Abstract
This study examines how an increase in home value affects fertility decisions of homeowners in China by exploiting regional heterogeneity in housing markets driven by local regulatory and geographic land constraints. In sharp contrast to the literature on developed countries, our instrumental variable results show a negative fertility response to house value growth driven by the recent housing boom in China, where a 100,000-yuan increase in lagged home values—about 43% of the average housing wealth at baseline—results in a 14% decrease in the likelihood of home-owning women giving birth. Further evidence suggests that underdeveloped credit markets may suppress the positive wealth effect of house value growth on childbearing.
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Notes
In this paper, we use the terms house price, house value, and home value interchangeably when referring to the total market price of a dwelling.
Under this policy, each couple was generally allowed to have only one child, especially in urban areas; couples meeting some eligibility criteria (e.g., ethnic minorities and rural couples whose first child was a girl) could bear two children, or otherwise, they had to pay a fine for above-quota births.
Wang et al. (2017) simulate that the fertility level would not substantially rebound under the two-child policy, implying that the OCP may not be a strictly binding factor when couples make fertility decisions in our study period.
Many emerging economies, such as South Korea, have experienced a housing boom during their rapid-growth transition periods (Chen and Wen 2017), and their fertility effects of home value increases deserve further exploration.
Evidence from developed countries has shown that housing price variations are important for multiple outcomes, including savings, consumption, education, labor supply, marriage, and long-term care insurance demand (e.g., Davidoff 2010; Farnham et al. 2011; Lovenheim 2011; Johnson 2014; Aladangady 2017; Harding and Rosenthal 2017; Zhao and Burge 2017).
All mortgage loans in China are adjustable-rate mortgages with no prepayment penalty.
For example, homeowners can realize their increased housing wealth by selling their homes and moving to a smaller house or a lower-priced real estate market in the future.
The indices include account ownership, making or receiving digital payments, saving at formal financial institutions, borrowers relying on formal credit, the capability of raising emergency funds through savings, among others.
Another potential reason is that housing cost is likely to be a greater portion of the cost of raising a child in developing countries. However, it does not seem plausible for China. Rogoff and Yang (2020) show that housing-related expenditure accounts for 23% of household consumption expenditure in China, which is similar to an average of around 22% in OECD and European Union countries in 2019 (OECD 2021).
The periods of \(t-2\), \(t-1\), and \(t\) then correspond with the waves of 2010, 2012, and 2014, the waves of 2012, 2014, and 2016, or the waves of 2014, 2016, and 2018, respectively. Following the method used by Ding and Lehrer (2010), we examine the potential bias related to panel attrition and find that women who were lost to follow-up are not systematically different from the remaining sample in terms of childbearing decisions.
This means that a woman should have reached age 20, the minimum legal marriage age in China, when first observed in wave \(t-2\). As a robustness check, we use the sample aged 20 to 45 (i.e., aged 16 to 41 in \(t-2\)) and obtain similar results.
Homeownership status is defined by owning a home solely or jointly with working units during the period from \(t-2\) to \(t-1\). The CFPS does not ask respondents directly whether they have moved from \(t-2\) to \(t-1\). We exclude those who moved across communities and within communities (i.e., the property right of the home is newly obtained, or the living spaces are different) between \(t-2\) to \(t-1\). Please refer to Appendix Table 11 for the detailed description of the sample selection.
The sample size for the main regression is larger than the number of women because some women have been observed in four or five waves. The results are robust to limiting the sample to unique observations, that is, including a woman only in her last three observed waves.
A detailed description of the CFPS wealth imputation can be found in Jin and Xie’s (2014) work. As a robustness check, we exclude households with imputed home values and obtain similar results as the main results.
The measure of home equity change is not used here for two reasons. First, whether having a mortgage loan and what the loan amount is may be endogenous. Second, only the CFPS of 2012 provides separate information on mortgages for owner-occupied units and other housing assets, while other waves ask about the total amount of house mortgage loans. In CFPS 2012, about 6.13% had a mortgage loan for the current residence, with an average amount of 78,754 yuan and a median amount of 57,297 yuan.
In China’s real estate market, there have been temporary slowdowns across cities between 2010 and 2014 because of tightening policies and administrative measures (Koss and Shi 2018). Thus, we observe a large proportion of perceived property value loss in the data. We have also decomposed home value changes in Eq. (1) into two variables for increases and decreases in home value. The results, not presented in this article, show little evidence that housing gains and losses have asymmetric effects on fertility in the context of China’s great housing boom.
We restrict the sample to women staying in the same housing unit during the period from \(t-2\) to \(t-1\) so that the change in home value is mostly driven by housing market fluctuations and not by moving residences. In the sensitivity analysis, we will show the robustness of our results to the inclusion of these movers.
The community questionnaire has been administered in CFPS 2010 and 2014.
The fertility policy adjustment and labor market shocks are generally homogeneous within a county in China.
This variable is derived from the Remotely Sensed Data of Land Use in China, collected by the Institute of Geographic Sciences and Natural Resources Research from The Chinese Academy of Sciences.
This variable is obtained from the China Stock Market & Accounting Research (CSMAR) Regional Economy Database.
We replace the total water area in the instrument with the percentage of water area in each prefectural city. The IV results are remarkably close to the results presented in Table 3, but the instrument becomes weaker in the first stage.
The Anderson–Rubin tests are implemented using the rivtest package in Stata.
Here total housing wealth refers to home equity of all housing units.
There are more than 520 communities in our sample, so it implies a large set of control variables given our limited sample size.
The upper and lower bounds are calculated using the imperfectiv command in Stata (Clarke and Matta 2018).
We also apply a permutation test to examine the exogeneity of our instrument. In the reduced-form regression of fertility on the excluded IV and other covariates, we randomly assign values to water area (or to budgetary revenue per capita). If budgetary revenue per capita (or water area) is correlated with the error term, we may observe a spurious relationship between the permuted IV and fertility. The distributions of t values for the excluded IV from permutating water area or budgetary revenue per capita for 1,500 times show that the p values are both smaller than 0.10 (p = 0.06 and 0.08), implying that the negative effect of the excluded IV on fertility in the reduced-form regression is very likely to be true rather than a biased result from an endogenous IV.
As the CFPS provides no information on household heads, the financial respondent is regarded as the household head. Usually, in Chinese culture, parents buy houses for their adult children, or married couples continue living in the parents’ house. In our data, about 65% of the respondents are household heads or spouses of household heads, 4.6% are daughters of household heads, and 21% are daughters-in-law of household heads.
In this analysis, we exclude those with missing information on community IDs and communities with fewer than eight households.
This is consistent with the literature’s estimates of approximately 90% of homeownership in China (Glaeser et al. 2017).
Dettling and Kearney (2014) also reveal a negative coefficient of Metropolitan Statistical Area (MSA)-level house price changes for renters and a positive estimate for homeowners relative to renters. However, contrastingly, they find a net positive effect on births among homeowners, which is indicative of the dominant wealth effects. However, given the net negative effect of community-level median home value changes among homeowners in China, we cannot rule out the possibility that the positive coefficient on \({\Delta HV}_{mct-1}\times {Own}_{imct-1}\) may be driven by the difference in the cost effects between renters and homeowners.
As a robustness check, we also estimate the results using the sample of women with children. The results are qualitatively similar to those reported in Table 6.
We exclude the top 0.5% of the observations with the highest level of household education expenditures. The average education expenditure is 4,911 yuan in the sample of homeowners with only one child.
Greater internet index values imply more advanced credit markets. Details of the index construction are explained in https://en.idf.pku.edu.cn/achievements/seriesofdigitafinanceindexes/490847.htm.
We include interaction terms for every subgroup so that the coefficient on each interaction term shows the estimated effect for the corresponding subgroup.
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Acknowledgements
The authors thank Wei Huang, Junjian Yi, Fangwen Lu, and seminar participants at the Peking University, Renmin University of China, Fudan University, and Central University of Finance and Economics for helpful comments. The guidance of editor Terra McKinnish and the anonymous referee is greatly appreciated.
Funding
This study was supported by the Key Project of the National Social Science Fund (No. 21ZDA098), the National Natural Science Foundation of China (No. 71673313 and No. 71703160), and the “Thematic Research Project on China’s Income Distribution” (No. 21XNLG03) with funding from the research fund of Renmin University of China.
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Liu, H., Liu, L. & Wang, F. Housing wealth and fertility: evidence from China. J Popul Econ 36, 359–395 (2023). https://doi.org/10.1007/s00148-021-00879-6
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DOI: https://doi.org/10.1007/s00148-021-00879-6