Economic Resources andRemaining Single: TrendsOver TimePearl A. Dykstra and Anne-Rigt Poortman

An influential hypothesis in family research is that having many economic resources

decreases women’s and increases men’s rate of entering a union. A more recent

hypothesis is that the strength of the association between economic resources and

union formation has weakened over time, given decreasing role differentiation by

gender. Rather than looking at the timing of union formation, we look at its

non-occurrence. Using the Netherlands Kinship Panel Study, we find that, as predicted,

high-resource women and low-resource men are more likely to remain single. Contrary

to predictions, university-educated men are also more likely to remain single. The

association between economic resources and permanent singlehood shows little

change over time. Several explanations for this unexpected finding are discussed.

Introduction

Decreasing marriage and rising divorce rates are

testimony to the major changes that family life has

been undergoing since the 1950s. These trends have

been explained in three ways: increased female

economic autonomy, rising consumption aspirations,

and increased individuation. As Lesthaeghe (1998)

has pointed out, the explanations are not mutually

exclusive. Ideational values and economic circum-

stances have a symbiotic relationship. We start from

the perspective that people’s economic standing

shapes their partnership preferences along with bio-

logical, religious, personality, and social factors

(Kalmijn, 1998; Fletcher et al., 1999; Press, 2004).Several authors have argued that women’s growing

economic independence has undermined the speciali-

zation gains arising from a gender-specific division

of labour in the family and has enabled women to

support themselves outside of marriage (Cherlin, 1979;

Becker, [1981] 1991; Ruggles, 1997). A view that is

less often put forward is that men’s deteriorated labour

market position has reduced specialization gains and

made it increasingly difficult to meet the minimally

required economic standards to enter marriage and

stay married (Cherlin, 1979; Oppenheimer et al.,

1997). The underlying assumption that women with

many and men with few economic resources are less

likely to enter marriage—and nowadays, a consensual

union—has become one of the most influential

hypotheses in research on union formation.The gains to marriage theory presuppose a tradi-

tional division of labour with men adopting the role

of breadwinner and women taking on the responsi-

bilities of homemaking and childcare. Gender roles

have changed: men’s contribution to unpaid work

has increased over recent decades just as women’s

participation in the labour force (Blossfeld and

Drobnic, 2001). For that reason, Oppenheimer (1988,

1995, 1997) has introduced a hypothesis about

change over time; having many economic resources

less strongly discourages women’s chances of union

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formation and less strongly encourages men’s chances

of union formation (see also Sweeney, 2002).So far, empirical evidence for the two hypotheses

has been mixed. Findings for men tend to show

that having many economic resources increases their

likelihood of marriage and (to a lesser extent) of

entering a consensual union, but most studies do not

find the expected negative effect of women’s economic

standing (see the review by Oppenheimer, 1997 for

American findings; see Blossfeld and Huinink, 1991;

Blossfeld, 1995; Bracher and Santow, 1998; Liefbroer

and Corijn, 1999; Kalmijn and Luijkx, 2005 for

European and Dutch findings). The second hypothesis

about historical change has been examined less

often. Findings show that the association between

women’s economic resources and marriage has turned

from being zero or negative in older cohorts into being

positive for younger cohorts (Goldstein and Kenney,

2001; Sweeney, 2002). Findings for men are mixed:

studies either report that men’s economic resources

have become less important for marriage (Sassler and

Goldscheider, 2004) or report no change (Sweeney,

2002; Kalmijn and Luijkx, 2005).We follow a novel approach to test both the

hypotheses. Rather than looking at the timing of

union formation, we look at its non-occurrence.

Event-history models, which are typically employed

in research on the entry into marriage, address both

the question of when to marry and the chances of

ever marrying. Determinants of never marrying can

in part reflect effects of marriage timing. For example,

a work-oriented high-earning woman can decide to

fully invest in her career first to discover at an older

age that no eligible partners are left. Nevertheless, and

as Yamaguchi (1998) has demonstrated, the determi-

nants of marriage timing and of ever marrying can

differ. His analyses revealed that birth cohort, age

of last child, and teenage marriage affected only the

probability of remarriage, whereas education affected

only remarriage timing. Recent literature has paid

virtually no attention to the phenomenon of remaining

single throughout life (DePaulo and Morris, 2005).

Using survey data from the Netherlands, we examine

the role of economic resources regarding the likeli-

hood of remaining single. In light of the rise in

cohabitation, remaining single is defined as not

having entered either a marriage or a consensual

union. Consistent with the two hypotheses, we ask

whether there is an influence of men’s and women’s

education and employment history on the likelihood

of remaining single and whether this influence has

changed across cohorts. There are two reasons why

a study of the association between economic resources

and remaining single contributes to existing research.First, such a study may provide a stronger test of

the hypotheses. Assuming that union formation

results from people’s partnership preferences and the

restrictions faced in realizing these preferences, a focus

on remaining single draws attention to variability

in preferences. Implicit in studies of the timing

of marriage or cohabitation is the notion that all

people desire to live with a partner, eventually, but

face varying opportunity structures in realizing their

preferences. A focus on singlehood starts from a

different premise: choosing not to enter marriage or

a consensual union because a partnership is not a

desirable option (Engelen and Kok, 2003). According

to Oppenheimer (1988, 1997), a number of economic

arguments, particularly those about the lower gains

of sharing a household in case of less specialization

(Becker, [1981] 1991), are arguments about the

desirability of marriage and cohabitation and, there-

fore, about their non-occurrence, not of their timing.

The role of restrictions, such as being a desirable

partner to others, also becomes more salient when

focusing on those who remain single; they may

have faced more severe restrictions than those who

succeeded in finding a partner. Conceivably, the

hypotheses will be better supported when tested

on remaining single rather than the timing of union

formation.A second reason to study permanent singlehood

is the lack of recent research on this topic. Clark

and Graham (2005) note that ‘there is little or no

research on predictors of choosing to remain unmar-

ried or uncoupled, or of the circumstances that

lead people to stay involuntarily unmarried’ (p. 133).

Our knowledge about who remains single stems

mostly from historical demography, and the findings

primarily pertain to the period before the second half

of the 20th century. In accordance with economic

theories, one of the most consistent findings is that

those who remained unmarried tended to be high-

resource females and low-resource males. Educated

women, working women, those in high-level jobs,

and those earning higher incomes were more likely

to remain single, whereas the opposite held for men

(e.g. De Jong Gierveld, 1969; Carter and Glick, 1970;

Havens, 1973; Spreitzer and Riley, 1974; Broschard,

1978; Bernard, [1972] 1982; Houseknecht et al., 1987;

Kiernan, 1988; Van Solinge and Van Poppel, 1995).

It remains an open question whether such a pattern

is also observed for cohorts entering the marriage

market in the second half of the 20th century.

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Theoretical Background

Like others, we assume that whether people remainsingle or not depends upon their preferences andopportunities (e.g. Goldscheider and Waite, 1986).If people have less desire to enter a partnership, theyare more likely to remain single. Given equalpreferences, people who are less likely to meetpotential partners or those who are less attractivecandidates are also more likely to remain singlebecause they face great difficulties in their searchfor a partner. Partnership preferences, in turn, areassumed to depend upon the costs and benefits ofentering a consensual union or marriage comparedto the alternative of remaining single. As describedearlier, we focus on economic costs and benefits.To the extent that people’s desire for a partnershipis driven by economic considerations, we assume thatthe greater the economic benefits and the lowerthe costs of a relationship, the more desirable apartnership is. To explain the role of economicresources in the likelihood of remaining singleand the change therein over time, we focus on the(changed) costs and benefits of union formationand the (changed) attractiveness as a potentialpartner. We draw, as much as possible, upon existingtheoretical arguments to explain why people withdifferent levels of economic resources might facedifferent union formation costs and benefits andwhy they might be more or less attractive candidates.

Most theories and hypotheses assume gender-specific roles and imply that high-resource womenand low-resource men face few economic benefitsand high economic costs in union formation. Micro-economic theory is perhaps the most influential(Becker, [1981] 1991). The theory posits that unionformation is beneficial because it allows for specializa-tion; dividing tasks between partners leads to eco-nomic benefits because of greater efficiency. Althoughthe theory is in principle gender neutral, it is usuallyargued that a traditional male breadwinner modelis most efficient because women have a comparativeadvantage at home and men in the labour market.Given that deviations from this traditional arrange-ment are less efficient, it follows that women withmany economic resources have little to gain economic-ally from union formation and the same holds forlow-resource men.

Historical demographers analysing marriagepatterns at the end of the 19th and beginning of the20th century also emphasize the economic costs thathigh-resource women and low-resource men face,given segregated gender roles. For high-resource

women, the incompatability of marriage and a joboutside the home brings a loss of investments ineducation and work and the necessity of having torenounce career ambitions (e.g. Havens, 1973; Freemanand Klaus, 1984; Franssen and Van Heezik, 1989). Forlow-resource men, union formation is costly becauseas presumed breadwinners they have to meet mini-mally required economic standards for setting up anindependent household (e.g. Dixon, 1971; Engelen andKok, 2003).

Besides these low economic benefits and high costs,mating preferences in a society with a gender-baseddivision of labour constitute restricted opportunitiesto enter a union for high-resource women and low-resource men. In such a society, women supposedlyprefer men with good economic prospects becausethe latter are expected to be the breadwinners, whereasmen supposedly prefer women who have not investedtoo much in career resources because they need awife who will assume homemaking responsibilities(Oppenheimer, 1997; Sweeney, 2002). Hence, womenwith many economic resources are unattractive candi-dates, and for men the reverse is true.

Reduced role differentiation by gender suggests thatthe association between economic resources and unionformation has changed (Oppenheimer, 1995, 1997;Sweeney, 2002; Ono, 2003; Sassler and Goldscheider,2004). High-resource women have presumably becomemore attractive candidates (Kalmijn, 1998). Now thatthe dual-earner family has become the norm, peoplemay prefer a higher standard of living or value wife’seconomic resources as a way to reduce the financialrisks of specialization (Oppenheimer, 1995, 1997;Sweeney, 2002). Also, the wife’s (potential) economiccontribution makes the husband’s economic resourcesless crucial for the household’s economic survival,thereby lessening the importance of men’s economicresources for his attractiveness as a marriage candidate(Sweeney, 2002; Press, 2004)—although the rise inthe desired standard of living may have attenuatedthis decrease in the importance attached to men’seconomic resources (Sweeney, 2002). These shifts inpartnership preferences imply that the positive effectof women’s economic standing on chances of remain-ing single has weakened over time, just as the negativeeffect of those of men.

Such a change is also to be expected on the basisof changing economic costs and benefits, and thusdesirability, of union formation. Because marriage andparticularly the more recent alternative of consensualunion—a type of partnership characterized by lessgender role segregation (South and Spitze, 1994; Brinesand Joyner, 1999)—no longer necessarily imply that

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women have to withdraw from the labour market,high-resource women no longer lose their humancapital investments, and may continue to pursue theiroccupational ambitions. As a result, high-resourcewomen might be more willing to enter a union. Forlow-resource men, the economic contributions ofwomen to the household income have made marriageand, particularly, a consensual union increasinglyaffordable.

To arrive at specific hypotheses, it is importantto consider the Dutch context (see also Ono, 2003).The Netherlands has for a relatively long time beencharacterized by a male-breadwinner model, and untilthe late sixties it was common policy to discouragewomen to continue working when they married or hadchildren (Blok, 1989; Pott-Buter, 1993). From the1970s onwards, which is relatively late comparedto other Western countries, women’s labour marketparticipation started to increase (Van der Lippe andVan Doorne-Huiskes, 1995). However, this increasemostly stemmed from an increase in part-time work(De Graaf and Van der Meulen, 1997), and Dutchwomen still have not reached the same level of eco-nomic independence as their male counterparts(Van Berkel and De Graaf, 1998; Siermann et al.,2004). We therefore expect that, in general, high-resource women and low-resource men are morelikely to remain single than low-resource women andhigh-resource men. However, the positive effect ofwomen’s and the negative effect of men’s economicresources on chances to remain single is expectedto be stronger for cohorts born before 1950 thanfor those born after 1950. The latter birth cohortentered the marriage market in the 1970s and 1980s,which was a period of rising female labour forceparticipation rates (Liefbroer and Dykstra, 2000).One might argue that the direction rather than themagnitude of the effect of women’s economicresources has changed in the sense that women withmany economic resources are nowadays less likelyto remain single. We do not expect so, however,because Dutch women’s labour market position is stillnot equal to those of men.

Methodology

Data

The data are from the public release file of themain sample of the Netherlands Kinship Panel Study(NKPS; Dykstra et al., 2005a, b). In 2002 and 2003,face-to-face computer-assisted interviews were con-ducted with over 8,161 men and women between

the ages of 18 and 80. At the end of the interview,respondents received a self-completion questionnairethat was picked up by the interviewer at a later date.The self-completion questionnaire mostly pertained tosubjective issues (e.g. attitudes, well-being). In 92 percent of the cases the self-completion questionnaire wasreturned and processed.

A random sample of addresses of private residencesin the Netherlands was used. The addresses werefrom all over the Netherlands, and not restricted tospecific regions or municipalities. The overall responserate of the NKPS study was 45 per cent, which iscomparable to the response rates generally achieved inDutch surveys on topics like families and well-being.A comparison of the sample with population figuresshows that the main sample is reasonably representa-tive with respect to urbanization and region but hasan overrepresentation of women, people in the middleage ranges, and people with children living at home(Dykstra et al., 2005a). The NKPS team constructedweights to make the sample representative of theDutch population with respect to household type, sex,age, region, and urbanization. Given that unweightedand weighted multivariate analyses yielded similarresults, we present the results of unweighted analyses,but for descriptive results (see Table 1) we use sampleweights.

The analyses reported in this study are based on2,142 men and 2,785 women who were between theages of 40 and 79 at the time of the interview (i.e. birthcohort 1923–1963), had valid data for their partnershiphistory, and who had a heterosexual orientation. Theage of 40 was selected as the minimum age to ensurea focus on people who were likely to remain single forthe rest of their lives. Preliminary analyses showed thatthe likelihood of entering a union for the first timeafter that age is quite low; of first marriages, only0.6 per cent were contracted after the age of 40, andof the never married who entered a consensual union,1.0 per cent did so after age 40.

Measures

Remaining single

Questions were asked about the current partnershipas well as all previous partnerships. We distinguishedrespondents who had ever shared a household witha partner for at least 3 months, either married orunmarried, and excluded individuals with a homo-sexual orientation (N¼ 129). Remaining single isdefined as never partnered by age 40, meaning thatthe respondent had never married or never lived ina consensual union by that age. As the descriptive

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statistics in Table 1 show, about 7 per cent of the menand 6 per cent of the women were never in a maritalor consensual union by the age of 40. The numberof single men is 205, of whom 89 were born before1950 and 116 in or after 1950 (unweighted data). Thenumber of single women is 211 (105 in the early and106 in the late cohort). These numbers are highenough to obtain reliable estimates but urge us to use

models that are as parsimonious as possible. Whenlooking only at those who never married by the ageof 40, the percentages are higher: 11 per cent of themen and 10 per cent of the women never married,which underscores the importance of also consideringconsensual unions. In preliminary analyses we checkedwhether results differ when looking at the likelihoodof remaining unmarried (instead of remaining never

partnered) at age 40. These analyses led to similar

conclusions, and because of the increased importance

of consensual unions we only present the results for

remaining never partnered.

Economic resources

We used information about respondents’ education

and employment history to measure their economic

resources. Although income and occupational status

are often used as indicators of economic resources,

the NKPS lacks retrospective measures for income

and occupational status. The indicators referring

to respondents’ employment history cover the period

from time since leaving school to the time of

the survey. Such measures are preferable to those

referring to respondents’ employment at the start

of their career because at that time their labour

market position might not yet be stable (Sassler and

Goldscheider, 2004). Our measures covering people’s

employment history up to the time of the survey

have the disadvantage that any association with

remaining single may be due to reversed causation,

particularly for women. Women’s careers, especially in

older cohorts, were intertwined with their partnership

career as they often (were legally required to) quit

work when they married or had children (Blok, 1989;

Pott-Buter, 1993). For men the problem of reversed

causation is far less severe, because entering a union

and becoming a parent have less of an impact on their

employment status and career.

Education

Information about the educational level of the

respondent was delineated via the question: ‘What

is the highest level of education that you pursued?’

Respondents were presented with 10 answer categories

ranging from (1) did not finish primary education

to (10) post academic education (e.g. Ph.D. degree).

For the purpose of the present analysis, we first

constructed four dummy variables by distinguish-

ing five categories: (i) up to primary education

(reference category), (ii) lower secondary education,

(iii) upper secondary education, (iv) higher vocational

education, and (v) university education. For reasons

of parsimony and because the highest and lowest

levels of education contain relatively few men and

women (see Table 1), particularly when the male

and female samples are broken down into cohorts

born before and after 1950, we also recoded the

original 10 categories into a single variable for

formally required years of education, ranging from 4

to 20 years.

Table 1 Descriptive statistics of the dependent andindependent variables

Men(per cent)

Women(per cent)

Single at age 40 7.2 5.7Control variables

Birth cohortBirth year (�1900)a 46.3 46.9Born 1923–1949 56.5 53.2Born 1950–1963 43.5 46.8Divorced parents 4.6 5.1One or morenever-married siblings

11.1 11.4

Economic resourcesEducation

Up to primary 10.5 12.6Lower secondary 25.4 38.8Upper secondary 25.7 24.6Higher vocational 26.2 19.5University 12.2 4.5Years of educationa 12.2 11.2

Employment historyNever employedb – 6.8First job at age 30

or over4.1 7.7

Over two unemploymentspells

9.4 5.8

Exit at age 40 or earlier 2.8 21.6Scoring on any

employment historyvariable

14.9 40.6

Unweighted N 2,142 2,785

Note: Data are weighted.

Source: NKPS (own calculations).aMean presented here.bNever employed men are included in the ‘Over two unemployment

spells’ category.

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Employment history

The survey had only limited information on respon-

dents’ occupational histories. Information was col-

lected on employment status (never gainfully

employed, currently gainfully employed, or previously

gainfully employed). In addition, information was

collected on the number and duration of periods of

unemployment, but not on their timing. The survey

also had questions on the age at entry into the labour

market, and for those who were not working at the

time of the interview there was information on

the age at which the labour force had been left. On

the basis of this information, we constructed four

dichotomous variables indicating poor economic

standing: (i) never had a paid job, (ii) late entry,

that is, the first job at age 30 or above, (iii) exit before

age 40, and (iv) discontinuous employment, that is,

unemployed twice or more often for at least a month.

Given that only few men never had a paid job

(N¼ 11), this variable only pertains to women, and

the never employed men are included in the ‘More

than two unemployment spells’ category. For the

same reason (see Table 1 for the relatively small

number of cases scoring a one on the employment

history dummies) as for constructing a single variable

for education, we also constructed a composite dummy

variable indicating whether respondents scored one

on any of the four separate dummy variables.Note that information is lacking on whether the

unemployment periods occurred specifically in early

adulthood (when respondents were of ‘marrying age’).

Also note that interruptions other than unemploy-

ment, such as leaving the labour market due to

occupational disability or full-time homemaking, could

not be identified. Finally, note that it cannot be

established whether those who left the labour market

before the age of 40 left the labour market perma-

nently. Respondents who had left the labour market

at the time of the interview might still re-enter at a

later point in time.

Control variables

The analyses take into account the birth year of

the respondent. We checked for non-linear effects of

birth year and used a non-linear specification where

they were found. To assess changes in the role of

economic resources over time, a less detailed distinc-

tion was made by distinguishing between respondents

who were born in the 1923–1949 period and those

born in the 1950–1963 period. Furthermore, we con-

trol for family background characteristics. As perma-

nent singlehood is rare, there is a stigma attached to it,

and stereotypes suggest that those who remain singlethroughout life are socially isolated and lack commu-nication skills (Byrne and Carr, 2005). However, thepenalty of remaining single varies across social contexts(Dixon, 1971; Koropeckyj-Cox, 2005). The analysestherefore control for family background variables thatindicate the degree to which permanent singlehoodis likely to meet social disapproval. In general, it isassumed that people in social contexts where single-hood meets less disapproval have a higher likelihoodof remaining single than people who face a higherpenalty. In preliminary analyses, we used several familybackground variables to measure the extent to whichremaining single would meet social disapproval(i.e. having non-religious parents, divorced parents,never-married parents, one or more never-marriedsiblings, and being resident of a big city at age 15). Forreasons of parsimony we chose to include only thosevariables that had significant effects on the likelihoodof remaining single: having divorced parents andnever-married siblings.

Divorced parents

In families where marriage for life is the norm, thepenalty of remaining single is likely to be greater thanin families with a history of diversity and complexityin marriage and parenting relationships. To accountfor a complex family history, we constructed a variableindicating whether respondent’s parents had divorcedbefore he or she reached the age of 21 (1¼ yes).

One or more never-married siblings

We also used the partnership history of siblings asa measure of whether respondents were likely to meetdisapproval when remaining single. We constructeda dummy variable indicating whether respondentshad at least one never-married sibling. Note that thenumber of respondents with one or more never-married siblings has probably been underestimatedbecause the NKPS survey does not have maritalhistory information on all siblings. For a maximumof two randomly selected siblings there is informationon current partner status. One of the responsecategories was ‘single, never married’. Given that notevery one has siblings, we control for being the onlychild or not.

Analytical Strategy

Logistic regression analyses were used to assess therole of education and employment history on thelikelihood of remaining single. The logistic regressions

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were carried out separately for men and women. We

estimated two models to assess the overall effect of

economic resources on remaining single. The first

model (Model A) incorporated the dummy variables

for education and employment history. The second

model (Model B) included the more parsimonious

measures for education (i.e. years of education) and

employment history (i.e. the composite variable

indicating whether respondents’ scored positive on

any of the dummy variables). To compare the fit of

the models, we calculated the Bayesian information

criterion (BIC); the more negative the BIC the better

the model fit (Raftery, 1996). The best fitting model

was used in the analyses to assess changes over time.The changing impact of education and employ-

ment across cohorts born before and in or after 1950

was examined by estimating an interaction model.

Interaction terms between birth cohort on the one

hand and the variables for education and employment

history on the other hand were included in this

interaction model. From this model we derived the

implied effects of education and employment history

for each cohort. Besides presenting the effects for

each cohort, we also show whether the effects differ

significantly between the two cohorts (as indicated

by the P-value of the interaction terms). Because

any difference in the effects between cohorts might

reflect differences in unobserved heterogeneity rather

than real differences (Allison, 1999), we replicated

our analyses using heteroscedastic probability models

that take unobserved heterogeneity into account. These

additional analyses show insignificant likelihood ratio

tests for unobserved heterogeneity and yield estimates

similar to those obtained from logistic regression

analyses. We, therefore, present the results of the

logistic regression analyses.

Results

First, we assess the overall impact of economic

resources on the likelihood of remaining single (see

Table 2). As the results for the control variables show,

the family background variables hardly affected the

likelihood of remaining single for both men and

women. Whether respondents had never-married

siblings was the only significant family background

determinant of singlehood (in both Models A and B).

As the parameters for birth cohort in Table 2 show,

the trend over time regarding the likelihood of

remaining single is linear for men. Men from younger

birth cohorts were more likely to remain single than

those from older birth cohorts. Among women, the

likelihood of remaining single decreased across thecohorts born before the late-1950s [turning point is1961 in Model A (¼ 0.244/(2�0.002)) and 1954 inModel B]. This trend reversed in the cohorts bornin the 1960s and after.

Turning to the variables of main interest, the resultsfor Model A show that men’s educational level hadthe expected negative effect on the likelihood ofremaining single. The higher a man’s educationallevel the less likely he was to remain single. Forexample, compared to men with only primary educa-tion, the odds that men with lower secondary orhigher vocational training remained single was about55 per cent lower, and even 60 per cent lower formen with upper secondary education [100 per cent�

(1� exp–0.927)]. University-educated men were anexception. Men with a university education wereas likely to remain single as were those with onlyprimary education.

The hypothesis that low-resource men are morelikely to remain single finds stronger support whenlooking at the effects of the employment historyvariables in Model A, as all indicators for poor eco-nomic standing had a positive effect on the likelihoodof remaining single. Particularly men who experienceddiscontinuities in their employment career or leftthe labour force early, were significantly more likelyto remain single than were men with stable careers,with the odds of permanent singlehood being morethan twice as high. A late entry also increased men’slikelihood of remaining single, but this effect was onlymarginally significant. In Model B, the dummies foreducation were replaced with a continuous variableindicating years spent in school. Because the dummiesshowed a non-linear relationship, a quadratic termwas included as well. In Model B the separate dummiesfor poor economic standing were replaced with asingle composite variable. The results for this modelalso show that men’s likelihood of remaining singlefirst decreased with additional years of schooling butstarted to increase again when men had more than13 years of schooling [¼ 0.349/(2�0.013)]. Moreover,men with poor economic standing were significantlymore likely to remain single. When comparing theBICs, Model B fitted the data better than Model A,and in the analyses of change over time we thereforeused the more parsimonious Model B for men.

Among women, the results for Model A showthat the likelihood of remaining single increased asthe level of educational attainment increased, whichsupports the hypothesis that high-resource womenare more likely to remain single than women withfew economic resources. Differences in the likelihood

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of remaining single turned out to be non-significant

between women with only primary and those with

lower secondary education. If we compare women at

the top of the educational ladder with those at the

bottom, highly educated women have over five times

higher odds of remaining single. The findings for

employment history are only partially consistent with

our hypotheses. As expected, women who were never

gainfully employed or left the labour force early

were less likely to remain single than women who

had a stable career, although the effect of never having

had a job was only marginally significant. Note that

this negative association may also arise because women

left the labour force when they married or became

mothers (reversed causation).Contrary to expectations, having experienced spells

of unemployment increased rather than decreased

women’s likelihood of remaining single. However,

because single women were far more likely to stay

attached to the labour market than women who

entered a union, the former automatically ran a

higher risk of experiencing unemployment at some

point in time than did the latter. The results for Model

B confirm that higher educated women were more

likely to remain single than lower educated women.

Although the effects of the separate dummies were

inconsistent, the composite measure for an unstable

career shows a marginally significant negative effect

on the likelihood of remaining single. As described

earlier, the effects of women’s employment history

should be interpreted with caution. Inspection of the

BICs shows that Model B fitted the data substantially

better than Model A. For that reason, Model B was

used in the analyses assessing change over time.Table 3 shows the results of the interaction models

testing whether the effects of economic resources

have changed across cohorts. As stated, the best fitting

model specification (Model B) was used for these

Table 2 Logistic regression of men’s (N¼ 2,142) and women’s (N¼ 2,785) likelihood of remaining single oneconomic resources and control variables: B-coefficients

Men WomenModel A Model B Model A Model B

Control variablesBirth year 0.028��� 0.029��� �0.244��� �0.216���

Birth year squared � � 0.002 0.002���

Divorced parents �0.379 �0.381 �0.856� �0.817�

One or more never-married siblings 0.542��� 0.537��� 0.628��� 0.624���

Economic resources

EducationLower secondarya

�0.764��� 0.406Upper secondary �0.927��� 1.238���

Higher vocational �0.816��� 1.819���

University �0.383 1.721���

Years of education �0.349��� 0.214���

Years education squared 0.013��� –

Employment historyNever employedb

� �0.832�

First job at age 30 or over 0.523� 0.205More than two unemployment Spells 0.695��� 0.597��

Exit at age 40 or earlier 0.792�� �0.810���

Scoring on any employment history variable 0.779��� �0.276�

Model Chi square 67��� 58��� 135��� 107���

Log likelihood �642 �647 �680 �694

BIC �15,059 �15,081 �20,628 �20,647

Source: NKPS (own calculations).aReference category is ‘Up to primary’.bNot estimated for men, never employed men are included in the ‘More than two unemployment spells’ category.�P50.10; ��P50.05; ���P50.01 (two-tailed test).

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analyses. Another reason for preferring Model B toModel A for analyses of historical change is that thelatter has too small numbers scoring one on thedummy variables to yield reliable estimates whenanalysing cases per birth cohort.

With respect to the control variables, the resultsmirror those found in Table 2. Having never-marriedsiblings strongly increased men’s and women’s like-lihood of remaining single. The effects of the controlvariables did not change across cohorts and are,therefore, constrained to be equal for both cohorts inthe interaction models. As to changes in the role ofeconomic resources, the role of education for men’slikelihood to remain single did not appear to havechanged much, which is contrary to our hypotheses. Ifanything, the effect of education became slightlystronger over time, but the interaction terms showthat the change in the effect of education acrosscohorts was not significant. A similar pattern isobserved for the effects of men’s employment history.Men with unsuccessful labour market careers weremore likely to remain single than men with successfulcareers in both cohorts, but the difference was morepronounced in the younger birth cohort. The changeacross cohorts was, however, not significant.

The results for women do not support ourhypotheses about change over time either. Contraryto expectations, the effect of education was stronger for

the younger than older birth cohort, with higher

educated women having a higher likelihood of remain-

ing single than lower educated women, but we note

that the change across cohorts is not significant. The

change in the effect of having few labour market

resources was significant and in the expected direction.

Women with few investments in the labour market

were less likely to remain single in the oldest cohort,

but their likelihood of remaining single did not differ

from that of women with a more stable career in

the youngest cohort. Although it would be tempting

to conclude that this result supports our hypotheses,

the weakening influence of women’s labour market

investments should probably be attributed to a reduced

importance of reversed causation. The explanation of

the negative effect of poor economic standing in the

older cohort presumably is that these women (had to)

quit working when marrying. For younger generations

this explanation is less plausible because women

nowadays are more likely to continue working when

they marry or have children.

Discussion and Conclusions

The hypothesis that possessing many economic

resources increases men’s and decreases women’s

likelihood of union formation was largely confirmed

Table 3 Logistic regression of men’s and women’s likelihood of remaining single on economic resources,control variables, and interaction terms between cohort and economic resources: B-coefficients

Men WomenCohort51950

Cohort�1950

Sign.historicalchangea

Cohort51950

Cohort�1950

Sign.historicalchangea

Divorced parents �0.360 b�0.802� b

One or more never-married siblings 0.558��� b 0.601��� b

Economic resourcesYears of education �0.285� �0.386�� ns 0.156��� 0.241��� nsYears of education squared 0.011 0.015�� ns – –Scoring on any employment

history variable0.575�� 0.949��� ns �0.675��� 0.267���

Model Chi square 54��� 96���

Log likelihood �649 �699BIC �15,054 �20,629N 1,167 975 1,520 1,265

Note: Coefficients are derived from an interaction model including interaction terms between the indicators for economic resources and a dummy

for birth cohort.

Source: NKPS (own calculations).aIndicates whether interaction terms between birth cohort and the indicators for economic resources are significant.bEffects of control variables constrained to be equal across cohorts (i.e. no interaction terms included).�P50.10; ��P50.05; ���P50.01, ns¼ not significant (two-tailed test).

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in our study. For men, we found that a low educationand a poor employment history were associated witha higher likelihood of remaining single. Men at thetop of the educational ladder were an exception. Theywere as likely to remain single as men at the bottomof the ladder. The opposite pattern emerged forwomen. Highly educated women and those withmany labour market resources were more likely toremain single. We note, however, that the findingswith respect to women’s employment history shouldbe interpreted with care because the association mayalso be due to reversed causation. Many women,particularly from older generations, quit working incase of family formation and, similarly, those whodid not enter a union had to continue working ormay even become more strongly focused on theircareer.

Our study replicates the few earlier findings onpermanent singlehood among less recent cohorts, butalso introduces new insights. Bernard’s ([1972] 1982)observation that spinsters are ‘the cream of the crop’holds for this study, but we also find that there aretwo rather than just one category of bachelors. Inaddition to Bernard’s ‘the bottom of the barrel’ group,there is a university-educated group of permanentlysingle men. The pattern that men at the top and atthe bottom of the educational ladder are most likelyto remain single has been reported in older US studies(e.g. Spreitzer and Riley, 1974). The arrival of a well-educated single male elite in the Netherlands was noteda decade ago by Van Solinge and Van Poppel (1995).The new singles in their study were urban dwellersand extensive consumers of culture.

The relatively high likelihood of remaining singlefor university-educated men is contrary to what weexpected. However, our reasoning was based on therole of economic resources, and education stands notonly for such resources but also for social-culturalcharacteristics, such as social skills, openness to newexperiences, and progressive attitudes (Lopata, 1973;Myers and Booth, 2002; Poortman and Van Tilburg,2005; Mirowsky and Ross, 2007). It might thus well bethat the highest educated group represents a culturalelite that is more open to and capable of a non-conformist lifestyle, such as singlehood, or has higherstandards for a long-term partner. An alternativeinterpretation comes from a gynocentric perspectivewhich focuses on women’s views for men (Press,2004). Well-educated women hoping to maintain ajob and to have children will search for a mate whowill share family work. Men who are strongly careeroriented and unwilling to contribute to householdlabour are unattractive to women who espouse an

egalitarian division of tasks. It is conceivable thatamong the university-educated men there is a sizablegroup of career males who are unwilling or reluctantto devote time to child care and housework. Thesemen are unattractive to potential partners, i.e. highlyeducated women who are strongly motivated to pursuea career of their own.

Unexpectedly, our findings with respect to changein the role of economic resources over time did notsupport the hypothesis that having many economicresources less strongly discourages women’s chancesof union formation and less strongly encouragesmen’s chances of union formation in younger thanolder cohorts. Highly educated women who were bornafter 1950 and entered the marriage market whenwomen’s labour market participation was on therise continued to be as likely to remain single ashighly educated women who were born before 1950.Similarly, the relation between men’s economicresources and the likelihood of remaining single didnot change much across cohorts. The question is whyour study failed to find support for the hypothesisabout historical change.

A first reason may be that the findings are speci-fically Dutch. Earlier work in the Netherlands on therole of economic resources on men’s marriage timingalso failed to find evidence for change over time(Kalmijn and Luijkx, 2005). Theoretically, historicalchange was to be expected because men’s and women’sroles have become more alike over the past decades.Although this trend towards less role differentiationby gender is also visible in the Netherlands, the risein women’s labour force participation started relativelylate and pertained mostly to part-time work (e.g. Vander Lippe and Van Doorne-Huiskes, 1995; De Graafand Van der Meulen, 1997; Cuijpers et al., 2004).Hence, it may be too early to observe change in therole of economic resources. Such change may onlybe observed when even younger cohorts are included.The younger cohort in this study entered the marriagemarket in the 1970s and 1980s and that time periodmarked the start of changing gender roles in theDutch context. Men’s and women’s economic stand-ing continues to be inegalitarian in the Netherlands(e.g. Cloın and Boelens, 2004), as witnessed by themost prevalent arrangement of men working full timeand women working part time. Hence, even amongyounger Dutch cohorts change might be less pro-nounced than in other countries.

A second reason for no historical change may bethat there are counteracting developments towardsan increase rather than decrease in the effects of eco-nomic resources on the likelihood of remaining single.

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This possibility is particularly relevant for interpretingthe findings for men. First, the rise in dual-earnerfamilies might imply the desire for a higher standardof living, which is not implausible given contempo-rary consumption patterns. If so, it is likely that men’seconomic resources continue to be equally or evenmore important than before (Sweeney, 2002). A secondcounteracting development relates to men’s remarriagepatterns. The past decades have shown an increasein the number of people separating or divorcingfrom their partner. The rate of remarriage (andre-cohabitation) is higher among men than women,and an increasing number of previously partneredmen finds a partner from the first marriage marketrather than the remarriage market (Ekamper et al.,2003; Rønsen and Skrede, 2006). This means thatcontemporary men who have not yet entered a unionface increased competition from men who are lookingfor a partner a second time and the group of menwho remain single may have become even moreselective in terms of poor educational and labourmarket qualifications.

A third explanation is that the timing and non-occurrence of union formation may be a differentphenomena, governed by different mechanisms (e.g.Dixon, 1978; Yamaguchi, 1998; Engelen and Kok,2003). To the extent that the delay of union formationleads to putting it off, findings for the timing of unionformation should largely resemble the findings forpermanent singlehood. As described earlier, permanentsinglehood may be confined to a selective group ofpeople who consist of (i) the least attractive candidateson the marriage market and (ii) those who do notwant to enter a partnership. As most people entera union at some point in time, the timing of unionformation might be more responsive to changes ingender roles. Permanent singlehood remains rare,however, and the group of people who stay singlethroughout their lives may continue to be the sameas before, despite changes in men’s and women’sroles. Health selection, though not addressed in thepresent study, requires greater attention. People inpoor health not only have lower probabilities ofunion formation but also tend to have less successfulemployment careers (Fu and Goldman, 1996; Waiteand Gallagher, 2000).

Our final explanation is methodological. We cannotrule out the possibility that the small numbers ofsingles in the cohorts under comparison are respon-sible for the absence of evidence showing a change inthe role of economic resources for union formation.This is a problem that is typical of research on smallgroups in a population.

The findings of this study, particularly when

compared to previous work on the timing of union

formation, call for a replication in other countries.

Such a replication should directly compare the effects

of economic resources on singlehood and on the

timing of union formation. That way one can find out

whether our results are typical for the Dutch context

or exemplify a more general pattern. A simultaneous

within-country examination of the timing and non-

occurrence of union formation could help to resolve

the issue of whether permanent singlehood is sub-

stantially different from the timing of union formation.

Acknowledgements

This study is based on data from the Netherlands

Kinship Panel Study (NKPS), which is funded by

the ‘Major Investments Fund’ of the Netherlands

Organization for Scientific Research (NWO; 480-10-

009). Financial and institutional support for the

NKPS also comes from the Netherlands Interdisciplin-

ary Demographic Institute, Utrecht University, the

University of Amsterdam, and Tilburg University. An

earlier version of this paper was presented at the 2004

meeting of the Gerontological Society of America, the

2006 meeting of the Dutch and Flemish Sociological

Associations in Tilburg, the NKPS seminar series,

Utrecht, and a 2007 seminar at the Vienna Demo-

graphic Institute.

Funding

The Netherlands Organization for Scientific Research

grant 451-03-064 (to A.-R.P).

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Authors’ Addresses

Pearl A. Dykstra (to whom correspondence should

be addressed), Netherlands Interdisciplinary

Demographic Institute, PO Box 11650, 2502 AR

The Hague, The Netherlands.

Email: dykstra@nidi.nlAnne-Rigt Poortman, Department of Sociology,

Utrecht University, PO Box 80140, 3508 TC

Utrecht, The Netherlands.

Email: a.poortman@uu.nl

Manuscript received: January 2008

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