We explore the determinants of the relative probabilities of labor force participation for British and Spanish married (or cohabiting) mothers. We further decompose these probabilities and find a substantial cross-national gap in participation rates which can be predominantly explained by higher wages and greater child care use in Britain.Mothers, Britain, Spain, participation.

Antonia Parera-Nicolau

Karen Mumford

English

Research Papers in Economics

 
 
 
 
 
 
 
 
 
 
 
 
Discussion Papers in Economics 
 
 
 
 
 
 
 
 
 
 
No. 2000/62 
 
Dynamics of Output Growth, Consumption and Physical Capital 
in Two-Sector Models of Endogenous Growth 
 
by 
 
Farhad Nili 
 
 
 
 
 
 
 
 
 
 
 
 
 
 
 
Department of Economics and Related Studies 
University of York 
Heslington 
York, YO10 5DD 
 
No. 2001/10 
 
The Labor Force Participation of Married Mothers in Spain and Britain 
 
by 
 
Karen Mumford and Antonia Parera Nicolau 
 The labor force participation of married mothers
in Spain and Britain.
Karen Mumford(1)
Antonia Parera-Nicolau(1,2)
(1) Department of Economics. University of York.
(2) Department of Economics.Universitat de les Illes Balears.
First version: January 2000. This version: August 2001
Abstract
We explore the determinants of the relative probabilities of labor force
participation for British and Spanish married (or cohabiting) mothers.
We further decompose these probabilities and …nd a substantial cross-
national gap in participation rates which can be predominantly explained
by higher wages and greater child care use in Britain.
Key words: mothers, Britain, Spain, participation, JEL J2
Corresponding author: Karen Mumford, kam9@york.ac.uk
Acknowledgement 1 The authors are grateful for helpful comments and
advice from Peter N. Smith and participants at the 6th Spring Meeting of
Young Economists (SMYE), Copenhagen, March 2001. Responsibility for
any mistakes or omissions is entirely our own.
11 Introduction
The relative proportion of women to men participating in the labor force has
been consistently smaller in Spain than it has in Britain. With the opening of
the Spanish economy and the relaxation of the regime’s moral controls in the
60’s, the participation rates of Spanish women increased slightly but steadily.
Since the …rst democratic elections in Spain, in 1979, the gap between Spain
and Britain has been decreasing at a higher rate. Nevertheless, there still exists
a substantial di¤erence in female participation rates between Spain and Britain
(Costa, 2000). This is especially true for married mothers, by the early 1990s
some 55% of British married mothers were participating in the labor market
whilst only some 32% of Spanish married mothers were.
In this paper we explore the determinants of the relative probabilities of la-
bor force participation for British and Spanish married (or cohabiting) mothers.
We further decompose these probabilities to consider the component due to dif-
ferences in recognizable characteristics and that component associated with dif-
ferences in response to these characteristics. In order to do so, we use individual
data on British and Spanish family’s members and implement a decomposition
devised by Even and Macpherson (1990 and 1993) in their analysis of declining
union membership. As the process of integration continues across these Euro-
pean countries, we also consider the implications of our results by indentifying
how policy might increase the participation rate in Spain.
Explanation of the econometric framework is presented in section 2, the data
sets are described in section 3, discussion and results of the probit models are
provided in section 4, section 5 includes analysis of the labor force participation
gap, and conclusions are presented in section 6.
2 The Econometric Framework
We estimate the following probit model of labor force participation (LFP):
2LFPij =
½
1 if Y ¤
ij > 0
0 if Y ¤
ij ￿ 0
¾
(1)
where Y ¤
ij = Xij¢¯j+uij is a latent (unobserved) variable which measures the
net utility gain of labor force participation for woman i of country j; j = (b;s),
b representing Britain and s for Spain; Xij is the vector, for each woman i, of
the variables thought to in‡uence LFP1; ¯j is the vector of parameters, who
measure the in‡uence on LFP of each explanatory variable in X ; and uij are
the error terms (assumed to be normally distributed).
The estimated probability that a female i of country j participates in the
labor market is:
^
Pij = ©
µ
Xij ¢
^
¯j
¶
; (j = b;s) (2)
where
^
¯j is the probit estimate of the parameter vector ¯j and © is the
standard normal cumulative density function. Following Even and Macpherson
(1990 and 1993), the average estimated probability of female LFP by country
is given by:
_
^
Pj =
1
Nj
¢
Nj X
i=1
©
µ
Xij ¢
^
¯j
¶
; (j = b;s) (3)
where Nj is the number of women in the country j sample.
The estimated probability of LFP if there are no di¤erences in response to
characteristics across British and Spanish women is:
1We follow the majority of authors investigating labour supply and adopt a standard
static work-leisure tradeo¤ model (see Killingsworth 1983 and the recent survey in Blundell
and Macurdy, 1999). Each individual is assumed to maximise a quasi-concave utility function
U(C, L, X ) where C, L, X are consumption, leisure, and individual attributes respectively.
3_
^
Po ´
1
Ns
¢
Ns X
i=1
©
µ
Xis ¢
^
¯b
¶
(4)
The British-Spanish women´s LFP-gap can be decomposed into two terms:
_
^
Pb ¡
_
^
Ps =
Ã _
^
Pb ¡
_
^
P0
!
+
Ã _
^
P0 ¡
_
^
Ps
!
(5)
where
Ã _
^
Pb ¡
_
^
P0
!
represents the portion of the gap associated with di¤er-
ences in characteristics which are expected to in‡uence labor force participation
(the explained component of the gap); and
Ã _
^
P0 ¡
_
^
Ps
!
is associated with di¤er-
ences in the impacts (or responses) of those characteristics on the probability
of labor force participation (the unexplained component of the gap).
The contribution that each individual explanatory variable k makes to the
explained portion of the total gap is given by:
_
^
Pek =
Ã _
^
Pb ¡
_
^
P0
!
¢
2
6 6 6
4
µ _
X
k
b ¡
_
X
k
s
¶
¢
^
¯
k
b
³ _
Xb ¡
_
Xs
´
¢
^
¯b
3
7 7 7
5
(6)
As discussed in Doiron and Riddell (1994), this procedure allocates shares
according to the relative size of the explanatory variable’s impact on participa-
tion2.
3 Data
The data for the British families are taken from the 1990/91 General Household
Survey (GHS). The GHS is an annual, multipurpose survey based on a sample
2Doiron and Riddell (1994) present an alternative decomposition which explicitly allows
for curvature of the probability of the dependent variable (in their case unionisation). We also
considered their decomposition and found little impact on our estimates (results are available
on request from the authors). Applying Okum’s Razor, we adopt the simpler decomposition.
4of around 10,000 private households in Great Britain. Interviews are conducted
with household members aged over 16. The 1990/91 year of this survey contains
a unique section on household child care choices necessary for the estimation
below.
For the Spanish families, the data has been obtained from the 1994/95 Eu-
ropean Community Household Panel (ECHP).The ECHP survey provides com-
parable micro-level data to the GHS3, and interviews are similarly conducted
with household members aged over 16. The 1994/95 sample is the earliest year
of data for this survey
From both data sets, we select all families in which the parents are married
or cohabiting, not self-employed, and in which there is at least one child aged
under 12. Once removing those observations for which complete information is
not available, we are left with 1531 families in the British sample and with 1113
in the Spanish.
The probit models of labor force participation (equation 2) that we estimate
include the following explanatory variables in the vector X :
² variables describing the mother’s human capital: age, age squared, uni-
versity quali…cation for mother (degree or higher), college quali…cation for
mother;
² a variable referring to the father’s human capital: university quali…cation
for father;
² variables re‡ecting opportunity cost: family’s unearned income (including
father’s labor income), and predicted net hourly wage (for the mother);
² a variable re‡ecting the demand-side of the labor market: unemployment
(regional unemployment rate);
3There is also a British sample in the ECHP. The sample size is, however, is too small to
obtain consistent and e¢cient estimates in our model.
5² variables describing the mother’s availability to participate in the labor
market: number of children aged under 5 years , number of children aged
between 5 and 12 years, number of adults present in the household, grand-
parent present in the household, use of formal child care, use of informal
child care only (unpaid for care o¤ered by friends and/or relatives)4;
² a variable referring to the father’s working arrangements: father’s work
pattern (not working, working part-time or working full-time);
² a variable referring to the marital status of the mother: cohabiting.
Means are provided in Table 1 (fuller discussion of the data is included
in an appendix available from the authors). In general, Spanish women are
slightly older; more likely to be formally married; are better educated with
fewer children; are less likely to use formal child care but are more likely to have
other adults including grandparents living in their home; have substantially
lower wages; and face higher regional unemployment rates than their British
counterparts.
4 Estimation of the Probit Model of LFP
Results for the probit estimates are presented in Table 15 (we report marginal
e¤ects rather than raw coe¢cient estimates, except for binary variables where
di¤erential e¤ects are reported). Columns 1 through 3 of Table 1 provide re-
sults for Britain, whilst columns 4 to 6 are for Spain. The overall test of the
explanatory power of the regressors is clearly signi…cant for all the regressions
and whilst the pseudo R2 measures are not high, they are comparable with those
4In both countries, there are many women working who do not appear to be using child
care, more so in Spain than Britain, this may be due to a view by survey respondents that
child care provided by families members does not classify as informal child care.
5The reference group for our analysis are those mothers who are married; left education
just after school; with a husband who did not go to university; are unemployed; have at least
one child aged under 12; with families who do not use formal or informal child care; and do
not have a grandparent living with them.
6found in other studies of labor force participation (see Blundell and Macurdy,
1999). Overall, the parameter estimates are generally well de…ned and of the
expected sign. We consider the results in more detail by addressing the impact
of the right hand side variables in turn.
What matters most for labor force participation in our results are the positive
impact of the mother’s own wage; the positive impact of the use of child care
(more so for formal child care than informal); and the negative e¤ect of number
of children in the household who are aged under 5 years old. These results are
true for both British and Spanish women. To a lesser extent, both British and
Spanish mothers are also less likely to participate in the labor market if their
alternative sources of family income (including their partner’s wage) are high.
For British women, we also found that older women are more likely to par-
ticipate (although at a decreasing rate), as are women with partners working
full time. Whereas university educated mothers in Britain are less likely to par-
ticipate (18% less likely), so are those with university educated partners (12%
less likely). This tendency of well educated women in developed countries to
lower their labor force participation rates during their child bearing years has
been recognized in several recent studies (Costas, 2000) and perhaps re‡ects the
di¢culty these women have …nding child care of a similar quality to that they
can provide themselves. The greater availability of family-friendly work prac-
tices such as extended maternity leave and career breaks may also encourage
the lower participation of well educated mothers in Britain
In neither countries’ estimates do the legal status of the partnership
(married or cohabiting), the number of adults present in the household, the
presence of grandparents, or the regional unemployment rate have a signi…cant
impact on the probability of the mother participating.
The average estimated probability of LFP for a married or cohabiting mother
of children aged under 12 is 0.54 in Britain. In Spain that estimated probability
7is, as expected, smaller at 0.32 (…nal row, Table 1). The total predicted LFP gap
(the di¤erence in estimated probabilities of LFP) between British and Spanish
women is therefore 0.22. Of this gap, 0.28 (or 124%) is due to di¤erences in
observed characteristics and -.05 (or -24%) is due to di¤erent responses to these
characteristics across the countries. In other words, if Spanish women had the
same observed characteristics as British women their participation rates would
exceed those of the Brits by almost 25%.
In addition to these general conclusions, we have also calculated the contri-
bution of each explanatory variable to the explained portion of the LFP Gap
(Table 1, column 7). As mentioned above, the own wage e¤ect is very strong
in our results. (In order to make allowance for only having wage information
on working mothers, we estimated a predicted wage for all mothers by max-
imizing a full information likelihood function, Blundell and Macurdy, 19996.)
If we increased the log of the Spanish womens’ average predicted wage by one
unit (bringing it to a very similar mean value as for British women, which is
equivalent to £3.64 per hour), the participation rate for Spanish women would
rise by 0.21 percentage points, removing the majority of the LFP gap across the
two nationalities. This is re‡ected in column 7 where the wage e¤ect makes up
69% (0.19/0.28) of the explained portion of the gap. Similarly, the use of child
care is related to 37% of this explained portion. Increasing the Spanish wage
and levels of child care to those levels o¤ered in Britain would, ceteris paribus,
result in dramatic increases in the participation of Spanish women. In contrast,
the total fall in participation amongst university educated mothers and those
with university educated partners constitutes only some -6% of this portion.
The unexplained portion of the gap is more di¢cult to interpret (Jones,
1983) and could be due to many factors omitted from our analysis For example,
cultural and legal di¤erences across countries will impact on the labor supply
6Discussion of this procedure and our results are included in the appendix available from
the authors.
8choices of mothers. These include the already brie‡y mentioned greater avail-
ability of family friendly work practices to British women; the possibly lower
quality child care provision in Britain; and the comparatively recent large scale
involvement of women in the Spanish labor market. As the process of inte-
gration continues across Europe, we would expect many of these di¤erences to
become further eroded.
5 Conclusions
We have found that married (or cohabiting) mothers of children aged under 12
in Spain are substantially less likely to participate in the labor market than are
similar British mothers. The majority of the gap in their predicted participation
gaps can, however, be explained predominantly by higher wages and greater
child care use in Britain. If the process of uni…cation across Europe leads to
common working conditions being introduced in the two countries, we expect
the size of this participation gap to fall dramatically generating many extra
productive workers in Spain.
References
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[3] Costa, Dora L., 2000. From mill town to board room: The rise of women’s
paid labor, Journal of Economic Perspectives, 14(4), 101-122.
[4] Doiron, Denise J., and Craig W. Riddell. 1994. The impact of unionization
on male-female earnings di¤erences in Canada, Journal of Human Resources,
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9[5] Even, William E., and David A. Macpherson. 1990, Plant size and the decline
of unionism, Economic Letters 32, 393-398.
[6] —– 1993, The decline of private-sector unionism and the gender wage gap,
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10TABLE 1.   Probit estimates of the determinants of mothers’ labour force participation.
Variables Britain  Spain Decomposition
effect t value  mean effect t value mean P0=P(Xbâs)
mother's age 0.055  2.46** 33.09  0.023  0.92  35.30  -0.048
mother's age squared -0.001  -2.15** 1135.01  -0.000  -1.13  1283.42  0.054
mother's university qualification -0.178  -1.93* 0.07  0.117  0.88  0.19  -0.013
mother's college qualification -0.026  -0.46 0.19  0.012  0.14  0.19  0.000
father’s university qualification -0.116  -2.53** 0.14  0.033  0.74 0.23  -0.003
ln(family unearned income) -0.188  -6.02*** 5.44  -0.074  -2.79***  5.13  -0.022
ln(mother's hourly wage) 0.442  3.49*** 1.27  0.206  2.09**  0.29  0.193
regional unemployment rate -0.003  0.29 8.44  -0.003  -1.00 23.56  0.052
number children aged under 5 -0.251  -8.86*** 0.78  -0.117  -3.77***  0.52  -0.029
number children aged 5 to 12 -0.032  -1.54 1.74  -0.005  -0.18  1.47  -0.001
number adults present in household -0.035  -0.99 2.14  0.038  1.60  2.44  -0.011
grand parent present in household 0.144  1.25 0.02  -0.053  -0.79  0.08  0.004
formal child care only 0.331  9.27*** 0.35  0.431  8.65***  0.13  0.080
informal (unpaid) child care 0.293  7.25*** 0.15  0.361  6.21***  0.08  0.023
father's work pattern 0.254  9.12*** 1.70  0.019  0.65  1.71  0.000
mother cohabiting (not married) -0.047  -0.84 0.08  0.017  0.16  0.02  0.001
Number of observations  1531 1113 Total explained 0.278
LR chi2(15)    452.04  283.77 Total unexplained -0.054
Log likelihood -829.05  -556.45  Total predicted gap 0.224
Pseudo R2    0.2142  0.2032
obs. P  0.5447  0.3207
pred. P   (at x-bar) 0.5435  0.2921
Marginal effects are reported except for binary regressors when differential effects are reported.   
*** significant at 99%,
** significant at 95%, 
* significant at 90%.

The Labor Force Participation of Married Mothers in Spain and Britain.

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