform with a dummy for being born in 1963 (the median birth year in the estimation
sample) or before. The resulting estimates appear in columns (1) and (6) of Table 5 for women
and men respectively: neither interaction term is significant. The estimated interaction terms
using different birth-year thresholds are also insignificant. We also interacted the treatment
effect with multiple age categories (in five- and ten-year bands) but continue to find no
significant interaction terms. All of these results are available upon request.
We second know that the relationship between fertility decisions and job insecurity likely
depends on education and earnings (Chevalier and Marie, 2017; Clark and Lepinteur, 2022),
and ask whether an analogous relationship is found for marriage. We thus interact the treatment
effect with dummies for “High-education” (for workers with post-secondary education) and
“High-wage” (for workers with above-median wage), measured in the pre-reform years only.
21
The estimated coefficients on these interactions in columns (2), (3), (7) and (8) of Table 5
suggest no significant difference in the effect of job insecurity on marriage by education or
wages.
22
Our last interactions concern pre-reform family characteristics. Columns (4) and (9) show
that the treatment effect is significantly larger for women who had at least one child before the
reform. One natural interpretation is that women with children are exposed to greater risk than
men with children, as in France the former ended up with sole custody of the children in 80%
of separations in the early 2000s. We also interact the treatment effects with the number of
children in the household in 1998 (i.e. just before implementation of the reform). The results
are displayed in columns (5) and (10). The estimated coefficients are positive and significant
21
We also interacted the treatment effects with a continuous measure of monthly wage (in logs), which produces
similar results. Using household income rather than the monthly wage also makes no difference.
22
With respect to hours of work, interactions with pre-reform hours either as a continuous variable or as a dummy
for part-time work (between 1 and 29 hours per week) yielded no significant estimates.
24
(and of the same size) only for women with one or two children (although that for women with
three or more children is not statistically different from that for mothers with one or two
children).
Why does the effect of job insecurity differ across some groups of workers? It is possible
that the impact of the change in the Delalande tax on job insecurity is stronger for some
workers, and especially those who already had children. We investigate in Appendix Table A6,
where we replicate our heterogeneity analysis using perceived job security as the dependent
variable. None of the interaction terms attracts a significant coefficient, so that impact of the
reform was the same in terms of perceived job security for all of the different types of workers
we consider.
A second possibility is that the relationship between perceived job security and marital
status varies across workers, for reasons of risk-aversion. In Görlitz and Tamm (2015), the
higher risk-aversion due to parenthood is larger for women. As such, mothers arguably
constitute the most risk-averse group of workers in our estimation sample, which might explain
why they are the most likely to get married after a rise in their job insecurity.
7. Risk-Sharing Theory
Our results above are in line with risk sharing theory, as the rise in marriage following
greater job insecurity only appears for the arguably most risk-averse workers (women with
children), although we have not yet provided any explicit test of this theory. In Hess (2004)
and Shore (2010), couples manage income risk by trying to ensure that the two partners’
exogenous income shocks are not perfectly positively correlated. In the context of the reform
analysed here, we thus expect a stronger treatment effect for women whose partner has a more
stable job.
25
This is what we test in Table 6, where we interact the reform effect with dummy variables
for different types of partners. Columns (1) and (4) show the average effect of the reform for
women and men, as in Panel B of Table 3, while columns (2) and (5) show the estimated
coefficients of the interaction of the reform with a dummy for “Employed Partner”. As risk-
sharing theory would predict, the shift from cohabitation to marriage (following the results in
Table A4) is only higher for women whose partners are currently working.
23
We find no
significant results for any of the groups of men.
We then take the treatment status of the partner into account. Under risk-sharing, household
members try to avoid correlated risks. The marriage incentives from job insecurity should then
be weaker for women whose partner is also affected by the reform. The estimates in column
(3) are in line with this prediction: marriage only rises significantly for women whose partner
is not affected by exogenously-higher job insecurity. We found no significant differences in
this respect for the partner working in the public vs. private sector or having a permanent vs.
temporary contract (although this latter might reflect a lack of statistical power, as only around
3% of partners in our estimation sample had temporary contracts). We continue to find no
significant marriage results for any of the groups of men (in the last column of Table 6).
8. Conclusion
Job insecurity increases the probability of marriage for women. The 1999 reform in the
French labour market taxed the layoffs of older workers, but at the cost of switching risk to
younger workers. As the higher layoff tax was applied only in larger firms, we can estimate
difference-in-difference regressions. We find that the reform-induced exogenous change in the
23
The interactions in Table 6 refer to the partner’s current labour-market position. As such, the employed in row
2 may have started work after their partner was treated. Equally, in rows 3 and 4, partner treatment may have
caused individuals to switch into more secure jobs (i.e. those that are not affected by the higher Delalande tax). If
we only consider the partner’s pre-reform employment status, we do not allow for these behavioural reactions (if
we do so we actually find similar estimates, although the coefficients are less precisely-estimated).
26
future probability of job loss both reduced perceived job security, and led to a robust significant
rise in the probability of being married for women. Our identification strategy here evaluates
the effect of job insecurity on the probability of being married for individuals who were in
employment both before and after the reform (and not those who changed their labour-force
status): as such, our insecurity is forward-looking and does not apply to events that have already
taken place.
Our results are novel as they are, to the best of our knowledge, the first that appeal to a
natural labour-market experiment to show that risk-sharing is one of the causes of marriage.
Job insecurity increased the probability of marriage for women, and more so for those who are
probably more risk-averse (mothers). In line with risk-sharing theory, we show that this
marriage effect was not found for couples in which the layoff risk rose for both partners after
the 1999 tax rise (i.e. couples where both members worked in treated firms). The lack of any
effect for men may reveal their lower risk-aversion with respect to job insecurity, or the greater
difficulty that insecure men face on the marriage market.
Part of the attraction of marriage then seems to be the risk-sharing it provides. Why then
don’t all couples get married? There are a wide variety of factors at play here, including cultural
norms. Some of these can be argued to be economic. In the same way that employment
protection legislation might discourage hiring due to the costs of firing, more rigid or expensive
divorce procedures may discourage some couples from marrying. In this case, more flexible
divorce procedures may lead to more people getting married, as would a more-advantageous
tax treatment of the married relative to the cohabiting. The flexibility of marriage and the labour
market can thus be considered as intertwined.
27
Ethical Statement: This study uses retrospective deidentified data collected by Eurostat. No
ethical approval is required.
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Figures and Tables
Figure 1: Marriage and Cohabitation Rates in France from 1990 to 2009
Notes: The marriage and cohabitation rates correspond respectively to the number of married
and cohabiting individuals divided by the size of the adult population. These rates were
calculated using data from the INED time series on the number of married and cohabiting
couples (https://www.ined.fr/fr/tout-savoir-population/chiffres/france/couples-menages-
familles/couples_menages_familles/) and the UN website
(http://data.un.org/Data.aspx?d=POP&f=tableCode%3A22) for the adult population.
33
Figure 2: Parallel-Trend Assumption – Panel Results
Panel A: Perceived Job Security
Panel B: Probability of being married
Notes: The data come from the French component of the ECHP panel. The analysis sample is made up of individuals
who work in the private sector, with a permanent contract and who are aged between 20 and 49. There are 10371
observations in the whole sample (2797 individuals), 4248 observations in the sample of women (1175 individuals)
and 6123 observations in the sample of men (1622 individuals). Each point shows the gap in the outcome between
the treatment group (i.e. being a younger worker in a firm with 50 or more employees) and the control group (being
a younger worker in a firm with fewer than 50 employees) in that year, as compared to the same gap in the omitted
year (1995). These numbers come from regression analyses that control for year and individual fixed effects, as well
as age dummies (in five-year bands), health status, the (lagged) number of children in the household, the log of the
monthly wage, weekly working hours, and occupation and region fixed effects. The error bars represent the 90%
confidence intervals. The dotted vertical line shows the date at which the increase in the Delalande tax was
announced, and the dashed vertical line the date of its implementation. Men and women were interviewed at the same
date; we have slightly left-shifted the points for men so that the confidence intervals can be seen.
34
Table 1: The Evolution of the Delalande Layoff Tax over Time
Worker’s age
50
51
52
53
54
55
56-57
58
59
July 1987-June 1992
All firm sizes
3
3
3
3
July 1992 - Dec. 1992
20 or more employees
1
1
2
2
4
5
6
6
6
Under 20 employees
0.5
0.5
1
1
2
2.5
3
3
3
Jan 1993-Dec 1998
All firm sizes
1
1
2
2
4
5
6
6
6
January 1999-2008
50 or more employees
2
3
5
6
8
10
12
10
8
Under 50 employees
1
1
2
2
4
5
6
6
6
Source: Legislative texts.
Notes: This table lists the amount of the tax to be paid by the firm to the unemployment-insurance system if it lays off a worker of a
given age. This tax is expressed as a multiple of the worker’s monthly gross wage. This tax applies to private-sector workers with
permanent contracts only.
35
Table 2: Descriptive Statistics of the ECHP Analysis Sample
Whole Sample
Women
Men
Mean
SD
Min
Max
Mean
SD
Min
Max
Mean
SD
Min
Max
Dependent Variable:
Perceived Job Security [1-6]
4.13
1.17
1
6
4.14
1.18
1
6
4.13
1.16
1
6
Married
0.57
0
1
0.58
0
1
0.56
0
1
Partnered
0.77
0
1
0.76
0
1
0.78
0
1
Partnered but not Married
0.19
0
1
0.18
0
1
0.21
0
1
Divorced or Separated
0.04
0
1
0.05
0
1
0.03
0
1
Never in a Relationship
0.20
0
1
0.19
0
1
0.20
0
1
Difference-in-differences Variables:
Treatment Group
0.47
0
1
0.44
0
1
0.49
0
1
Post Period
0.37
0
1
0.36
0
1
0.38
0
1
Individual Characteristics:
Age
34.6
6.85
20
49
34.4
20
49
34.8
6.81
20
49
Female
0.41
0
1
-
0
1
-
High Education (Post-Secondary)
0.24
0
1
0.28
0
1
0.21
0
1
No. of Children in Household (lagged)
0.95
0.97
0
8
0.87
0
8
1.00
1.03
0
8
Health Status [1-5]
3.88
0.72
1
5
3.85
1
5
3.90
0.72
1
5
Job Characteristics:
Weekly Working Hours
39.4
7.86
2
96
36.3
7.78
2
90
41.5
7.20
4
96
Monthly Wage in French Francs (log)
8.98
0.48
5.08
11.60
8.80
0.49
5.08
11.60
9.10
0.44
5.41
11.07
Observations
10371
4248
6123
Individuals
2797
1175
1622
Note: The data come from the French component of the ECHP panel. The analysis sample is made up of individuals who work in the private sector, with a permanent contract and
who are aged between 20 and 49.
36
Table 3: The Delalande Tax, Job Security and the Probability of Being Married – Panel Results
Panel A: Job Security
Whole Sample
Women
Men
(1-6)
(1)
(2)
(3)
(4)
(5)
(6)
Reform Announcement
-0.188
***
-0.149
*
-0.225
***
(0.057)
(0.090)
(0.075)
Reform Implementation
-0.202
***
-0.159
*
-0.240
***
(0.053)
(0.086)
(0.067)
Total Effect of the Reform
-0.198
***
-0.156
**
-0.236
***
(0.048)
(0.077)
(0.062)
Panel B: Married
Whole Sample
Women
Men
(Dummy)
(1)
(2)
(3)
(4)
(5)
(6)
Reform Announcement
0.018
0.031
0.010
(0.012)
(0.020)
(0.016)
Reform Implementation
0.015
0.047
*
-0.007
(0.015)
(0.025)
(0.019)
Total Effect of the Reform
0.016
0.042
**
-0.002
(0.013)
(0.021)
(0.017)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis sample is
made up of individuals who work in the private sector, with a permanent contract and who are aged between 20 and 49. There
are 10371 observations in the whole sample (2797 individuals), 4248 observations in the sample of women (1175 individuals)
and 6123 observations in the sample of men (1622 individuals). The announcement effect of the reform covers the treatment
from the start of 1998, when the Delalande-tax reform we consider was announced, up to its implementation on January 1
st
1999; the subsequent reform-implementation effect refers to the implementation period starting on January 1
st
1999. These
two effects respectively correspond to α
2
and α
3
in Equation (1). The total effect of the reform corresponds to β
2
in Equation
(2). Standard errors in parentheses are clustered at the individual level. All of the regressions control for year and individual
fixed effects, as well as age dummies (in five-year bands), health status, the (lagged) number of children in the household, the
log of the monthly wage, weekly working hours, and occupation and region fixed effects. *, ** and *** indicate significance
at the 10%, 5% and 1% levels respectively.
37
Table 4: The Rise in the Delalande Tax and the Probability of being Married – Robustness Checks
20+
employees
Spanish
sample
Italian
sample
Danish
sample
OLS
Conditional
FE Logit
Donut
DiD
(1)
(2)
(3)
(4)
(5)
(6)
(7)
Panel A: Whole Sample
Total Effect of the reform
0.009
-0.003
0.011
0.025
0.033
0.193
0.014
(0.018)
(0.013)
(0.013)
(0.026)
(0.020)
(0.370)
(0.015)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
8897
8938
11373
4682
10371
10371
7523
Individuals
2360
3127
3411
1432
2797
2797
2067
Panel B: Women
Total Effect of the reform
0.050
**
-0.040
0.022
-0.014
0.058
*
1.530
**
0.043
(0.024)
(0.025)
(0.023)
(0.041)
(0.033)
(0.772)
(0.028)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
3494
3084
4265
1739
4248
4248
3018
Individuals
944
1134
1340
572
1175
1175
860
Panel C: Men
Total Effect of the reform
-0.019
0.010
0.006
0.055
0.022
-0.176
-0.005
(0.025)
(0.016)
(0.015)
(0.033)
(0.025)
(0.498)
(0.019)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
5403
5584
7108
2943
6123
6123
4505
Individuals
1416
1993
2071
860
1622
1622
1207
Notes: These are linear regressions, except in column (6). The dataset is the French component of the ECHP panel (except in columns (2), (3) and (4), where it is
respectively the Spanish, Italian and Danish ECHP components). The analysis samples consist of individuals who work in the private sector, with a permanent
contract and are aged between 20 and 49. The total effect of the reform corresponds to β
2
in Equation (2). Standard errors in parentheses are clustered at the individual
level, except in column (6). The conditional FE logit coefficients in column (6) refer to the log of the odds ratio. All of the regressions control for year and individual
fixed effects, as well as age dummies (in five-year bands), health status, the (lagged) number of children in the household, the log of the monthly wage, weekly
working hours, and occupation and region fixed effects. The individual time-invariant controls are gender and education dummies. *, ** and *** indicate significance
at the 10%, 5% and 1% levels respectively.
38
Table 5: The Rise in the Delalande Tax and the Probability of being Married – Heterogeneity Analysis
Women
Men
(1)
(2)
(3)
(4)
(5)
(6)
(7)
(8)
(9)
(10)
Total Effect of the Reform
0.062
0.014
0.019
-0.040
-0.040
0.018
-0.002
-0.012
0.014
0.014
(0.039)
(0.022)
(0.023)
(0.033)
(0.033)
(0.031)
(0.018)
(0.029)
(0.024)
(0.024)
Interacted with:
Born before 1963
-0.031
-0.033
(0.045)
(0.036)
High Education
0.080
-0.018
(0.055)
(0.049)
High Monthly Wage
0.039
0.014
(0.039)
(0.036)
Parent beforehand
0.114
***
-0.023
(0.042)
(0.032)
1 child beforehand
0.113
**
-0.015
(0.053)
(0.050)
2 children beforehand
0.124
**
-0.047
(0.052)
(0.037)
3+ children beforehand
0.065
0.022
(0.046)
(0.041)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis sample is made up of individuals who work in the
private sector, with a permanent contract and who are aged between 20 and 49. There are 4248 observations in the sample of women (1175 individuals) and 6123
observations in the sample of men (1622 individuals). The total effect of the reform corresponds to β
2
in Equation (2). Standard errors in parentheses are clustered at
the individual level. All of the regressions control for year and individual fixed effects, as well as age dummies (in five-year bands), health status, the (lagged) number
of children in the household, the log of the monthly wage, weekly working hours, and occupation and region fixed effects. *, ** and *** indicate significance at the
10%, 5% and 1% levels respectively.
39
Table 6: The Rise in the Delalande Tax and the Probability of being Married by Partner’s Employment Status – Panel Results
Women
Men
(1)
(2)
(3)
(4)
(5)
(6)
Total Effect of the Reform
0.042
**
-0.016
-0.015
-0.002
-0.018
-0.020
(0.021)
(0.031)
(0.031)
(0.017)
(0.021)
(0.021)
Interacted with:
Employed Partner
0.093
**
0.032
(0.036)
(0.027)
Employed Partner Not Currently Affected by the Higher Delalande Tax
0.101
**
0.029
(0.039)
(0.028)
Employed Partner Currently Affected by the Higher Delalande Tax
-0.017
-0.008
(0.042)
(0.055)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis sample is made up of individuals who work in the private
sector, with a permanent contract and who are aged between 20 and 49. There are 4248 observations in the sample of women (1175 individuals) and 6123 observations in
the sample of men (1622 individuals). The total effect of the reform corresponds to β
2
in Equation (2). The standard errors in parentheses are clustered at the individual level.
All of the regressions control for year and individual fixed effects, as well as age dummies (in five-year bands), health status, the (lagged) number of children in the household,
the log of the monthly wage, weekly working hours, and occupation and region fixed effects.*, ** and *** indicate significance at the 10%, 5% and 1% levels respectively.
40
Appendix
Figure A1: Sample Selection
-
Note: Each box shows the number of observations in the French ECHP sample, from the raw data (95 171 observations) to the
sample that is used in our empirical analyses (10 371 observations).
41
Figure A2: The Distribution of Perceived Job Security in the ECHP
Notes: The data come from the French component of the ECHP panel. The analysis sample is
made up of individuals who work in the private sector, with a permanent contract and who are
aged between 20 and 49.
42
Figure A3: Average Perceived Job Security and the Probability of being Married by Treatment Group
Panel A: Perceived Job Security
v
Panel B: Probability of being married
Note: The dotted vertical line indicates the date at which the rise in the Delalande tax was announced, and the dashed vertical
line the date of its implementation.
43
Table A1: The Delalande Tax and Marriage – Panel results with the full set of controls
Whole Sample
Women
Men
(1)
(2)
(3)
(4)
(5)
(6)
Reform Announcement
0.018
0.031
0.010
(0.012)
(0.020)
(0.016)
Reform Implementation
0.015
0.047
*
-0.007
(0.015)
(0.025)
(0.019)
Total Effect of the reform
0.016
0.042
**
-0.002
(0.013)
(0.021)
(0.017)
Age: 26-30
0.109
***
0.109
***
0.090
***
0.089
***
0.124
***
0.123
***
(0.020)
(0.020)
(0.031)
(0.031)
(0.027)
(0.027)
Age: 31-35
0.113
***
0.113
***
0.088
**
0.087
**
0.132
***
0.132
***
(0.024)
(0.024)
(0.034)
(0.034)
(0.032)
(0.032)
Age: 36-40
0.097
***
0.097
***
0.063
*
0.062
*
0.122
***
0.122
***
(0.026)
(0.026)
(0.036)
(0.036)
(0.035)
(0.035)
Age: 41-45
0.061
**
0.061
**
0.003
0.002
0.101
***
0.100
***
(0.028)
(0.028)
(0.042)
(0.042)
(0.038)
(0.038)
Age: 46-49
0.006
0.006
-0.080
-0.081
0.065
0.064
(0.032)
(0.032)
(0.051)
(0.050)
(0.042)
(0.042)
Monthly Wage (in log)
0.021
0.021
-0.007
-0.007
0.054
**
0.054
**
(0.016)
(0.016)
(0.023)
(0.023)
(0.021)
(0.021)
Weekly Working Hours
-0.001
**
-0.001
**
-0.000
-0.000
-0.001
**
-0.001
**
(0.000)
(0.000)
(0.001)
(0.001)
(0.001)
(0.001)
Self-Assessed Health [1-5]
-0.001
-0.001
-0.006
-0.006
0.004
0.004
(0.004)
(0.004)
(0.007)
(0.007)
(0.005)
(0.005)
Number of Children in the Household (lagged)
0.054
***
0.054
***
0.057
***
0.057
***
0.052
***
0.052
***
(0.007)
(0.007)
(0.012)
(0.012)
(0.009)
(0.009)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis sample is made up of individuals
who work in the private sector, with a permanent contract and who are aged between 20 and 49. There are 10371 observations in the whole sample
(2797 individuals), 4248 observations in the sample of women (1175 individuals) and 6123 observations in the sample of men (1622 individuals).
The announcement effect of the reform refers to the treatment from the beginning of 1998, when the reform to the Delalande tax was announced,
up to its implementation on January 1
st
1999; the reform-implementation effect considers the implementation treatment starting on January 1
st
1999. These two effects correspond to α
2
and α
3
in Equation (1). The total effect of the reform corresponds to β
2
in Equation (2). Standard errors
in parentheses are clustered at the individual level. All of the regressions control for individual, year, occupation and region fixed effects. *, **
and *** indicate significance at the 10%, 5% and 1% levels respectively.
44
Table A2: Differences in observable characteristics between the treatment and control groups before and after the announcement of the reform
Before the announcement of the
1999 rise in the Delalande tax
After the announcement of the
1999 rise in the Delalande tax
Difference-in-
Differences
Treatment
Control
Difference
Treatment
Control
Difference
(1)
(2)
(3)
(4)
(5)
(6)
(7)
Age
34.128
33.092
1.036
***
36.140
35.143
0.997
**
-0.040
[6.082]
[6.517]
(0.175)
[7.024]
[7.302]
(0.199)
(0.266)
Female
0.386
0.441
-0.056
***
0.380
0.425
-0.045
***
0.011
[0.487]
[0.497]
(0.014)
[0.486]
[0.494]
(0.014)
(0.019)
High Education
0.266
0.187
0.079
***
0.298
0.217
0.081
***
0.002
[0.442]
[0.390]
(0.012)
[0.458]
[0.412]
(0.012)
(0.017)
No. Children in Household (lagged)
0.943
0.943
-0.000
0.958
0.959
-0.000
-0.000
[0.984]
[0.969]
(0.027)
[0.970]
[0.954]
(0.027)
(0.038)
Health Status
3.901
3.916
-0.015
3.839
3.851
-0.013
0.002
[0.740]
[0.752]
(0.021)
[0.706]
[0.673]
(0.019)
(0.028)
Weekly Working Hours
40.108
39.516
0.592
***
38.851
39.097
-0.246
-0.838
***
[6.898]
[9.151]
(0.218)
[6.557]
[8.285]
(0.218)
(0.309)
Monthly Wage (log)
9.004
8.818
0.227
***
9.143
8.930
0.214
***
-0.013
[0.440]
[0.471]
(0.013)
[0.472]
[0.482]
(0.013)
(0.018)
Notes: There are 10371 observations in the whole sample (2797 individuals). Standard errors are in parentheses and standard deviations are in square brackets.
*, ** and *** indicate significance at the 10%, 5% and 1% levels respectively.
45
Table A3: Differences in characteristics between the treatment and control groups before and after the announcement of the reform by gender
Before the announcement of the
1999 rise in the Delalande tax
After the announcement of the
1999 rise in the Delalande tax
Difference-in-
Differences
Treatment
Control
Difference
Treatment
Control
Difference
(1)
(2)
(3)
(4)
(5)
(6)
(7)
Panel A: Women
Age
33.925
32.902
1.024
***
35.712
35.117
0.595
*
-0.428
[6.117]
[6.617]
(0.278)
[7.105]
[7.327]
(0.318)
(0.422)
High Education
0.312
0.219
0.093
***
0.338
0.255
0.082
***
-0.011
[0.464]
[0.414]
(0.019)
[0.473]
[0.436]
(0.020)
(0.027)
No. Children in Household (lagged)
0.821
0.915
-0.093
**
0.804
0.930
-0.126
***
-0.033
[0.886]
[0.859]
(0.038)
[0.85]
[0.873]
(0.038)
(0.054)
Health Status
3.888
3.876
0.012
3.801
3.835
-0.034
0.045
[0.718]
[0.754]
(0.032)
[0.710]
[0.678]
(0.030)
(0.044)
Weekly Working Hours
37.701
35.646
2.054
***
36.755
35.635
1.119
***
-0.935
*
[6.725]
[9.122]
(0.355)
[6.220]
[8.023]
(0.319)
(0.478)
Monthly Wage (log)
8.901
8.625
0.277
***
8.993
8.759
0.235
***
-0.042
[0.447]
[0.468]
(0.020)
[0.472]
[0.493]
(0.021)
(0.029)
Panel B: Men
Age
34.257
33.243
1.014
***
36.402
35.162
1.240
***
0.226
[6.058]
[6.434]
(0.227)
[6.963]
[7.286]
(0.256)
(0.343)
High Education
0.236
0.161
0.075
***
0.274
0.189
0.085
***
0.010
[0.425]
[0.368]
(0.014)
[0.446]
[0.392]
(0.015)
(0.021)
No. Children in Household (lagged)
1.019
0.966
0.054
1.053
0.980
0.073
**
0.020
[1.034]
[1.047]
(0.037)
[1.022]
[1.010]
(0.037)
(0.053)
Health Status
3.910
3.948
-0.038
3.862
3.864
-0.002
0.036
[0.753]
[0.748]
(0.026)
[0.703]
[0.669]
(0.026)
(0.037)
Weekly Working Hours
41.619
42.573
-0.954
***
40.138
41.661
-1.123
***
-0.569
[6.571]
[7.940]
(0.260)
[6.428]
[7.508]
(0.257)
(0.365)
Monthly Wage (log)
9.134
8.970
0.164
***
9.235
9.056
0.179
***
0.015
[0.411]
[0.414]
(0.015)
[0.448]
[0.434]
(0.016)
(0.022)
Notes: There are 4248 observations in the sample of women (1175 individuals) and 6123 observations in the sample of men (1622 individuals). Standard errors
are in parentheses and standard deviations are in square brackets. *, ** and *** indicate significance at the 10%, 5% and 1% levels respectively.
46
Table A4: The Delalande Tax and Other Types of Partnerships – Panel Results
Panel A: Whole Sample
Married
Partnered
Partnered
but not
married
Divorced
or
Separated
Never in a
relationship
(1)
(2)
(3)
(4)
(5)
Total Effect of the Reform
0.016
-0.023
-0.009
-0.004
0.013
(0.013)
(0.015)
(0.013)
(0.007)
(0.011)
Panel B: Women
Married
Partnered
Partnered
but not
married
Divorced
or
Separated
Never in a
relationship
(1)
(2)
(3)
(4)
(5)
Total Effect of the Reform
0.042
**
-0.013
-0.052
**
0.006
0.010
(0.021)
(0.023)
(0.025)
(0.013)
(0.020)
Panel C: Men
Married
Partnered
Partnered
but not
married
Divorced
or
Separated
Never in a
relationship
(1)
(2)
(3)
(4)
(5)
Total Effect of the Reform
-0.002
-0.007
-0.002
-0.009
0.013
(0.017)
(0.014)
(0.018)
(0.009)
(0.012)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis
sample is made up of individuals who work in the private sector, with a permanent contract and who are aged
between 20 and 49. There are 10371 observations in the whole sample (2797 individuals), 4248 observations in
the sample of women (1175 individuals) and 6123 observations in the sample of men (1622 individuals). The total
effect of the reform corresponds to β
2
in Equation (2). Standard errors in parentheses are clustered at the individual
level. All of the regressions control for year and individual fixed effects, as well as age dummies (in five-year
bands), health status, the (lagged) number of children in the household, the log of the monthly wage, weekly
working hours, and occupation and region fixed effects. *, ** and *** indicate significance at the 10%, 5% and
1% levels respectively.
47
Table A5: The Rise in the Delalande Tax and Perceived Job Security – Robustness Checks
20+
employees
Spanish
sample
Italian
sample
Danish
sample
OLS
BUC Ordered
Logit
Donut
DiD
(1)
(2)
(3)
(4)
(5)
(6)
(7)
Panel A: Whole Sample
Total Effect of the reform
-0.167
**
-0.018
-0.049
0.032
-0.110
***
-0.475
***
-0.142
***
(0.069)
(0.064)
(0.058)
(0.077)
(0.042)
(0.114)
(0.055)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
6657
8938
11373
4682
10371
16479
7523
Individuals
1740
3127
3411
1432
2797
2797
2067
Panel B: Women
Total Effect of the reform
-0.202
*
0.143
-0.065
0.167
-0.131
*
-0.379
**
-0.095
(0.109)
(0.103)
(0.105)
(0.145)
(0.070)
(0.187)
(0.087)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
2599
3084
4265
1739
4248
6546
3018
Individuals
691
1134
1340
572
1175
1175
860
Panel C: Men
Total Effect of the reform
-0.150
*
-0.092
-0.037
-0.019
-0.103
**
-0.549
***
-0.192
***
(0.090)
(0.081)
(0.069)
(0.092)
(0.052)
(0.144)
(0.073)
Individual Time-Invariant Controls
.
.
.
.
Yes
.
.
Individual Fixed Effects
Yes
Yes
Yes
Yes
.
Yes
Yes
Observations
4058
5854
7108
2934
6123
9933
4505
Individuals
1049
1993
2071
860
1622
1622
1207
Notes: These are linear regressions, except in column (6). The dataset is the French component of the ECHP panel (except in columns (2), (3) and (4), where it is respectively the Spanish,
Italian and Danish components of the ECHP). The analysis samples are made up of individuals who work in the private sector, with a permanent contract and who are aged between 20
and 49. The total effect of the reform corresponds to β
2
in Equation (2). Standard errors in parentheses are clustered at the individual level. The BUC ordered logit coefficients in column
(6) refer to the log of the odds ratio and the number of observations is artificially higher due to the estimation method. All of the regressions control for year and individual fixed effects,
as well as age dummies (in five-year bands), health status, the (lagged) number of children in the household, the log of the monthly wage, weekly working hours, and occupation and
region fixed effects. The individual time-invariant controls are gender and education dummies. *, ** and *** indicate significance at the 10%, 5% and 1% levels respectively.
48
Table A6: The Rise in the Delalande Tax and Perceived Job Security – Heterogeneity Analysis
Women
Men
(1)
(2)
(3)
(4)
(5)
(6)
(7)
(8)
(9)
(10)
Total Effect of the Reform
-0.150
-0.181
**
-0.203
-0.036
-0.034
-0.248
***
-0.219
***
-0.302
***
-0.137
-0.138
(0.117)
(0.089)
(0.135)
(0.138)
(0.138)
(0.091)
(0.071)
(0.089)
(0.106)
(0.106)
Interacted with:
Born after 1963
-0.018
0.015
(0.154)
(0.123)
High Education
0.116
-0.061
(0.170)
(0.138)
High Monthly Wage
0.071
0.194
(0.167)
(0.125)
Parent beforehand
-0.163
-0.137
(0.165)
(0.129)
1 child beforehand
-0.192
-0.148
(0.193)
(0.156)
2 children beforehand
-0.036
-0.109
(0.184)
(0.156)
3+ children beforehand
-0.688
-0.170
(0.433)
(0.200)
Notes: These are linear regressions. The data come from the French component of the ECHP panel. The analysis sample is made up of individuals who work in the
private sector, with a permanent contract and who are aged between 20 and 49. There are 10371 observations in the whole sample (2797 individuals), 4248 observations
in the sample of women (1175 individuals) and 6123 observations in the sample of men (1622 individuals). The total effect of the reform corresponds to β
2
in Equation
(2). Standard errors in parentheses are clustered at the individual level. All of the regressions control for year and individual fixed effects, as well as age dummies (in
five-year bands), health status, the (lagged) number of children in the household, the log of the monthly wage, weekly working hours, and occupation and region fixed
effects. *, ** and *** indicate significance at the 10%, 5% and 1% levels respectively.
