ECONOMIC RESEARCH CENTERDISCUSSION PAPER
E-Series

No.E14-4

Is a Daughter Really Like Water Spilled on the Ground?
Adult Children’s Gender, Filial Support, and Parental
Mortality in Rural China
byEiji Mangyo

February 2014

ECONOMIC RESEARCH CENTERGRADUATE SCHOOL OF ECONOMICSNAGOYA UNIVERSITY

Is a Daughter Really Like Water Spilled on the Ground?
Adult Children’s Gender, Filial Support, and Parental Mortality in Rural China
Eiji Mangyo *AbstractLong-standing patrilocal traditions hold that sons in rural China reside with their parents
after marrying, whereas daughters join their husbands’ families. Knowing they will relyon their adult children for support when they are old, rural Chinese parents might preferto bear sons rather than daughters. Using longitudinal data [2002, 2005, and 2008/2009]of the elderly from 22 rural Chinese provinces, this study examines whether evidence
endorses such a preference. Evidence shows no correlation between the gender of adultchildren and mortality of their elderly parents. Although evidence reveals that sons are
somewhat more likely than daughters to support their elderly birth parents financially,physically, and emotionally, it also reveals that many daughters do so, countering theperception that married daughters become part of husbands’ families.

JEL classification: J14, J16
Keywords: parental mortality, child gender, filial support*

Graduate School of Economics, Nagoya University, Furo-cho, Chikusa-ku, Naogoya,464-8601 Japan Email: eijim@soec.nagoya-u.ac.jp
I am grateful to participants of the Hitotsubashi-Nagoya Joint Work Shop of Income
Inequality for useful comments on earlier drafts. Data used for this research wasprovided by the study entitled “Chinese Longitudinal Healthy Longevity Survey”(CLHLS) managed by the Center for Healthy Aging and Development Studies, Peking
University. CLHLS is supported by funds from the U.S. National Institutes on Aging(NIA), China Natural Science Foundation, China Social Science Foundation, andUNFPA.
1

1.

Introduction

Traditionally and today in rural China, where pensions are rare and non-familial
assistance is limited, the elderly rely on their children for daily financial, physical, andemotional support. In that regard, the Chinese have a saying that captures the patrilocal
tradition of daughters’ abandoning their birth families and joining their husbands’families upon marriage: “A daughter is like water spilled on the ground.” Given thiscombination of tradition and elderly necessity, Chinese parents might have an incentive
to bear and raise boys rather than girls.
This paper examines if there is any factual basis for such a preference by
examining how the gender of adult children relates to the welfare of the elderly in rural
China. It first examines whether the existence of and the number of sons are related toparental longevity and finds no systematic difference in mortality between elderly
parents of sons and daughters. Next, it examines whether the gender of adult offspringcorrelates with systematic differences in filial support of elderly parents. Evidence
indicates that sons may be more likely than daughters to support their birth parents
financially, physically, and emotionally. However, the evidence also shows that many
married daughters do not forsake their birth parents and provide support after marriage.
Evidence further shows that differences between sons’ and daughters’ support are in2

some cases statistically and economically indistinguishable.
This paper is organized as follows. The section Background describes the malecenteredness of traditional Chinese culture as the basis for the hypothesis studied in thispaper: elderly parents with sons enjoy greater longevity and welfare than those withdaughters. This section also reviews previous literature in this field. The section Data
and Research Methodology discusses the indicated topics. The section Results presents
research results. The final section concludes.
2.

Background

China traditionally has been a patrilocal, patrilineal, and patriarchal culture.
Following a field survey in a rural Central China, Junhong (2001) noted that the birth ofa son was celebrated more than that of a daughter and elevated the mother’s position
among family and neighbors. The family line is preserved through the male linage, andsocial and cultural customs prompt preferential household treatment of sons overdaughters. Sons are raised as successors, future breadwinners, and future heads offamilies. After marrying, a son (typically the eldest) is expected to reside with his wife
in his parent’s household. In contrast, daughters are reared as temporary residents of
their parents’ household because they are expected to join their husbands’ families after
marriage. A woman is expected to obey her father in youth, her husband after marriage,and her son in old age (Arnold & Zhaoxiang, 1986).
3

Alongside socio-cultural and economic incentives for couples to prefer sons,introduction of China’s one-child policy in 1979 triggered a rise in the ratio of male tofemale births (Ebenstein, 2010). According to the World Bank (2006), the ratio of malebirths per 100 female births was stable around 106 until 1979 and rose to 107 in 1982,
to 115 in 1990, and to nearly 120 in 2000. Micro-level studies provide further evidenceof preference for sons. Using data from fertility surveys in China, Johansson andNygren (1991) showed that the ratio of male births rose with parity during the 1980s.
Using Chinese census data, Ebenstein (2010) found that intervals between births werelonger for boys than for girls. Using the 1990 Chinese census, Yi et al. (1993)
documented that the ratio of male births was higher among couples who had only
daughters or who had more daughters than sons. Findings from Johansson and Nygren(1991), Ebenstein (2010), and Yi et al. (1993) endorse the conjecture that Chineseparents have been seeking gender-selective abortions. This conjecture is further
supported by Junhong’s (2001) survey of second pregnancies in rural central China: hefound that none of 157 male fetuses was aborted, whereas 46% of 123 female fetuses
was aborted after an ultrasound B-scan.
The cited studies indicate that rural couples have sought to control the genderof their offspring as well as their fertility since the introduction of China’s one-child4policy in 1979 and especially following wider rural access to ultrasound diagnosis in the
1990s. Even before ultrasound was widely available, Chinese couples exercised
imperfect control over the gender composition of their families by reproducing, until
they delivered a boy (Junhong, 2001). In short, evidence has accumulated that rural
Chinese couples have sought to bear sons rather than daughters. As evidence suggests
they discriminate against daughters in allocation of household resources (Brown & Park,
2002; Jianghong Li, 2004; Song & Appleton, 2006).
These findings present an issue for statistical research. Unless properly
controlled for, couples’ economic condition produces a downward bias in the potential
protective effect of sons on mortality of elderly parents. More generally, parents’preference for sons—which is unobservable by researchers—may engender statisticalbias by correlating both with the gender of adult offspring and the welfare of elderly
parents. However, Lee (2008) and Ebenstein (2010) found that the gender ratio among
first borns approximated the natural male–female ratio (1.05) in Korea and China. Lee(2008) used the gender of first-born children as the instrument for sibling size to
examine its effect on investment in child education in Korea. Li and Wu (2011) used thegender of the first-born child as the instrument for the existence of sons in examining
the effect of having a son on women’s bargaining power inside households and women’s
5nutrition and health in rural China.
Literature concerning the relation between parents’ mortality and the gender of
their children is sparse. Drawing data from rural Bangladesh, Rahman (1999) showedthat after controlling the number of kin residing within the same and adjacenthouseholds, living with two or more sons was negatively correlated with the mortalityof elderly fathers and mothers, whereas the number of co-resident daughters had nocorrelation. Analyzing the same data, Rahman (2000) further stated that the beneficial
effect of co-resident sons on parental mortality was stronger for elderly mothers than forfathers and did not differ with household wealth. Hurt, Ronsmans, and Quigley (2006)
found that mortality for both mothers and fathers negatively correlated with the numberof surviving sons using data from a poor rural area in Matlab, Bangladesh. Using datafrom Taiwan and China, Pham-Kanter and Goldman (2012) found no correlations
between parental mortality and several measures of gender of adult children. 1
Literature concerning the relation between child gender and filial support isalso limited, especially for mainland China. Lee, Parish, and Willis (1994) and Lin et al.
(2003) found that in Taiwan, most adult children, both sons and daughters, financially

1

For China, Pham-Kanter and Goldman (2012) used the same data source as this study(the Chinese Longitudinal Healthy Longevity Survey). However, this study uses thefollow-up window to the 2008 survey, whereas theirs used the follow-up window to the2005 survey.
6

supported their parents, with sons being the main contributors. Using data from a
Chinese township, Yang (1996) stated that adult children’s financial support to husbands’
parents was made in appreciation of parents’ previously raising and caring for them,whereas financial support to wives’ parents was in appreciation for their current
contribution to housework and child rearing. To all appearances, previous studies ofrelations between children’s gender and filial support used adult children as the unit ofanalysis. No previous studies analyzed the relation between the availability or amountof filial support and child gender using elderly parents as the unit of analysis.
3.

Data and Research Methodology

3.1 Data
The Chinese Longitudinal Healthy Longevity Survey (CLHLS) was started in1998 to study the determinants of healthy longevity of human beings. It initiallysampled 8,959 persons aged 80 or older throughout 22 Chinese provinces. 2 In 2000,
CLHLS resurveyed its original sample and recorded health, family, and living
conditions of deceased and surviving individuals. CLHLS interviewed 4,824 new elders,bringing the combined sample size for the 2000 survey to 11,161 (surviving)
respondents aged 80 or older. Besides following up on elderly respondents from the2

The 22 provinces are Anhui, Beijing, Chongqing, Fujian, Guangdong, Guangxi, Hebei,
Henan, Helongjiang, Hubei, Hunan, Jiangsu, Jiangxi, Jilin, Liaoning, Shaanxi,
Shangdong, Shanghai, Shanxi, Sichuan, Tianjian, and Zhejiang.
7

2000 survey, CLHLS in 2002 interviewed 4,894 new individuals aged 65–79 and 4,920
new individuals aged 80 or older. The sample for the 2002 survey was 16,057(surviving) respondents aged 65 or older. CLHLS conducted follow-up surveys in 2005
and 2008–2009. Both recorded the health, family, and living conditions of deceased andsurviving respondents.
The sample of elderly individuals surveyed in 2002 and subsequently surveyed
is used for analysis in this study. Only those in rural locales are examined because
China’s male-centric culture may have faded in urban areas. After excluding the urban
elderly, the sample size is 8,670. In addition, 521 respondents reported having nochildren throughout their lives, reducing the sample to 8,149. Because observations are
missing for some variables, sample size differs in different analyses. 3
Table 1 shows descriptive statistics of this study’s main variables. The sample
size for most variables is 8,136. Table 1 lists sample sizes only for variables with sample
sizes differing from 8,136. The sample contains more females (58%) than males(41.8%). The oldest elder was born in 1885 and the youngest in 1939. Years ofschooling ranged from 0 to 25, but the distribution is skewed leftward, as illustrated by
the sample mean (1.368). Residency is scattered across the 22 provinces, with the3

The sample size for analyzing elderly mortality is in most cases 7,269, and the sample
size for analyzing child support to parents ranges from 7,709 to 8,136.
8

largest percentage (15.2%) living in Guangxi and the smallest (0.7%) living in Tianjian.
The children of elderly respondents were born between 1907 and 1983. The
one-child policy initiated in 1979 likely did not affect birth rates, as the youngest elder
in the sample was 40 in 1979 and probably past the procreative phase of life. The
number of reported children (including deceased children) averaged 4.6, and the numberof surviving children averaged 3.6 in 2002. The ratio of males among first births is 0.58,
far higher than the natural ratio at birth. This finding may illustrate male bias in Chinese
culture. In the 2002 round of CLHLS, elderly respondents could report up to 11 children,
but respondents may have been less likely to report daughters than sons.
Less than 7% of the sampled elders received pensions. Approximately 66%
died before the 2008 survey. In 2002, 62% did not reside with children, and on anaverage, they had received 1,034 RMB from their children during the previous year
(2001). Approximately 5%, 16%, and 7% of the sampled elders had no family member
to care for them, share their feelings or opinions, or ask for help when they had
difficulties, respectively. 4 Among 2,259 sampled elders who acknowledged difficulties
with activities of daily living (ADL), 8% had no family to help with basic activities
such as bathing, dressing, toileting, transferring or moving, continence, and feeding.
4“Family members” are the sampled elders’ spouse, son, son’s wife, daughter,daughter’s husband, grandchildren, and grandchildren’s spouses.
93.2 Research methodology
This section discusses methods used to analyze relations between the gender ofadult children and welfare of their elderly parents. It first describes the analysis of therelation between adult children’s gender and elderly parents’ mortality. Second, it
analyzes the relation between gender and support to elderly parents.
I used the Cox proportional hazard model to examine whether adult children’s
gender correlates with parental mortality. The date of the 2002 interview is the entry
date, and the date of death is the exit date for the hazard model of elders who died
before the 2008 survey. Right censoring occurred at the interview date of the 2008
survey for sampled elders who survived through that day. 5
The parametric part of the Cox proportional hazard model isg(c, x, n, x, z) = exp�β0 + β1 c + β2 n + β3 x + β4 z + ε�,(1)where c is one among several indicators of adult children’s gender indicated below andn is the total number of children. 6 In addition, x is a vector of control variables,
including the elder’s age in 2002 divided by 10, its square, and elder’s years ofschooling. Finally, z is a vector of 21 province dummies with Beijing as the reference
5

Because the date of death displays a greater frequency of missing observations thanthe month of death, longevity is measured in months (monthly data) in the hazard
model.
6
I used either the total of reported children or the total of surviving children (until the
2002 survey) for different regressions.
10

category. 7 Indicators of adult children’s gender (c) are one of the following: number ofreported sons (living and deceased), number of surviving sons (until the 2002 survey),
at least one reported son, at least one surviving son, and a boy as the first-born child.
The number of children (𝑛𝑛) is the number of reported children or the number ofsurviving children. I separately estimated the hazard model for fathers and mothers toaccommodate differential baseline hazards and the differential effects of independent
variables between parents’ genders.
This study cannot establish causality between children’s gender and parents’mortality because children’s gender (c) and number of children (𝑛𝑛) would beendogenous in determining parental mortality. As noted, before ultrasound became
widely available and before introduction of the one-child policy, rural Chinese couples
generally assured that they sired at least one son, by having babies until a son was born
(Junhong, 2001). The most recent childbirth in the CLHLS data occurred in 1983, whenultrasound was generally unavailable in rural China for the sampled elders. This implies
that the probability of having another baby was higher if the ratio of boys to total
number of living children for the couple was lower.
Appendix Table 1 confirms endogeneity of number of children (𝑛𝑛) and7

Footnote 2 names the 22 provinces.
11children’s gender (c) by regressing a dummy for the existence of the next child on theratio of boys to the existing number of children with control variables (parental age in2002 divided by 10, its square, years of schooling, and 21 provincial dummies with
Beijing as the reference province). The linear probability model shows that theprobability of having the next child is higher when the ratio of boys to the number ofexisting children is lower for the second through the seventh births, and that thecorrelations are statistically significant for the second to sixth births. This finding is
consistent with previous studies (Johansson & Nygren, 1991; Ebenstein, 2010; Yi et al.,1993).
To examine the path from children’s gender to parental mortality, the relation
between children’s gender and forms of filial support to parents was analyzed.
Following is the linear probability model used:
ns = β0 + β1 c + β2 n + β3 x ′ + β4 z + ε,(2)where ns is one of the indicators of filial support to birth parents, and the independent

variables are the same as in Equation (1). For the outcome variable, I use five dummy
variables indicating filial physical and emotional support to parents: no co-residencewith children, no family member to care for sick parents, no family member to sharefeelings and opinions, no family member to ask for help when the parent has difficulties,
12and no family member to help with ADL. 8 There is one measure of financial support:
the amount of money received from children during 2001. For this outcome variable, Ialso used OLS.
I constructed the six outcome measures using the 2002 survey data. The
independent variables are the same as in Equation (1) except x ′ , which includes adummy for elders currently residing with spouses in addition to the original control
variables x (elder’s age in 2002 divided by 10, its square, and elder’s years ofschooling). Equation (2) is separately estimated for fathers and mothers to reveal
potential differential effects of the right-hand side (RHS) variables between parental
genders.
Children’s gender (c) and number of children (𝑛𝑛) are also endogenous in

Equation (2). Unobserved determinants of parents’ preference for sons would correlatewith both gender of adult children (c) and number of children (𝑛𝑛) as RHS variables
and filial support to birth parents as the outcome variable. Again, this study cannot
establish causality from children’s gender to filial support.
4.

Results

Table 2 presents the coefficient estimates (not odds ratios) from analyzing

8

ADL measures in the 2002 CLHLS are bathing, dressing, toileting, transferring(moving), continence, and feeding.
13

parental mortality. Columns in Table 2 use different measures of children’s gender. The
initial two columns use the number of reported sons after controlling the number ofreported children. The next two use the number of surviving sons after controlling thenumber of surviving children. Results show that number of reported or surviving sonshas no strong correlation with the probability of death for both elderly fathers andmothers.
Columns (5) and (6) use “at least one reported son” and “at least one reported
daughter” as measures of adult children’s gender. Among fathers, “at least one reported
daughter” is associated with lower mortality at the 10% significance level, whereas “atleast one reported son” has no relation with mortality. However, the difference inassociation between the two is not statistically discernible at conventional levels ofconfidence. Among mothers, “at least one reported son” is correlated with higher
mortality at the 10% significance level, whereas “at least one reported daughter” has nocorrelation with mortality. The difference between the two coefficients is notstatistically significant at conventional levels of significance.
Columns (7) and (8) use “at least one surviving son” and “at least one
surviving daughter” as measures of adult children’s gender and find that “at least one
surviving daughter” negatively correlates with maternal mortality at the 10%
14significance level. No other correlations are statistically significant with respect topaternal and maternal mortalities. Finally, Columns (9) and (10) find that “first-born son”
has no correlation with mortality for fathers or mothers.
Overall, no systematic correlation appears between adult children’s gender andparental mortality for fathers and mothers. The existence of one daughter negatively
correlates in some cases with parental mortality, and the existence of one son positively
correlates with maternal mortality in one case. This finding contradicts the hypothesisthat sons are axiomatically associated with a higher level of welfare among elderly
parents.
For control variables, mortality risk non-linearly increases with age for fathers
and mothers. Years of schooling are unrelated to mortality risk for the elderly in thissample. Four provinces (Henan, Guangxi, Shaanxi, and Shangdong) have lower
mortality risks than Beijing (the base province in Equation (1)) only for mothers. There
are no statistically significant differences in mortality risk across provinces for fathers(Table 2 does not report coefficient estimates of province dummies).
I repeated the estimation with stiffer controls for initial health status andchildhood socioeconomic status (SES), but results concerning parental mortality remainrobust after including them (results not reported). Specifically, I standardized responses
15

to 14 questions about ADL and instrumental activities of daily living (IADL) acrosselders sampled in 2002 and calculated the individual average over 14 responses. I used
this individual average of ADL and IADL responses as the measure of initial health in
2002. Furthermore, I created five dummies from self-reported health in the 2002 survey(one is the reference category in the regressions). For childhood SES, a dummy
indicates whether respondents’ primary water source during childhood was tap water ora well versus a river, lake, spring, pond, or pool. Another measure for childhood SES isa set of three dummies indicating the father’s primary occupation before the age of 60. 9
One additional measure of childhood SES is number of siblings. Although initial healthin 2002 strongly correlates with the subsequent mortality of sampled elders, controlling
initial health and childhood SES does not affect the correlation between parental
mortality and children’s gender. 10
9

One dummy indicates employment in agriculture, fisheries, forestry, or animalhusbandry. The second dummy indicates whether respondents’ fathers were professional,
technical, governmental, institutional, or managerial personnel, staff/service workers,
industrial workers, self-employed, or military personnel. The third dummy indicates
employment at home, unemployed, other, or missing. Overwhelmingly (88%),
respondents reported farming as their father’s main job before the age of 60.
10
Preferably, childhood health and childhood SES should be controlled. Because
measures of childhood health are unavailable in the CLHLS, I used health measures in2002 as indicators of initial health. However, health status in 2002 may be endogenous
in examining the relation between parental mortality and child gender. For example,
couples who have only daughters may work more to save for old age, and greater work
effort might affect health in 2002 and subsequent mortality. If children’s gender does
affect parents’ mortality, it might gradually affect parents’ health since the time when
they have babies. For this reason, I prefer the results without controlling for initialhealth in 2002. Moreover, most measures of childhood SES are not statistically16

I also controlled for “male main occupation before the age of 60” as a proxy ofhousehold wealth among the sampled elderly. Specifically, own occupation is used formale elders, and spouse’s occupation is used for female elders except for those who
remained unmarried throughout life, where own occupation is used even for female
elders. Four dummies indicate “male’s main occupation before the age of 60.” 11 Thisadditional set of controls does not meaningfully change the results (results not reported).
Table 3 presents the results from analyzing the relation between children’s
gender and filial support to birth parents. As measures of children’s gender, number ofsurviving sons, at least one surviving son, at least one surviving daughter, and first-born
son are used. Table 3 does not report estimates of coefficients of the control variables(elder’s age in 2002 divided by 10, its square, elder’s years of schooling, and a dummy
for living with spouse). For each dependent variable, three regressions are separately
estimated for fathers and mothers. The first row uses the number of surviving sons, thesecond row using “at least one surviving son” and “at least one surviving daughter,” andthe last row using “first-born male” as measures of adult children’s gender after

significant in the mortality regressions.
11
One dummy indicates employment in agriculture, fisheries, forestry, or animalhusbandry. The second dummy indicates whether respondents (or spouses) were
professional, technical, governmental, institutional, or managerial personnel. The third
dummy indicates whether respondents (or spouses) were staff/service workers,
industrial workers, self-employed, or military personnel. The fourth dummy indicates
employment at home, unemployed, other, or missing.
17

controlling for the number of surviving children.
In Column (1), the probability that mothers reside without children declines as
the number of surviving sons rises (after controlling the number of surviving children).
In addition, fathers and mothers with “at least one surviving son” are less likely to live
without children, whereas “at least one surviving daughter” is uncorrelated to livingwith adult children. Equality of estimated coefficients of “at least one surviving son”
and “at least one surviving daughter” is strongly rejected with a p-value of near 0 for
both fathers and mothers. The gender of the first-born child is unrelated to theprobability of not residing with children for both fathers and mothers. This finding is
consistent with the traditional patrilocal nature of living arrangements among China’s
rural elderly, who are “supposed” to live with a son. However, the gender of the firstborn seems irrelevant to parental co-residence with children.
Column (2) shows that the amount of money received from adult children
correlates with none of the measures of children’s gender after controlling for thenumber of surviving children. In Panel 2, the difference between coefficients of “at least
one surviving son” and “at least one surviving daughter” also is not statisticallysignificant at conventional levels.
Results in Column (3) indicate that number of surviving sons does not affect18

the availability of family care for sick elderly parents, but “at least one surviving son”
and “at least one surviving daughter” correlate with providing care for sick parents.
However, the effect of “at least one surviving son” exceeds that of “at least one
surviving daughter,” with the equality of the coefficient estimates rejected at 9% forfathers and at nearly 0% for mothers.
Results in Columns (4) (emotional support) and (5) (general support) indicatethat the existence of both sons and daughters is correlated with the availability of filial
support for mothers, but that only the existence of sons is correlated with filial support
at conventional levels for fathers. For both fathers and mothers, the effect of sons isstronger than that of daughters with the equality of the coefficient estimates rejected atconventional levels.
Sample sizes in Column (6) dwindle because responses are available only fromsampled elders with at least some ADL difficulties. Incidence of ADL difficulties ismore frequent among elderly females than males (Murtagh & Hubert, 2004; Crimmins,
Kim, & Solé-Auró, 2011; Gu & Zeng, 2004). Average lifespan is longer for females
than for males. Thus, the sample size for mothers (1,571) is more than double than thatof fathers (688). Given the smaller sample of fathers, no statistically significant results
emerge. Among mothers, both “at least one surviving son” and “at least one surviving
19

daughter” are associated with a lower probability of no family support for ADL
difficulties. The presence of a son has a larger effect than the presence of a daughter,although the difference is not statistically significant at conventional levels.
Correlations between filial support and adult children’s gender remain robust
after including the additional control for an elder’s health in 2002 (results not reported).
Because health may be endogenous in determining the relation between filial support
and adult children’s gender, results without controlling for the elder’s health seem
preferable. 125.

Conclusion

Previous studies have documented the patrilocal, patrilineal, and patriarchal
natures of Chinese culture and have found evidence that Chinese couples had madefertility decisions that resulted in an abnormally high birth ratio of boys to girls. Without
a doubt, couples’ desire for male offspring relates to concerns about their security in oldage. It is believed that under patrilocal living arrangements, elderly parents reside withand are financially, physically, and mentally supported by their eldest son and daughter
in-law. In contrast, it is believed that a daughter joins her husband’s family upon
marriage, and therefore birth parents cannot rely on daughters for support in old age.
12First, the elder’s health and filial support may be determined simultaneously. Second,children’s gender may affect the elder’s health, which in turn may affect filial support.
See footnote 10.
20This widely spread image that sons and not daughters provide security for their elderly
parents in rural China suggests that the gender of adult children is relevant to themortality of elderly parents. However, this study finds no correlation between thegender of adult children and the mortality of elderly parents in rural China. It shows thatmany daughters financially, physically, and emotionally contribute to their elderly birthparents.
The following description of filial support from sons vis-à-vis daughters is
consistent with this study’s findings: the patrilocal custom of living arrangements isprevalent in rural China; therefore, sons are much more likely to live with elderly
parents, and daughters’ contribution in this regard is minimal. However, daughters’
financial contributions are statistically no different from sons’. Sons’ contributions toelderly parents’ emotional and physical support typically exceed daughters’, andsometimes the differences are discernible statistically. Even so, daughters’ contributions
are not negligible. Findings in this study are consistent with previous studies concerning
adult children’s support for elderly parents in Taiwan using adult children as the unit ofanalysis (Lee, Parish, & Willis 1994; Lin et al., 2003).
Findings of this study suggest that the traditional interpretation ofson-dominated filial support to parents is overblown. This study suggests a more21

realistic position of daughters vis-à-vis sons: Sons are the primary providers of filial
support to birth parents, whereas daughters provide supplemental support or primary
support if sons cannot take the responsibility. In contravention of conventional wisdom,
many daughters remain in contact with birth parents after marriage.
This paper cannot establish causality between the gender of adult children andthe mortality of elderly parents or between the gender of adult children and availabilityof filial support. This is because rural Chinese couples could manipulate somewhat thegender of their offspring before ultrasound became widely available, and couples today
can almost perfectly control the gender of their offspring with wider access toultrasound technology. An innovative study design is necessary to establish causality.
If daughters’ contributions to birth parents were traditionally non-existent andhave recently arisen as the result of economic development, including improved
transportation and communication, this may help to elevate daughters’ traditionallylesser status in rural Chinese households. If parents realize that daughters contribute totheir old-age security, their motivation to invest in sons rather than daughters could
diminish.

22

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in rural Bangladesh. Population Studies, 53, 227–235.

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in rural Bangladesh. Journal of biosocial science, 32, 89–98.

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24school enrollment? World Development, 34, 1639–1653.

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Yi, Z., Ping, T., Baochang, G., Yi, X., Bohua, L., & Yongpiing, L. (1993). Causes andimplications of the recent increase in the reported sex ratio at birth in China. Population
and Development Review, 19, 283–302.

25

Table 1: Descriptive statistics of elderly individuals
Mean
Female (dummy)

Std. dev.

Min

Max

11.912 1885

1939

0.582

Birth year

1915.244

Age in 2002

86.21511.90962

116

Years of schooling

1.368

2.635

0

25

Province of residence (dummy):
Beijing

0.008

Tianjian

0.007

Hebei

0.013

Shanxi

0.011

26

N

Liaoning

0.031

Jilin

0.024

Helongjiang

0.013

Shanghai

0.010

Jiangsu

0.129

Zhejiang

0.091

Anhui

0.061

Fujian0.029

Jiangxi

0.027

Shangdong

0.049

Henan

0.052

27

Hubei

0.033

Hunan

0.053

Guangdong

0.059

Guangxi

0.152

Chongqing

0.036

Sichuan

0.098

Shaanxi

0.016Year of the first birth

1938.961

12.742 1907

1978 7,990

Year of the last birth

1954.033

12.732 1910

1983 7,564

Num. reported sons

2.434

1.534

0

9

Num. reported daughters

2.194

1.558

0

10

28

Num. reported children

4.628

2.184

1

11

Num. surviving sons

1.850

1.381

0

8

Num. surviving daughters

1.716

1.377

0

8

Num. surviving children

3.566

1.963

0

10

First child is boy (dummy)

0.579

8,098

Dead by the 2008 survey (dummy)

0.659

7,269

Have pension (dummy)

0.064

8,130

No co-residence with child (dummy)

0.624

7,709

Amount of money received from children

1034.391 1425.287

No family member to take care when sick (dummy)

0.052

No family member to share feelings (dummy)

0.161

0 40000 7,997
8,123

29

No family member to ask for help (dummy)

0.068

No family member to help ADL difficulties (dummy)

0.084

1.

2,259

The sample size is 8,136 elderly individuals, except when another sample size is indicated.

30

Table 2: Parental mortality (Cox proportional hazard model)

Num. reported sonsNum. surviving sons

(1)

(2)

(3)

(4)

(5)

(6)

Fathers

Mothers

Fathers

Mothers

Fathers

Mothers

0.016

0.002

(0.022)(0.018)
0.005

0.006

(0.025)(0.021)-0.0540.110*(0.075)(0.062)-0.117*-0.009(0.071)(0.058)At least one reported son

At least one reported girl31

Num. reported children

0.002-0.027**0.026**-0.033***(0.015)(0.012)(0.013)(0.010)

Num. surviving children

Age/10(Age/10)2

Years of schooling-0.009-0.025*(0.017)(0.015)3.190***

3.003***

3.191***

3.018***

3.174***

3.018***(0.389)(0.319)(0.388)(0.320)(0.389)(0.319)-0.139***-0.127***-0.140***-0.128***-0.139***-0.128***(0.022)(0.018)(0.022)(0.018)(0.022)(0.018)-0.0010.010-0.001

0.011-0.001

0.011(0.008)(0.016)(0.008)(0.016)(0.008)(0.016)

32

N

3,085

4,184

3,085

4,184

3,085

4,184

1. Standard errors are in parentheses.
2. *** p < 0.01, ** p < 0.05, * p < 0.1
3. All explanatory variables are measured as of 2002.
4. Twenty-one province dummies are also included as controls, but the coefficient estimates are not reported in the table. (Beijing is
the reference province.)

33

At least one surviving son

At least one surviving girlFirst child boy

Num. reported children

(7)

(8)

(9)

(10)

Fathers

Mothers

Fathers

Mothers-0.0730.033(0.068)(0.052)-0.099-0.086*(0.066)(0.052)-0.0690.057(0.048)(0.039)0.006-0.028***(0.011)(0.008)34

Num. surviving children

Age/10(Age/10)2

Years of schooling

N

0.015

-0.014(0.016)(0.015)3.161***

3.036***

3.337***

3.009***(0.389)(0.320)(0.402)(0.325)-0.138***-0.129***-0.147***-0.127***(0.022)(0.018)(0.023)(0.018)-0.0010.010-0.0020.007(0.008)(0.016)(0.008)(0.017)

3,085

4,184

2,915

4,025

1. Standard errors are in parentheses.
2. *** p < 0.01, ** p < 0.05, * p < 0.1
3. All explanatory variables are measured as of 2002.
35

4.

Twenty-one province dummies are also included as controls, but estimated coefficient are not reported in the table. (Beijing is thereference province.)

36

Table 3: Adult children’s support to parents (OLS)(1)

Dependent variable

Fathers or mothers?

(2)

(3)

Amount of money received

No family member to take carefrom children

when sickNo co-residence with children

Fathers

Mothers

Fathers

Mothers

Fathers

Mothers-0.012-0.037***13.2136.227-0.001-0.005(0.008)(0.007)(22.022)(21.794)(0.004)(0.003)-0.000-0.00694.044***82.723***-0.014***-0.013***(0.006)(0.006)(16.290)(15.627)(0.003)(0.002)

3,270

4,439

3,361

4,636

3,398

4,725

Num. surviving sonsNum. surviving childrenN37

-0.197***-0.293***19.05833.741-0.047***-0.071***(0.028)(0.024)(72.395)(62.801)(0.012)(0.010)

0.033

0.001

9.571

-10.502-0.022**-0.018*(0.026)(0.022)(67.167)(59.324)(0.011)(0.009)0.003-0.00498.380***83.675***-0.008***-0.006**(0.005)(0.005)(14.933)(15.073)(0.002)(0.002)p-val. from test of equality

0.000

0.000

0.915

0.557

0.082

0.000

N

3,270

4,439

3,361

4,636

3,398

4,725

0.006

0.002

12.318-2.5870.002-0.004(0.016)(0.015)(44.950)(41.826)(0.007)(0.007)At least one surviving son

At

least

one

survivingdaughterNum. surviving children

First child boy

38

-0.006-0.024***101.361***86.358***-0.014***-0.014***(0.005)(0.004)(11.947)(11.522)(0.002)(0.002)

3,262

4,420

3,352

4,610

3,389

4,696

Num. surviving children

N

1. Standard errors are in parentheses.
2. *** p < 0.01, ** p < 0.05, * p < 0.1
3. Age divided by 10, age divided by 10 squared, years of schooling, a dummy of currently living with spouse, and 21 province
dummies are included as controls in all regressions. (Beijing is the reference province.)

39

(4)

(5)

(6)

No family member to share

No family member to ask forNo family member to help ADL

feelings

help

difficulties

Dependent variable

Fathers or mothers?

Fathers

Mothers

Fathers

Mothers

Fathers

Mothers-0.004-0.010*-0.007*-0.008**-0.003-0.001(0.006)(0.006)(0.004)(0.004)(0.011)(0.008)-0.011**-0.016***-0.012***-0.014***-0.009-0.017***(0.004)(0.004)(0.003)(0.003)(0.008)(0.005)

3,404

4,732

3,404

4,732

688

1,571

Num. surviving sonsNum. surviving childrenN40

-0.066***-0.088***-0.080***-0.088***-0.026-0.079***(0.019)(0.016)(0.013)(0.011)(0.030)(0.020)-0.024-0.035**-0.019-0.028***-0.027-0.061***(0.018)(0.015)(0.012)(0.010)(0.030)(0.019)-0.005-0.008**-0.007***-0.006**-0.0060.000(0.004)(0.004)(0.003)(0.003)(0.007)(0.005)p-val. from test of equality

0.068

0.005

0.000

0.000

0.980

0.451

N

3,404

4,732

3,404

4,732

688

1,571

-0.011-0.003-0.015*-0.008-0.018-0.008(0.012)(0.011)(0.008)(0.007)(0.020)(0.014)At least one surviving son

At

least

one

survivingdaughterNum. surviving children

First child boy

41

-0.013***-0.020***-0.016***-0.017***-0.011**-0.018***(0.003)(0.003)(0.002)(0.002)(0.005)(0.004)

3,395

4,703

3,395

4,703

684

1,557

Num. surviving children

N

1. Standard errors are in parentheses.
2. *** p < 0.01, ** p < 0.05, * p < 0.1
3. Age divided by 10, age divided by 10 squared, years of schooling, a dummy for currently living with spouse, and 21 province
dummies are included as controls in all regressions (Beijing is the reference province).

42

Appendix

Appendix Table 1: Endogenous nature of gender of adult children (OLS)

Ratio existing boys

Parent age/10(Parent age/10)2

2nd child

3rd child

4th child

5th child

6th child

7th child-0.037***-0.049***-0.074***-0.099***-0.080**-0.064(0.006)(0.010)(0.016)(0.024)(0.033)(0.045)-0.101***-0.032

0.085

0.040

0.158

0.400***(0.039)(0.049)(0.064)(0.082)(0.104)(0.135)0.005**

0.001-0.006-0.002-0.009-0.021***(0.002)(0.003)(0.004)(0.005)(0.006)(0.008)43

Years of schoolingN-0.001-0.003*-0.005***-0.002-0.006*-0.009**(0.001)(0.001)(0.002)(0.002)(0.003)(0.004)

7,849

7,296

6,521

5,418

3,998

2,682

1. Standard errors are in parentheses.
2. *** p < 0.01, ** p < 0.05, * p < 0.1
3. Twenty-one province dummies are included as controls, but coefficient estimates are not reported (Beijing is the reference
province).
4. Parent age in 2002.

44