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How private sector participation improves retirement preparation: A case from China

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Abstract

This paper shows empirically how private sector participation improves the adequacy and equality of retirement preparation in a three-pillar retirement system. We develop a three-layer replacement rate approach based on the China Health and Retirement Longitudinal Study, a nationwide representative household survey of middle-aged and old aged population. Our empirical evidence shows that private sector participation increased the mean (median) replacement rate in 2013 from 35.4% (15.4%) to 69.8% (48.7%). The evidence also suggests that annuitising home equity is responsible for a large portion of this increase. Surprisingly, private sector participation also mitigates the inequality of retirement preparation between the formal and informal sectors. Our empirical findings emphasise the importance of annuitisable private savings for the retirement income security of the one-fifth of the global population who live in a representatively high-growth and rapidly ageing economy.

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Notes: PAYGO: pay as you go; PF: partially funded; FF: fully funded

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Notes

  1. See Bridges and Choudhury (2009), Neelakantan and Chang (2010), and Knoef et al. (2016) for discussions on the protection gaps of minorities, females, and immigrants, respectively.

  2. Wang et al. (2014a, b) present and discuss the looming financing crisis and the unfairness in the Chinese pension system.

  3. Conventionally, private sector participation in retirement preparation involves Pillar II, i.e. employer-provided private pensions, and Pillar III, which includes individually arranged private pensions and potentially annuitisable private savings. In this paper, we further consider annuitising private savings through the additional premium contribution options in the public pension programme (Resident’s Basic Pension Programme, Pillar I) as a special type of private sector participation.

  4. See, e.g. Wang et al. (2014a, b), Song et al. (2015), and Chen and Groenewold (2017).

  5. Two consulting reports in Chinese also contribute to understanding Chinese retirement adequacy (Insurance Association of China 2015; China Center for Insurance and Risk Management of Tsinghua University 2016). Our analyses fundamentally differ from theirs in terms of methodology, sample representativeness, and findings. Feng et al. (2011) explore micro-level data; however, their focus is on the trade-off between public pension and household savings for urban residents.

  6. Monetary transfer and services from children to parents are not included in our replacement rate estimation; rather, we control for these elements in a second-stage regression to reveal the substitution effects between family transfer and pension benefits and savings. We refer to Cai et al. (2006) and Oliveira (2016) for a closer examination of family transfer.

  7. Forteza and Ourens (2012) develop an indicator describing the return to contribution of Latin American pension programmes; however, they do not focus on retirement preparation.

  8. Wang et al. (2014a, p. 25).

  9. The Rural Resident’s Basic Pension and Urban Resident’s Basic Pension were merged to form the Resident’s Basic Pension in China in 2014. Thus, we do not distinguish the detailed differences in the premium contribution and the pension benefits of these programmes but instead aggregate and analyse them as if they had always been one programme, i.e., the Resident’s Basic Pension.

  10. The spouse of the primary respondent may be younger than 45 but is also interviewed as a respondent.

  11. In the main part of the paper, we use CHARLS (2013) to conduct most of our analyses. We repeat all analyses with CHARLS (2011, 2015) in a robustness test, and all results support our conclusions (see the section on robustness tests). CHARLS (2011) contains limited information on the Resident’s Basic Pension, which was underdeveloped with a low coverage rate in 2011 and may thus result in sample selection bias. CHARLS (2015) was partially released in May 2017; however, the sub-data set of family information has not yet been released. We would have to approximate the number of children and number of siblings to use CHARLS (2015). In the regression analyses, a full sample result including all three waves is also presented.

  12. The minimum statutory retirement age is 40 for workers involved in unhealthy special activities, and the maximum statutory retirement age is 70 for senior government officials and other senior positions; the regular statutory retirement age is 60 for males, 50 for female workers, 55 for female cadres, and 60 for other females.

  13. Before the third step, we supplement missing values by taking full advantage of information in CHARLS (2011, 2013, 2014 and 2015). For example, we supplement missing values of time-invariant variables in one year with the corresponding values in another year.

  14. Hurd and Rohwedder (2011). Alternatively, we truncate the replacement rate in percentiles of 2.5/97.5 and 1/99. The results are consistent with our conclusions and are available from the authors on request.

  15. The observations dropped from Steps 1, 2, 3, and 4 are 13%, 16%, 47%, and 10% of the previous step sample, respectively.

  16. The three layers of replacement rates are defined differently from the three pillars of the retirement system. The 2-Layer PR includes the public pension (Pillar I), the employer-provided private pension (Pillar II), and individually arranged private pension (a part of Pillar III). The 3-Layer PR additionally includes the annuitisable private savings.

  17. We also calculate the household replacement rate as a robustness test, and the results support our conclusions.

  18. Both programmes have a design of income redistribution. The benefits of the Employee’s Basic Pension are partially determined based on the average employment income and the years of contribution (the part from the pooling account), while the yearly premium is calculated as a percentage of individual employment income. Thus, for the same amount of benefits, high-income employees contribute a greater premium than low-income employees. The redistribution effect of the Resident’s Basic Pension is indirect, as reflected by the general tax subsidy, to which rich people contribute more than poor people.

  19. For 1281 respondents, the 3-Layer PRs are at least twice the 2-Layer PR, among 2372 respondents in Sample B.

  20. Yuh (2011), Knoef et al. (2016). Caution must be observed when comparing these fractions because the replacement rate estimations differ in several aspects. However, the differences do not change the relative positions of the three countries.

  21. The coefficient of financial assets remains insignificant in 2013 and in the full sample even if we drop the home equity variable in the regression equation.

  22. We note that this equality-improving effect of private savings does not vary across gender or over generations, as evidenced by the insignificant coefficients of the interaction terms Formal sector × Birth year and Formal sector × Male in Column 6, Table 5.

  23. Low fertility rates in recent years may weaken the trend of raising children for old-age support in China. An alternative explanation might be that parents do not have spare money to contribute to pension programmes when they have more children.

  24. Following Yuh (2011), we use the characteristics of the main respondent in one family as the characteristics of that family, including occupational sector, birth year, education, and ethnic group.

  25. The majority is actually the “exception” because the respondents in our sample were mostly born between 1944 and 1964, and thus, their children were mostly born between 1960 and 1995. The one-child policy was formally implemented in the late 1970s, and thus, less than half of the respondents were restricted by this policy. Moreover, exceptions are widely present in rural China, as the one-child policy officially allows for two children for rural residents if the first child is a girl.

  26. Wang et al. (2014a, p. 33).

  27. We exclude from this subsample respondents who retired after the age of 60 because most of them had the Resident’s Basic Pension benefits for free, without contributions when the programme was rolled out.

  28. For example, a person who retires at 60 has paid CNY 200 for 12 years as a premium contribution to the Resident’s Basic Pension programme. We simulate a scenario in which they would pay an additional CNY 750 a year (i.e. 950 in total) for an additional 10 years (i.e. 22 years in total), and thus, the simulated increments are CNY 3817. The calculation formulas and standard tariff are documented in Ministry of Human Resources and Social Security (2014).

  29. Ideally, the net (after-tax) pre-retirement permanent income should decrease as the premium contribution to the public pension programme increases. We, however, do not consider this impact on income. This simplification will not affect our conclusion because a decreasing permanent income will further increase the new 1-layer PR, reinforcing our conclusions. We truncate the non-zero replacement rates at the 5th and 95th percentiles (Hurd and Rohwedder 2011).

  30. In Sample C, the average yearly premium contribution is CNY 232, and the average number of years of contribution is 7.8.

  31. Participants of the Employee’s Basic Pension have, on average, 21 years of contribution in Sample A, a minimum of 15 years of contribution to receive any pension benefits, and a fixed individual payroll tax of 8%. If participants in the Resident’s Basic Pension follow the same contribution plan, then the yearly premium contribution should be CNY 988 (8% × CNY 12,354, the average income from work in Sample C). Thus, we simulate the scenario of additional CNY 750 premiums for an additional 10 years, which approximates the 8% contribution for 18 years (between 15 and 21 years).

  32. The comparison is based on the new 1-layer PR scenario with the additional CNY 750 premium contribution for an additional 10 years (mean 64.0% and median 45.0%). The original 3-layer PR for the informal sector in 2013 is 59.1% (mean) and 37.7% (median).

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Acknowledgements

The authors thank Christian Biener, Martin Eling, Katja Hanewald, Peng Jing, Chong Liu, Gene C. Lai, Ze Song, Yi Yao, and participants of the ARIA and APRIA conferences in 2016 for their helpful inputs and comments. This study is supported by the key grant of Chinese Ministry of Education - Commercial Pension in China: Institution and Operation (Grant No. 14JZD027), the National Natural Science Foundation of China (Grant No. 71703003), and the seeds fund of School of Economics, Peking University.

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  1. Department of Risk Management and Insurance, School of Economics, Peking University, Yiheyuan Rd. 5, Beijing, 100871, China

    Wei Zheng, Zining Liu & Ruo Jia

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  1. Wei Zheng
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Correspondence to Ruo Jia.

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Zheng, W., Liu, Z. & Jia, R. How private sector participation improves retirement preparation: A case from China. Geneva Pap Risk Insur Issues Pract 44, 123–147 (2019). https://doi.org/10.1057/s41288-018-0110-7

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