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The Strengthening Impact of Earnings on the Transition to Parenthood

Evidence from Norway 1994-20091

Rannveig V. Kaldager2 rvk@ssb.no

Paper prepared for the XXVII IUSSP International Population Conference, Busan, Republic of Korea

26.-31. August 2013

Preliminary – please do not cite or quote

Abstract This study describes how the impact of earnings on the transition to parenthood changes in

the period 1994-2009 in Norway. By comparing changes in the earnings-fertility relationship

for men and women, I cast light on whether the mechanisms linking earnings and the

transition to parenthood have become similar for men and women over time. Discrete-time

hazard regressions are estimated on highly accurate data from the Norwegian population

registers, covering all men and women at risk of having a registered first birth in the period

1994-2009. Results show a monotonously positive and substantial impact of annual earnings

on first birth rate for men throughout the period. The impact of earnings on the transition to

parenthood strengthens over time for both men and women, but more so for women. At the

end of the period, the earnings-first birth relationship is similar for men and women. The

results show that despite generous benefit schemes and stable economic growth, earnings

have become increasingly important for the transition to parenthood. The similarity across

gender indicates that the mechanisms linking earnings and fertility in Norway have become

gender neutral.

1 I am grateful to Torkild Hovde Lyngstad, Øystein Kravdal, Trude Lappegård, Marit Rønsen and Arnstein Aassve for

helpful comments. 2 Research Department, Statistics Norway and Institute of Sociology and Human Geography, University of Oslo.

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1. Introduction

Position in the labour market affects life chances through several mechanisms – one of which

is the impact of earnings on the transition to parenthood. In his microeconomic theory of

fertility, Becker (1991) argues that this mechanism operates in opposite ways for men and

women: For men, the impact of earnings on the transition to fatherhood is a process of

cumulative advantage, where doing well in the labour market facilitates starting a family. For

women, the earnings-fertility relationship is expected to be negative, as taking time off work

to raise children is more costly when earnings are higher. Challenging the assumption of

gender specialisation, Oppenheimer (1997) argued that the mechanisms linking labour market

position and family formation are similar for men and women, giving a process of cumulative

advantage for women as well as men. As women increase their efforts in paid work and men

increasingly partake in unpaid work, the degree of gender specialisation in the family has

decreased in Western societies (see e.g. Thévenon 2009, Hook 2006). These trends are

particularly evident in the Scandinavian countries (Leira 2006), where the state actively

supports men’s participation in childrearing and mothers’ labour force participation. Thus,

gender-neutral theories are expected to be increasingly relevant to explain the earnings-

fertility relationship, while theories based on gender specialisation are expected to loose

predictive power over time.

Cross-country comparisons (e.g. Andersson, Kreyenfeld and Mika 2009, Berninger 2013)

indicate that contextual factors, such as the generosity and design of welfare schemes, may

shape the impact of earnings on fertility. Some such welfare schemes, like cash allowances

for parents and free or subsidized schooling and health care, reduce the cost of childbearing

and may thus weaken the impact of earnings on transition to parenthood. On the other hand,

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welfare schemes that reduce the alternative costs of childbearing – such as low-cost high

quality child care – may strengthen the impact of earnings on transition to motherhood, as

high-earning women who would otherwise have remained childless choose to enter

parenthood. Higher overall societal wealth may decrease the impact of earnings on fertility

due to loosening budget constraints, or increase the impact of earnings on fertility if

consumption preferences grow faster than budget constraints loosen. Though contextual

factors vary both over time and between countries, variation over time has rarely been used to

cast light on the mechanisms linking earnings and fertility.

This study is based on using highly accurate register information on the annual earnings and

first births of all Norwegian men and women born 1955-1988 who were at risk of a first birth

in the period 1994-2009 (N ~ 12 million person years). I estimate the impact of earned income,

measured in quintiles, on the transition to parenthood, using discrete time hazard regression.

The extraordinarily rich data set allows for estimating separate models by year and gender and

for describing changes over time separately by gender. The main contribution of the study is a

comparison of the impact of earnings on first birth rates across sexes and periods. Arguing

that the degree of gender specialisation in Norway has decreased in the period of study, I will

explore whether this has led to the earnings-fertility relationship becoming more similar for

men and women.

2 Theoretical perspectives on earnings and the transition to

parenthood

Earnings are expected to affect the transition to parenthood through three main mechanisms:

Firstly, earnings potential may affect whether men and women are attractive as partners.

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Secondly, his and her earnings may affect both whether and when couples have a first child.

Finally, the propensity to have first child without living with a partner may vary with earnings

potential.

2.1 Earnings and union entry and stability

First birth rates are higher among individuals who are currently in a co-residential relationship,

both because of pre-relationship fertility desires and because of a jointly developed desire for

children (Wetlesen, 1991; Marsigilio, 2007). For men, both the theory of gender specialisation

(Becker, 1991) and the theory of pooling of resources (Oppenheimer, 1997) predict that

higher earnings facilitate union entry and stability. These theories differ, however, in their

prediction for women: Becker (1991) assumes that high-earning women have less to gain

from union formation and therefore have a lower propensity to form unions and higher

propensity for union dissolution. In the theory of pooling of resources, it is assumed that men

as well as women prefer high-earning partners, and a positive impact of earnings on union

entry and stability is therefore expected to appear for both her earnings as well as his earnings.

For men, the theoretical prediction has been supported by several empirical studies showing

that men’s earnings potential is associated with higher marriage rates (Sweeney, 2002; Xie et.

al., 2003; Kalmijn and Luijkx, 2005; Petersen et. al., 2011) and lower divorce rates (Hoffman

and Duncan, 1993; Jalovaara, 2003; Lyngstad, 2004; Kalmijn et. al., 2007). For women,

empirical results are more mixed: While Sweeney (2003) and Jalovaara (2012) find that

economic resources have a positive impact on women’s propensity to marry, Xie et. al. (2003)

find no such positive impact. As predicted by the theory of gender specialisation, women’s

higher earnings is generally found to elevate divorce risk (Lyngstad (2004) for Norway,

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Jalovaara (2003) for Finland and Kalmjin et. al. (2007) for the Netherlands), though Hoffman

and Duncan (1993) find no impact of wives’ wages on divorce risk in the US.

In the period of study, childbearing to cohabiting couples has become increasingly common in

Norway (e.g. in 2011 49 per cents of first births were to cohabiting parents, compared to 32

per cent to married couples3). A positive impact of earnings on entry into cohabitation is

found in the Nordic countries for men as well as women (Bracher and Santow, 1998;

Jalovaara, 2012). As for marriage, men’s higher earnings potential has consistently been

found to stabilize cohabitating unions (Jalovaara, 2011; Texmon, 1999; Brines and Joyner,

1999; Kalmijn m.fl., 2007). While the impact of women’s earnings on marriage stability seem

to be largely in line with the specialisation theory, empirical results for earnings and

cohabitation dissolution lends more support to the theory of pooling of resources: If the

earnings level of the spouses are similar, the risk of cohabitation dissolution is reduced

(Kalmijn et. al. 2007, Brines and Joyner 1999, Jalovaara 2011).

2.2 Earnings and household fertility decisions

According to conventional microeconomic theory of fertility (Becker, 1991), the earnings

potential of the spouses affects a couple’s demand for children through two opposing

mechanisms: First, children are costly – both in terms of time and money. Higher earnings

make it possible both to cover monetary costs of childrearing, and to reduce working hours to

take care of children while maintaining an acceptable standard of living. This leads to a

positive income effect of earnings on fertility. Second, the cost of childbearing is mainly

driven by the cost of the time allocated to childrearing, and taking time off work is more

costly when wages are higher. This substitution mechanism represents a negative effect of 3 http://www.ssb.no/emner/02/02/10/fodte.

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earnings potential on fertility. Under gender specialisation, the income effect dominates for

men, while the substitution effect dominates for women.

In Becker’s theory of income and fertility, earnings potential is treated as a static

characteristic that affect the preference for total number of children (completed fertility). In

the Norwegian context, with low levels of childlessness and generous welfare schemes that

cover a large proportion of the cost of childrearing, earnings are expected to affect when

couples have a first child rather than whether they have a child4. Dynamic theories of fertility

address the impact of earnings on timing decisions (Hotz et al 1997). Under complete gender

specialisation, fertility timing is optimal when his earnings are as high as possible, while her

opportunity costs are minimized. As earnings increase over the life course, his earnings are

highest if the first birth is postponed until the end of fertile years (Happel et. al. 1984). The

optimal timing of motherhood, with respect to minimizing opportunity costs, is less clear

(Gustavsson 2001). Under gender equal division of labour (or partial gender specialisation),

optimal timing of parenthood must take both opportunity costs and income effects into

account with respect to both spouses’ earnings.

In the Nordic context, with high work-family compatibility and thus lower opportunity costs

of childbearing, empirical studies have found a positive correlation between female earnings

and the transition to parenthood (Andersson 2000 (Sweden), Vikat 2004 (Finland), Andersson,

Kreyenfeld and Mika 2009 (Denmark)). Using predicted hourly wage, Rønsen (2004) finds a

weak negative impact on the transition to motherhood in Norway, and Grott and Pott-Butter

(1992) obtain a similar result for Netherland. Heckman and Walker (1990) find a strong

negative impact of female wages on the transition to motherhood in Sweden, though Tasiran

4 A possible exception is that the substitution effect could lead some top-earning women to forgo motherhood.

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(1994) argues that a non-significant or even positive impact is found for Sweden using better

data on female wages. All in all, while observed earnings are positively correlated with the

transition to motherhood in the Nordic context, predicted wages have a weaker positive or

even negative impact on the same transition. As earnings are plausibly endogenous to the

fertility decisions, it is unsurprising that models using predicted (i.e. plausibly exogenous)

wage rates may yield different results. Contrary to this expectation, Kravdal (1994) found that

among Norwegian women, the importance of predicted wages and actual wages on the

transition to motherhood are strikingly similar. However, as women’s labour supply decisions

are expected to be highly endogenous to fertility decisions, and yearly earnings reflect labour

supply decisions while wages do not (or to a much lesser extent), endogeneity problems could

be more severe in studies using observed earnings than when using observed wages are used.

Heckman and Walker (1990) find a weak positive impact of men’s income on the transition to

fatherhood in Sweden, but the estimate is insignificant after control for marital status.

Similarly, Merrigan and St.-Pierre (1998) find no significant impact of men’s earnings on first

birth rate in Canada after controls for woman’s wages and marital status. Re-estimating

Heckman and Walkers models with better data, Tasiran (1994) find that the impact of men’s

income is negative in some specifications. Lappegård and Rønsen (2013) find positive impact

annual earnings on the transition to parenthood. Overall, while the empirical studies of male

fertility are scarcer, they seem to support the theoretical expectation that earnings either

facilitates, or is less important for, the transition to fatherhood.

2.3 Earnings and childbearing outside unions

Non-union childbearing is associated with socioeconomic disadvantage among women (see

e.g. Perelli-Harris et al., 2010). Particularly, Kravdal (1994) find that women with lower

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wages have higher risk of having a non-union first birth. Research on men’s non-union

childbearing is scarce, but points in the same direction. Non-residential fatherhood is

associated with lower socioeconomic status (Nelson, 2004; Skrede, 2004), and though this is

partly due to the socioeconomic gradient in union dissolution risk, it also indicates that non-

union childbearing is associated with socioeconomic disadvantage. Previous studies indicate

that men with lower income have higher risk of contraceptive failure (see Nelson (2004) for

an overview). As non-union births are more often unplanned (Hayford and Guzzo, 2010), and

thus likely to more often be a result of contraceptive failure, men with lower income may

have an elevated risk of fathering an unplanned child. A conception outside union does not

necessarily lead to a non-union birth if it is carried to term, as the parents-to-be may choose to

form a union before the child is born. Expectant mothers may be more interested in forming a

union if the father-to-be has higher earnings than if he has lower earnings (Ermisch, 2003).

This would also give a higher risk of non-union birth among men with lower earnings

potential.

Though the impact of earnings on non-union childbearing is expected to be negative for men

as well as for women, it should be noted that this mechanism will be of modest importance for

the overall impact of earnings on the transition to parenthood, as only approximately 1 in 10

first births take place outside unions in the period of study.

2.4 Possible sources of bias

For men, the estimated impact of earnings on fertility is expected to be biased upwards, for

two main reasons: Firstly, an intention to enter fatherhood could make men to increase their

effort in paid work. Secondly, some unobserved personal characteristics (such as willingness

to work hard, good health etc.) are expected to be positively correlated with both earnings and

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entry into fatherhood. Comparing estimates for plausibly exogenous earnings measures (e.g.

Heckman and Walker 1990) with estimates for individual earnings (e.g. Andersson and Scott

2007) supports this expectation of upwards bias. For women, the expected direction of bias is

less clear. Anticipatory effects – the phenomenon that women may reduce their efforts in paid

work because they expect to have a child soon – will bias the estimates downwards. The fact

that studies using predicted wages overall yield weaker estimates than studies using annual

earnings points towards that the latter earnings measure may give estimates that are biased

upwards.

2.5 Summary and expected associations

A positive association between earnings and the probability to enter fatherhood is expected,

for two main reasons. First, men with higher earnings potential are likely more attractive as

partners, as shown by their higher union entry and lower union dissolution rates. Second,

couples postpone the transition to parenthood until his earnings are relatively high, to have

sufficient income to cover the costs of childrearing. As the estimates are expected to be biased

upwards, they can be interpreted as an “upper bound” of the causal effect of earnings on the

transition to fatherhood in the Norwegian context (see Manski 1995).

For women, the direction of the expected association between earnings and fertility depends

on whether the income effect or the substitution effect dominates. If the income effect

dominates, the impact of her earnings on the transition to parenthood will be positive, while it

will be negative if the substitution effect dominates. In addition to these mechanisms,

selection may bias the estimates for women in either direction. In line with previous studies

from the Nordic context, I expect the impact of women’s earnings on the transition to

motherhood to be positive.

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3 Theoretical perspectives on change over time

This section discusses three types of societal changes that may have changed the impact of

earnings on the transition to parenthood over the period of study 1994-2009: Changing

fathering practices, changes in welfare schemes and changes in the cost of and relative

preference for children.

3.1 Changing fathering practices

In the period of study, actively encouraged by family policies, the time fathers spend on

childcare and housework has increased (Kitterød 2012). In fact, a small fatherhood wage

penalty has emerged (Cools and Strøm 2011), indicating that parenthood now implies

alternative costs for men as well as for women. The emergence a opportunity costs for men

could affect men’s preferences for the timing of parenthood. While postponement maximizes

household income at the transition to fatherhood, having a child earlier in the career could

reduce (or be perceived to reduce) opportunity costs. It is, however, questionable that the

magnitude of the wage penalty is sufficiently strong to be of major importance for such

timing decisions.

From 1993 to 2004, there has also been a minor increase in both the proportion of mothers

who are working, and the number of hours worked by employed mothers (Ministry of

Children and Family Affairs 2007). Studying the period 1996 to 2010, Rønsen and Kitterød

(2012) find that women return faster to work over time. In theory, this stronger attachment to

the labour market could affect opportunity costs of childbearing, and thus fertility timing

decisions. However, the observed increase in mothers’ employment and working hours is

arguably to small to cause of a marked change in fertility timing.

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3.2 Welfare schemes and changes in the cost of childrearing over time

While welfare schemes such as cash transfers and tax allowances reduce the monetary cost of

childbearing, the alternative (time use) cost of childbearing is reduced by schemes such as

available high-quality childcare (McDonald 2002). If the monetary cost of childrearing is

reduced, I expect earned income to be less important for the transition to parenthood.

Reducing the opportunity cost of childbearing is expected to strengthen the impact of

earnings on the transition to motherhood, as it may induce career-oriented women who would

otherwise have forgone childbearing to have a child. This section gives an overview of

changes in (the relative value of) welfare schemes that affected monetary and non-monetary

cost of childbearing in the period of study, and the expected consequences of these changes

for the earnings-first birth relationship.

[Figure 1 about here]

Three welfare schemes were central in reducing the monetary cost of childbearing in Norway

the period of study. Child allowances are given from the first child, and are not means tested.

While these child allowances made a substantial contribution to covering the monetary cost of

childbearing in the beginning of the period of study, they have not been indexed to the growth

in real wages and consumption (NOU 2004 and 2009). Figure 1 shows that the CPI-adjusted

value of the child allowance has declined in the period of study. In 1998, a cash-for-care

allowance was introduced for 1-year (and, from 1999, 2-year)-olds who were not enrolled in

publicly financed child care (Bakken and Myklebø 2010). Though the nominal value of the

cash-for-care benefit has fluctuated slightly after its introduction, the CPI-adjusted value of

the benefit has remained largely unchanged. Finally, mothers who have not earned rights to

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parental leave allowance receive a lump-sum transfer (“engangsstønad”) upon birth. After a

marked increase in the beginning of the period, the nominal value of this lump-sum transfer

has been virtually unchanged through the period. After adjusting for price growth, the value

of this lump-sum transfer is found to decrease over time. As transfers cover a smaller share of

the monetary cost of childbearing over time, earned income is expected to be increasingly

important for the transition to parenthood.

While every second child aged 1-5 were enrolled in kindergarten at the beginning of the

period of study, the proportion had risen to 9 out of 10 in the same age group in 20095. Using

Norwegian data for the period 1973-1998, Rindfuss et al (2010) find that childcare

availability has a positive impact on fertility at all parities. As childcare reduces the

alternative cost of childbearing, it weakens the (negative) substitution effect of women’s

wages on fertility. Thus, increased childcare availability is expected to strengthen the impact

of women’s earnings on fertility.

Women who have been employed for at least 6 of the 10 last months before giving birth are

eligible for parental allowance in Norway. In 1993, the parental allowance gave 100%

earnings replacement for 42 weeks, or at 80% replacement for 52 weeks6. During the period

of study (up to 2008) the maximum length of the paid leave and the fathers’ quota were each

expanded by 2 weeks, leaving the maximum available share for the mother unchanged

(Bringedal and Lappegård 2012). The value of the parental allowance far exceeds that of the

lump-sum transfer for women who work full-time or long part-time. The value of the parental

allowance increases with the growth in real wages. Thus, the value of the parental allowance 5 http://www.ssb.no/a/minifakta/no/main_07.html#fig0701 6 Earnings are not replaced over a cap of six times the National Insurance basic amount (approximately 50 000 EUR in 2009)

employers top up the replacement for social security if necessary. (http://www.nav.no/805369034.cms). Some employers “top up” the parental allowance for employees earning above this cap.

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relative to the lump-sum transfer has increased over time. This provides an incentive for

women to gain foothold in the labour market before entering parenthood – which would

strengthen the earnings-first birth relationship for women.

Overall, changes in benefit schemes in the period of study are expected to have strengthened

the impact of earnings on fertility, for two main reasons: First, as the value of schemes that

reduce the monetary cost of childbearing has declined over relative to the value of earned

income, the income effect is expected to have become stronger. Second, the increasing

availability of child care has reduced the opportunity cost of childbearing, dampening the

(negative) impact of the substitution effect for women.

3.3 Increased investment in child quality and changing relative preferences

Using data from the US, Kornrich and Furstenberg (2013) find that parental spending on

children has increased over the last 40 years. Gauthier et al (2004) find that the time parents

spend with children has increased in the same period in 16 industrialized countries. Thus, the

the cost of childrearing, both in terms of time and money, has increased. One strategy to

obtain sufficient income to cover these costs is to postpone childbearing to later ages when

earnings are higher. Higher household income can be invested in high-quality childrearing, or

allow for reduced working hours to invest time in children while still maintaining a relatively

high standard of living.

Throughout the period of study, there has been a general upsurge in wealth in Norwegian

society. The loosening of budget constraints is expected to weaken the earnings-fertility

association, but this holds only if the overall level of wealth increases more quickly than

consumption preferences: As pointed out by Crimmins et al (1991), the combination of a

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stronger preference for consumption and a constant preference for children will lead to a

weaker relative preference for children – that is, the actual preference for children when

alternative costs are taken into account will decrease. A stronger preference for consumption

may lead couples to postpone childbearing to a later age where the forgone consumption

associated with childbearing is lower.

3.4 The welfare state versus real wages

Changes in the generosity of welfare schemes are expected to have strengthened the impact of

earnings on fertility for women over the period of study, while changes in fathering practices

may have weakened the impact of earnings on the transition to fatherhood. Thus, the impact

of earnings on the transition to parenthood is expected to have become more similar across

gender. Increasing investment in child quality and increasing consumption preferences could

contribute to earnings remaining important for the transition to parenthood.

4 Method and data

4.1 Data

The analysis is based on data on births, earnings (defined as the sum of earnings from

employment and primary and secondary business income), unemployment benefits, health

related benefits, and educational level/enrolment for all men and women born 1955-1987

from the Norwegian population registers. The data set further is restricted to persons who

have at least one Norwegian-born parent, who are Norwegian citizens, and who did not have a

first child before age 20 or year 1994. First births are observed in the period 1994-2009, and

observations are censored at whatever occurs first of a first birth, age 50 or calendar year 2009.

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The study sample consists of 5 127 672 person years for women, and 6 986 150 person years

for men.

4.2 Method

Discrete-time hazard regression models for first birth rates are estimated with the baseline rate

(hazard) specified as a linear spline with 5-year knots. After data are transformed into person

years, logistic regression models are estimated in Stata, using the logit command. To facilitate

comparison across models (Mood 2010), and to ease the interpretation of interaction terms,

results are reported as average marginal effects, computed with the command margins,

dydx().

4.3 Variables

Earnings quintile is defined based on the position in the earnings distribution relative to all

individuals (i.e. both parents and (currently) childless persons) of the same sex and age in the

same year. Calculations are done separately by year and age to avoid that the earnings

variable captures period and age effects. Missing earnings are included as a separate category.

Educational attainment and enrolment may confound the association between annual earnings

and first birth risk, and are thus included as controls. Persons who are in education for at least

4 months of a year are defined as students during that year. I also included controls for

aggregate unemployment and reception of unemployment benefits. A dummy for reception of

disability pension or rehabilitation transfers was constructed to capture health limitations that

affect earnings potential7. Calendar time is included as a categorical variable with 4-year

7 The health dummy is based on a measure from FD Trygd, which includes old age pensions as well as child allowances.

However, as childless persons under age 50 do not have the right to neither of these additional benefits the measure

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categories. A set of dummies for region of birth is included, to capture regional variation in

earnings level and fertility that may confound the estimates for earnings.

A couple’s decision to get married may result from an intention to have a first child, and if so,

a control for marital status would be a control for an intention to have a child (see Rindfuss

and St. John (1983) for a discussion of this). Including marital status in the model would then

control out any indirect effect of earnings potential on fertility that is mediated by marriage.

For these reasons, controls for marital status are omitted. A covariate for marital status would

also make comparisons over time less clear due to the increase in first births to cohabitants:

Non-marital births in the first part of the period will have to a larger extent be non-union

births than non-marital births towards the end of the period (Noack 2010:30).

5 Results

Summary statistics of person years are shown in table 1. Persons with missing information on

earnings and educational level are included in the analysis, using missing as a separate

category. The mean yearly probability of a first birth is 5 per cent among men, and close to 7

per cent among women, reflecting that women on average have a first child earlier than men.

[Table 1 about here]

Figure 2 shows mean CPI-adjusted earnings within each earnings quintile over the period of

study. The most striking feature of the figure is the overall growth in purchasing power, for

men as well as women. The figure also shows an increase in earnings inequality over time, as

constitutes a fairly good proxy for reception of health benefits in this group. “Sykepenger”, which is given for the first year of sickness absence, is included in the earnings variable.

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the difference in mean earnings between the earnings quintiles increases over period. While

mean earnings increase over time, the CPI-adjusted value of cash transfers has been stable or

decreasing (Figure 2). Thus, the value of cash transfers relative to earnings has decreased

markedly. This illustrates that the value of benefits linked to earnings, such as parental

allowance, has increased relative to the value of other cash transfers in the study period.

[Figure 2 about here]

5.1 The impact of earnings on the transition to parenthood

Results from the main model (with all controls) are shown in table 2 as average marginal

effects. The impact of earnings on the transition to fatherhood is positive, of considerable

magnitude, and increases monotonously with earnings quintile. The impact of earnings on the

transition to motherhood is positive, but not monotonous: The first birth probability is lower

among women in the 2nd earnings quintile than in the lowest. The (relatively) high first birth

probability of women in the lowest earnings quintile could plausibly be driven by anticipatory

effects, that is, women who intend to have a child in the near future may decrease their efforts

in paid work. For the higher quintiles, the estimates for men and women go in the same

direction, but the impact of earnings is stronger for men. It is also noteworthy that missing

earnings are correlated with low first birth probabilities among both men and women.

[Table 2 about here]

The positive estimate for having received unemployment benefits may at first sight seem

surprising. However, this variable also captures an income effect, where those who are

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unemployed and have earned rights to such benefits both have higher income, and possibly

are more likely to re-enter the labour market, than those who are unemployed without having

earned such rights. The estimate for having received health benefits is negative for both men

and women, indicating that health limitations both suppress earnings and lead men and

women to postpone the transition to parenthood. The estimates change little, and most often

not significantly, when the control variables are added consecutively (not shown). Though

selection on unobservables plausibly remains, this modest change indicates that the estimates

are not fully driven by selection. Particularly, it is noteworthy that the introduction of a

control for health selection, one of the more important potential confounders that is rarely

observed, had little effect on the estimates.

The positive estimates for men are similar to the estimated impact of earnings on the second

birth rate obtained by Andersson and Scott (2007) and the positive impact of logged earnings

on the first birth rate found by Lappegård and Rønsen (2012). Taken at face value, this

indicates that there is a substantial income effect on fertility for men. However, as discussed

in section 2.4, the estimates for men may be biased upwards (compared to the causal effect of

income on fertility), due to unobservable characteristics that have a positive effect of earnings

as well as fertility.

For women, the findings are in line with previous studies of women’s earnings in the Nordic

context (Andersson 2000 (Sweden), Vikat 2004 (Finland), Andersson, Kreyenfeld and Mika

2009 (Denmark)), indicating an income effect of women’s earnings on transition to

motherhood. The impact of male earnings is more strongly positive than the impact of female

earnings, indicating that a substitution effect may still be dampening the income effect among

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women. However, the results for women should be interpreted with more caution, as the

expected direction and strength of bias is hard to predict.

5.3 Changes over time

The results presented in the previous section give an average of the impact of earnings on

fertility over the period 1994-2009. This average may hide changes over time that gives

valuable information about the mechanisms linking earnings and the transition to parenthood.

I therefore now turn to the main question of this study – an investigation of any changes over

time in the impact of earnings on transition to parenthood. The extraordinarily large data set

allows for estimation of model 1 separately by year and gender, thus allowing for full

interactions between period and gender and all other independent variables. Results from 16

separate period regressions are shown in figure 3a (men) and 3b (women) (a table with all

year-specific estimates is found in the Appendix). Again, estimates are presented as average

marginal effects to facilitate comparison across models.

[Figure 3 a and b about here]

Starting with the results for men, figure 3a shows that the impact of earnings is strong and

monotonously positive across quintiles throughout the period. The magnitude of the earnings

estimates is relatively stable over time – the impact of earnings weakens somewhat towards

the end of the 1990s, and then increases towards the end of the period. For women, the pattern

of change over time looks different. Throughout the period, the importance of earnings for the

transition to motherhood strengthens markedly. While the impact of earnings is substantially

weaker for women than for men in the beginning of the period, the estimates look strikingly

similar across gender at the end of the period. The indications of anticipatory effects (i.e. the

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fertility of women in the lowest quintile is higher than the fertility of women in the 2nd

quintile) have disappeared entirely toward the end of the period, and the impact of earnings on

the transition to motherhood is now monotonously positive by quintile.

6 Discussion and conclusion

The impact of earnings on the transition to parenthood has become similar for men and

women over time, as earnings have become increasingly important for the transition to

motherhood. The empirical patterns lend little support to explanations based on gender

specialisation, such as Becker’s microeconomic theory of fertility. The hypothesis that the

positive impact of earnings on transition to fatherhood would weaken as alternative costs of

childbearing increase for men receives no support. The patterns observed in the data points

toward two main questions: Firstly, why have earnings become so important for the transition

to motherhood? And secondly, why does earnings remain so important for the transition to

parenthood in a context where overall level of wealth is high, and a large proportion of the

cost of childbearing is covered by the welfare state?

Throughout the article, two main mechanisms that link earnings to the transition to

parenthood have been discussed: Firstly, earnings may affect union entry and stability. If this

mechanism has led to the strengthening earnings-fertility relationship observed, this implies

an overall stronger selection on earnings potential into parenthood. Theoretically, this seems

unlikely: Even if both men and women prefer high-earning spouses, there is no reason to

expect that low-earning men and women should not form unions. Also, such a stronger

selection should lead to a lower proportion of first births, the opposite of what is observed in

21

aggregate statistics: In the period of study, there has been a slight increase in Total Fertility

Rate (TFR)8, and the proportion of first birth to all births has also increased slightly9.

The second mechanism that may have affected the impact of earnings on fertility is the impact

of earnings on couple’s decisions of when and possibly whether to have a first child. Again

looking to aggregate statistics, the transition to parenthood has been postponed with

approximately 2 years for both men and women in the period of study10, implying that men

and women alike postpone the transition to parenthood until earnings are higher. The changes

over period may thus imply that women, like men, increasingly time the transition to

parenthood in order to maximise the income effect of earnings. Two changes in the welfare

“package” may have facilitated this development: Firstly, the increasing availability of high-

quality, low-cost child care has reduced the opportunity cost of childbearing, weakening the

impact of the substitution effect. Secondly, the increasing value of earnings-based parental

allowances relative to other flat-rate transfers to families with children is expected to have

made the income effect of women’s earnings more important. Together, weakening the

importance of the substitution effect and strengthening the importance of the income effect is

expected to strengthen the impact of earnings on the transition to motherhood – making it

similar to the impact of earnings for men.

This development could imply a conflict between two of the main goals of the Norwegian

welfare state – redistribution and gender equality. Possibly, the welfare polices have

contributed to a shift in timing, so that women increasingly gain a foothold in the labour

market before having a first child. However, if these incentives lead some women to never

8 StatBank Norway, https://www.ssb.no/en/statistikkbanken, table 04232 9 StatBank Norway, https://www.ssb.no/en/statistikkbanken, table 05523 10 StatBank Norway, https://www.ssb.no/en/statistikkbanken - table 07278

22

having a first child because they did not gain foothold in the labour market – or because

childbearing had become too late when earnings were sufficiently high – this implies that

welfare schemes have generated a process of cumulative advantage with respect to women’s

family formation. This is linked to the question of whether the strengthened impact of

earnings affects whether – or just when – women have a first child. The answer to this

question will be available only when the women who are currently in their main childbearing

years have completed their fertile period.

Earnings have become more important for men as well as women over the period of study.

Why is income increasingly important for childbearing in a period of high economic growth

and high economic security? Providing a full answer to this puzzle is outside the scope of this

paper, but the increasing value of children could be among the mechanisms driving this

development. Increasing investment in children – both in terms of time and money – depends

on a relatively large household budget. The extent to which the norms and practices around

investment in children has changed over the last decades in Norway, and how this is linked to

the impact of earnings on fertility, remains a question for future research.

23

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26

Figures and tables

Table 1: Summary statistics – person years MEN WOMEN Freq. % Freq. % First birth in current year No 6 629 471 94,9% 4 775 395 93,1% Yes 356 679 5,1% 352 277 6,9% Earnings quintile Missing 840 997 12,0% 813 26 15,9% Q1 1 489 386 21,3% 722 924 14,1% Q2 1 277 964 18,3% 739 392 14,4% Q3 1 154 895 16,5% 733 02 14,3% Q4 1 114 491 16,0% 914 179 17,8% Q5 1 108 417 15,9% 1 204 897 23,5% Educational attainment Higher education, higher degree 338 379 4,8% 230 419 4,5% Higher education, higher lower degre 1 171 349 16,8% 1 386 025 27,0% Primary education 938 826 13,4% 643 547 12,6% Secondary education 4 479 718 64,1% 2 828 805 55,2% Missing 57 878 0,8% 38 876 0,8% Educational enrollment No 4 786 297 68,5% 2 997 256 58,5% Yes 2 199 853 31,5% 2 130 416 41,5% Recieved unemployment benefits No 6 164 078 88,2% 4 770 900 93,0% Yes 822 072 11,8% 356 772 7,0% Recieved health-related benefits No 6 511 950 93,2% 4 780 217 93,2% Yes 474 2 6,8% 347 455 6,8% Period 1994-1997 1 674 299 24,0% 1 232 483 24,0% 1998-2001 1 734 386 24,8% 1 274 224 24,8% 2001-2005 1 796 065 25,7% 1 318 274 25,7% 2006-2009 1 781 400 25,5% 1 302 691 25,4% Mean (S.E.) Mean (S.E.)

Age 29.356 0.003 28.382 0.003

N 6 986 150 5 127 672

27

Table 2: Model 1: Discrete time hazard regression of the probability of first birth.

MEN

WOMEN

Earnings quintile Quintile 2 0.342

-0.096

(0.329, 0.355) (-0.112, -0.0805) Quintile 3 0.623

0.203

(0.610, 0.636) (0.188, 0.217) Qunitile 4 0.79

0.507

(0.777, 0.803) (0.494, 0.521) Quintile 5 0.892

0.57

(0.879, 0.905) (0.557, 0.583) Missing earnings -1.226

-1.398

(-1.260, -1.191) (-1.429, -1.367) Educational attainment Higher education, higher degree 0.145

0.165

(0.132, 0.158) (0.150, 0.180) Higher education, lower degree 0.0641

0.0809

(0.0550, 0.0731) (0.0723, 0.0894) Primary education 0.0452

0.194

(0.0314, 0.0590) (0.178, 0.211) Missing educ. Info 0.0667

0.422

(0.0267, 0.107) (0.385, 0.459) Enrolled in education -0.254

-0.548

(-0.264, -0.244) (-0.558, -0.539) Recieved unemployment benefits 0.0419

0.165

(0.0308, 0.0529) (0.152, 0.177) Recieved health benefits -0.538

-0.417

(-0.560, -0.517) (-0.438, -0.397) Log unemployment rate -0.00131

0.0235

(-0.0171, 0.0145) (0.00477, 0.0423) Period dummies Yes

Yes

Region dummies Yes

Yes N 6 986 150

5127672

Estimates are reported as Average Marginal Effects. 95% confidence intervals in parentheses. Estimates not significant at the 0.001%-level are in italics.

28

Figure 1:

Sources:. Data on cash-for-care from Bakken and Myklebø (2010), and gives the yearly benefit level for a child who do not attend state supported kindergarten at all. Data on child allowances are taken from NOU 1996 p. 133 and 436. (up to 1996) and NOU 2009:232 (after 1996). The time series on “engangsstønad” is compiled from NOU 1996:143 (1993-1996), http://www.regjeringen.no/nb/dep/bld/dok/regpubl/otprp/19961997/otprp-nr-48-1996-97-/3/3.html?id=300894 (1997), http://www.stortinget.no/no/Saker-og-publikasjoner/Publikasjoner/Innstillinger/Stortinget/2000-2001/inns-200001-061/ (1998-2000), http://www.regjeringen.no/nb/dep/bld/dok/regpubl/stprp/20022003/stprp-nr-1-2002-2003-/8.html?id=287805, (2001-2002) http://www.statsbudsjettet.no/Statsbudsjettet-2009/Statsbudsjettet-fra-A-til-A/Fodsel-og-adopsjon--engangsstonad/ (2003-2008). CPI is obtained from https://www.ssb.no/statistikkbanken/, table 08184 Figure 2:

29

Figure 3a:

See Table A1 in Appendix for a table of estimates and details on model specification.

Figure 3b:

See Table A2 in Appendix for a table of estimates and details on model specification

30

Appendix

Table A1: Marginal effects for the impact of earnings on the transition to parenthood – men (as shown in Figure 3a). Standard errors in parentheses. Models are estimated separately by year, and each row thus gives estimates from a separate regression model. The baseline hazard is specified as a linear spline with 5-year knots. All models include controls for educational enrolment and attainment, region of birth, a dummy for reception of health benefits and a dummy for reception of unemployment benefits.

YEAR Q2 Q3 Q4 Q5 MISSING

1994 0.012 (0.001) 0.028 (0.002) 0.038 (0.002) 0.042 (0.002) -0.041 (0.001) 1995 0.013 (0.001) 0.03 (0.002) 0.04 (0.002) 0.045 (0.002) -0.04 (0.001) 1996 0.014 (0.001) 0.032 (0.002) 0.041 (0.002) 0.046 (0.002) -0.04 (0.001) 1997 0.015 (0.001) 0.032 (0.002) 0.041 (0.002) 0.046 (0.002) -0.04 (0.001) 1998 0.015 (0.001) 0.033 (0.002) 0.041 (0.002) 0.047 (0.002) -0.04 (0.001) 1999 0.017 (0.001) 0.033 (0.002) 0.041 (0.002) 0.047 (0.002) -0.039 (0.001) 2000 0.017 (0.002) 0.033 (0.002) 0.042 (0.002) 0.049 (0.002) -0.039 (0.001) 2001 0.017 (0.002) 0.033 (0.002) 0.045 (0.002) 0.053 (0.002) -0.039 (0.001) 2002 0.018 (0.002) 0.034 (0.002) 0.045 (0.002) 0.054 (0.002) -0.039 (0.001) 2003 0.019 (0.002) 0.034 (0.002) 0.046 (0.002) 0.055 (0.002) -0.038 (0.001) 2004 0.019 (0.002) 0.034 (0.002) 0.046 (0.002) 0.056 (0.002) -0.036 (0.001) 2005 0.02 (0.002) 0.037 (0.002) 0.05 (0.002) 0.059 (0.002) -0.036 (0.001) 2006 0.02 (0.002) 0.038 (0.002) 0.051 (0.002) 0.059 (0.002) -0.033 (0.001) 2007 0.022 (0.002) 0.039 (0.002) 0.051 (0.002) 0.059 (0.002) -0.03 (0.002) 2008 0.023 (0.002) 0.043 (0.002) 0.056 (0.002) 0.064 (0.002) -0.029 (0.002) 2009 0.03 (0.002) 0.052 (0.002) 0.069 (0.002) 0.082 (0.003) -0.028 (0.002)

31

Table A2: Marginal effects for the impact of earnings on the transition to parenthood – women (as shown in figure 3b). Standard errors in parentheses. Models are estimated separately by year, and each row thus gives estimates from a separate regression model. The baseline hazard is specified as a linear spline with 5-year knots. All models include controls for educational enrolment and attainment, region of birth, a dummy for reception of health benefits and a dummy for reception of unemployment benefits. Q2 Q3 Q4 Q5 Missing

1994 -0.014 (0.002) -0.001 (0.002) 0.019 (0.002) 0.024 (0.002) -0.067 (0.001) 1995 -0.012 (0.002) -0.001 (0.002) 0.019 (0.002) 0.025 (0.002) -0.065 (0.001) 1996 -0.011 (0.002) 0.003 (0.002) 0.023 (0.002) 0.025 (0.002) -0.064 (0.001) 1997 -0.011 (0.002) 0.003 (0.002) 0.024 (0.002) 0.027 (0.002) -0.062 (0.001) 1998 -0.01 (0.002) 0.005 (0.002) 0.024 (0.002) 0.028 (0.002) -0.061 (0.001) 1999 -0.008 (0.002) 0.01 (0.002) 0.031 (0.002) 0.031 (0.002) -0.059 (0.001) 2000 -0.008 (0.002) 0.012 (0.002) 0.032 (0.002) 0.031 (0.002) -0.057 (0.001) 2001 -0.007 (0.002) 0.012 (0.002) 0.033 (0.002) 0.032 (0.002) -0.054 (0.001) 2002 -0.006 (0.002) 0.013 (0.002) 0.033 (0.002) 0.037 (0.002) -0.051 (0.001) 2003 -0.005 (0.002) 0.016 (0.002) 0.038 (0.002) 0.038 (0.002) -0.047 (0.001) 2004 -0.004 (0.002) 0.016 (0.002) 0.038 (0.002) 0.046 (0.002) -0.046 (0.002) 2005 -0.004 (0.002) 0.017 (0.002) 0.042 (0.002) 0.049 (0.002) -0.046 (0.002) 2006 0.001 (0.002) 0.018 (0.002) 0.044 (0.002) 0.05 (0.002) -0.041 (0.002) 2007 0.004 (0.002) 0.028 (0.002) 0.045 (0.002) 0.051 (0.002) -0.038 (0.002) 2008 0.007 (0.002) 0.038 (0.002) 0.06 (0.002) 0.067 (0.002) -0.033 (0.002) 2009 0.018 (0.003) 0.051 (0.003) 0.084 (0.003) 0.086 (0.003) -0.033 (0.002)