Trajectories to high income: Growth dynamics in Japan, the People's Republic of China, and the Republic of Korea

Murach, Michael; Wagner, Helmut; Kim, Jungsuk; Park, Donghyun Working Paper Trajectories to high income: Growth dynamics in Japan, the People's Republic of China, and the Republic of Korea ADB Economics Working Paper Series, No. 622 Provided in Cooperation with: Asian Development Bank (ADB), Manila Suggested Citation: Murach, Michael; Wagner, Helmut; Kim, Jungsuk; Park, Donghyun (2020) : Trajectories to high income: Growth dynamics in Japan, the People's Republic of China, and the Republic of Korea, ADB Economics Working Paper Series, No. 622, Asian Development Bank (ADB), Manila, https://doi.org/10.22617/WPS200276-2 This Version is available at: https://hdl.handle.net/10419/246699 Standard-Nutzungsbedingungen: Die Dokumente auf EconStor dürfen zu eigenen wissenschaftlichen Zwecken und zum Privatgebrauch gespeichert und kopiert werden. 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If the documents have been made available under an Open Content Licence (especially Creative Commons Licences), you may exercise further usage rights as specified in the indicated licence. https://creativecommons.org/licenses/by/3.0/igo/ ASIAN DEVELOPMENT BANK ASIAN DEVELOPMENT BANK 6 ADB Avenue, Mandaluyong City 1550 Metro Manila, Philippines www.adb.org Trajectories to High Income Growth Dynamics in Japan, the People’s Republic of China, and the Republic of Korea Japan and the Republic of Korea quickly graduated from middle-income to high-income status and thus offer valuable lessons for the People’s Republic of China (PRC). This paper analyzes the economic growth patterns of the PRC, Japan, and the Republic of Korea in the postwar period. The authors use Cobb–Douglas production functions to assess the long-run equilibrium relationships between per capita gross domestic product, capital, and labor by means of cointegrated vector autoregressive models. The analysis uncovers a striking similarity between the growth experiences of the PRC and the Republic of Korea. About the Asian Development Bank ADB is committed to achieving a prosperous, inclusive, resilient, and sustainable Asia and the Pacific, while sustaining its efforts to eradicate extreme poverty. Established in 1966, it is owned by 68 members —49 from the region. Its main instruments for helping its developing member countries are policy dialogue, loans, equity investments, guarantees, grants, and technical assistance. TRAJECTORIES TO HIGH INCOME GROWTH DYNAMICS IN JAPAN, THE PEOPLE’S REPUBLIC OF CHINA, AND THE REPUBLIC OF KOREA Michael Murach, Helmut Wagner, Jungsuk Kim, and Donghyun Park ADB ECONOMICS WORKING PAPER SERIES NO. 622 October 2020 ASIAN DEVELOPMENT BANK ADB Economics Working Paper Series Trajectories to High Income: Growth Dynamics in Japan, the People’s Republic of China, and the Republic of Korea Michael Murach, Helmut Wagner, Jungsuk Kim, and Donghyun Park No. 622 | October 2020 Michael Murach ([email protected]) is a research and teaching assistant at FernUniversität in Hagen, Germany. Helmut Wagner (Helmut.Wagner@ fernunihagen.de) is a professor of Economics and President of CEAMeS at FernUniversität in Hagen, Germany. Jungsuk Kim ([email protected]) is a professor at Sejong University, Seoul. Donghyun Park ([email protected]) is a principal economist at the Asian Development Bank, Manila. We would like to thank participants of the 15th Annual Conference of the European Economics and Finance Society, 2016, Amsterdam; the 2nd CEAMeS Workshop on East Asia Macroeconomic Studies, 2018, Xiamen; the 3rd CEAMeS Workshop on Macroeconomic Development and Trade in East Asia, 2019, Hagen; and members of the Chair of Macroeconomics at the FernUniversität in Hagen. Creative Commons Attribution 3.0 IGO license (CC BY 3.0 IGO) © 2020 Asian Development Bank 6 ADB Avenue, Mandaluyong City, 1550 Metro Manila, Philippines Tel +63 2 8632 4444; Fax +63 2 8636 2444 www.adb.org Some rights reserved. Published in 2020. ISSN 2313-6537 (print), 2313-6545 (electronic) Publication Stock No. WPS200276-2 DOI: http://dx.doi.org/10.22617/WPS200276-2 The views expressed in this publication are those of the authors and do not necessarily reflect the views and policies ofthe Asian Development Bank (ADB) or its Board of Governors or the governments they represent. 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Since papers in this series are intended for quick and easy dissemination, the content may or may not be fully edited and may later be modified for final publication. CONTENTS TABLES AND FIGURES iv ABSTRACT vi I. INTRODUCTION 1 II. LITERATURE REVIEW 2 III. EMPIRICAL ANALYSES 6 A. Developing Theoretical Hypotheses 6 B. Data Description and Historical Events 7 C. Univariate Properties of the Data 12 D. Structural Break Tests 12 E. Econometric Framework 13 F. Estimation 14 IV. CONCLUSION 31 APPENDIX 33 REFERENCES 39 TABLES AND FIGURES TABLES 1 Summary of Potential Structural Breakpoints 4 2 Residual Analysis—Diagnostic Testing of the Unrestricted Vector Autoregression 15 (1) Model, People’s Republic of China 3 The Just-Identified Long-Run Cointegration Relations for r = 1, People’s Republic of China 16 4 The Overidentified Long-Run Cointegration Relation for r = 1, People’s Republic of China 16 5 Residual Analysis—Diagnostic Testing of the Unrestricted Vector Autoregression 21 (1) Model, Republic of Korea 6 Likelihood-Ratio Trace Test for the Unrestricted Vector Autoregression (1) Model, 21 Republic of Korea 7 The Just-Identified Long-Run Cointegration Relations for r = 1, Republic of Korea 22 8 The Overidentified Long-Run Cointegration Relations for r = 1, Republic of Korea 22 9 Residual Analysis—Diagnostic Testing of the Unrestricted Vector Autoregression 26 (1) Model, Japan 10 Likelihood-Ratio Trace Test for the Unrestricted Vector Autoregression (1) Model, Japan 26 11 The Just-Identified Long-Run Cointegration Relations for r = 1, Japan 27 12 The Overidentified Long-Run Cointegration Relations for r = 1, Japan 27 A1 Structural Break Tests 33 FIGURES 1 Catching-Up Potential of Japan, the People’s Republic of China, and the Republic of Korea 3 to the United States 2 Growth Trajectories of Japan, the People’s Republic of China, and the Republic of Korea 8 3 Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse 9 of the Employment Rate, People’s Republic of China 4 Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse 10 of the Employment Rate, Republic of Korea 5 Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse 11 of the Employment Rate, Japan 6 Cointegrating Relationship for the People’s Republic of China 17 7 Test of Beta Equal to the ‘Known Beta’, People’s Republic of China 18 8 One-Step Prediction Test for the Concentrated Model, People's Republic of China 18 9 Test of Coefficient Constancy, People’s Republic of China 19 10 Cointegrating Relation for the Republic of Korea 23 11 Test of Beta Equal to the ‘Known Beta’, Republic of Korea 24 12 One-Step Prediction Test for the Concentrated Model, Republic of Korea 24 13 Test of Coefficient Constancy, Republic of Korea 25 14 Second Cointegrating Relation for Japan 28 15 Test of Beta Equal to the ‘Known Beta’, Japan 28 16 One-Step Prediction Test for the Concentrated Model, Japan 29 17 Test of Coefficient Constancy, Japan 30 A1 Time Series of Capital and Employment 35 A2 First Cointegration Relation, Japan 36 A3 Coefficient Stability for the First Cointegration Relation, Japan 37 ABSTRACT We analyze and compare the patterns of economic growth and development in the Japan, the People’s Republic of China, and the Republic of Korea in the postwar period. The geographical proximity and cultural affinity between the three countries, as well as the key role of the development state in the economies, suggest that an analytical comparison would be a meaningful and valuable exercise. Furthermore, Japan and the Republic of Korea are two of the few economies that have jumped from middle income to high income in a short period and thus offer potentially valuable lessons for the PRC. We use Cobb–Douglas production functions to assess the long-run equilibrium relationships between per capita gross domestic product, capital, and labor by means of cointegrated vector autoregressive models. We show that such equilibrium relationships cannot be rejected for all three countries, while the evidence is stronger for the PRC and the Republic of Korea than for Japan. Our hypothesis tests show that the estimated Cobb–Douglas production functions display coefficients of capital and employment that sum up to 1 and broken linear trends that can be attributed to structural breaks and (changes in) total factor productivity growth. We observe a striking similarity between the experience in the Republic of Korea and the PRC, which gives some optimism that the PRC may be capable of graduating to high income, like the Republic of Korea. Keywords: aggregate production function, comparative economic growth, economic development, Japan, People’s Republic of China, Republic of Korea JEL codes: E23, O47, O53, O57, P52 I. INTRODUCTION Since the introduction of market reforms in 1978, decades of world-topping economic growth have transformed the People’s Republic of China (PRC) into the world’s second-biggest economy and an upper-middle-income economy. The PRC’s remarkable economic transformation, triggered by a systemic shift from a centrally planned economy to a more market-oriented economy, may indeed be the most significant development in the global economic landscape since the Second World War. However, since the global financial crisis of 2008–2009, the PRC’s growth has slowed down visibly, although it continues to grow at a healthy pace. While the slowdown is partly due to a less benign external environment, it is largely due to structural factors, such as rebalancing toward domestic demand and consumption, rapid income convergence toward high-income countries, population aging, and tertiarization. The PRC is already an upper-middle-income country with an income level at which growth typically slows down (see, for example, Eichengreen, Park, and Shin 2012; 2014). Therefore, to some extent, the slowdown is a necessary transition to a more balanced and sustainable growth paradigm, not least against the many imbalances that built up during the high-growth decades.1 At the same time, there is no guarantee that the PRC’s transition from middle income to high income will be as smooth and fast as its transition from low income to middle income. In fact, economic theory suggests that sustaining rapid growth will be difficult because marginal returns to capital eventually decline as an economy grows richer and acquires a larger stock of capital. The gains from shifting workers from low-productivity agriculture to higher-productivity manufacturing also eventually decline. Furthermore, as countries approach the global technology frontier, they must begin to develop new technology on their own instead of relying exclusively on importing advanced technology from abroad. Generally, the essence of economic growth shifts from input accumulation— that is, deploying more capital, labor, and other inputs—to total factor productivity growth—that is, using all those inputs more efficiently. Empirically, a large number of middle-income countries have failed to graduate to high-income status in a reasonable period. This well-known stylized fact has given rise to the concept of the middleincome trap. Of 101 middle-income countries in 1960, only 13 proceeded to high-income status by 2008.2 Will the PRC be able to follow in their footsteps? Of the 13 economies mentioned, only a few appear to be comparable with the PRC. Albert, Jude, and Rebillard (2015) point out that only Japan; the Republic of Korea; Taipei,China; and Israel followed a growth strategy similar to that of the PRC: export-led growth paired with strong investment. Of special interest and relevance to the PRC is the experience of Japan and the Republic of Korea, which are relatively large countries. Although both economies are nowhere near as large as the PRC, they are much larger than Singapore and Hong Kong, China and substantially larger than Taipei,China. The central objective of our paper is to assess empirically the PRC’s prospects for transcending the middle-income range by looking in the rearview mirror and comparing the PRC’s past experiences with those of Japan and the Republic of Korea. Accordingly, we analyze and compare the economic growth and structural transformation trajectory of the PRC, the Republic of Korea, and Japan on the macroeconomic level by estimating Cobb–Douglas production functions in a multivariate 1 See Wagner (2017) and Maliszewski and Zhang (2015). 2 See World Bank (2013). 8 ADB Economics Working Paper Series No. 622 Figure 2: Growth Trajectories of Japan, the People’s Republic of China, and the Republic of Korea GDP = gross domestic product, PRC = People’s Republic of China. Note: The x-axis is the logarithms of capital intensity (K/L), and the y-axis is the logarithms of per capita GDP (Y/N). Source: Authors’ calculation with Penn World Table version 9.0 data. A more detailed view of the individual country time series is presented in Figures 3–5. The variables Y_P_CHN, K_EMP_CHN, and G_CHN are the logarithms of the PRC’s per capita GDP, the logarithms of the capital–employment ratio, and the logarithms of the inverse of the employment rate in the PRC. For each variable, we present the level values (continuous lines; left axis) and the first differences (dashed lines; right axis). In the following, we refer to the respective country with the variable endings CHN for the PRC, KOR for the Republic of Korea, and JP for Japan. Figure 3 shows that the PRC’s per capita growth was comparably volatile before 1979. The same is true for the growth in the capital–employment ratio, which became increasingly strong over the sample period, with the exception of a strong slump around the end of the 1980s.11 The inverse of the employment rate decreased over the whole sample period, indicating that the overall employment in the population increased. 11 This is also in line with the assessment by Chow and Lin (2002) in section II. Republic of Korea; 1953 1980 1984 1997 Republic of Korea; 2014 Japan; 1950 1956 1962 1973 1985 1990 1997 Japan; 2014 PRC; 1952 1970 1976 1992 1994 1997 2008 2011 PRC; 2014 6.5 7.0 7.5 8.0 8.5 9.0 9.5 10.0 10.5 11.0 7.5 8.5 9.5 10.5 11.5 12.5 Trajectories to High Income 9 Figure 3: Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse of the Employment Rate, People’s Republic of China Note: Values are logarithms of levels and their first differences. Sources: Penn World Table version 9.0; Authors’ calculations. –0.4 –0.3 –0.2 –0.1 0.0 0.1 0.2 0 1 2 3 4 5 6 7 8 9 10 1952 1954 1956 1958 1960 1962 1964 1966 1968 1970 1972 1974 1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 Y_P_CHN d_Y_P_CHN 0.00 0.02 0.04 0.06 0.08 0.10 0.12 0.14 0 2 4 6 8 10 12 1952 1954 1956 1958 1960 1962 1964 1966 1968 1970 1972 1974 1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 K_EMP_CHN d_K_EMP_CHN –0.06 –0.05 –0.04 –0.03 –0.02 –0.01 0.00 0.01 0.02 0 0.2 0.4 0.6 0.8 1.0 1.2 1952 1955 1958 1961 1964 1967 1970 1973 1976 1979 1982 1985 1988 1991 1994 1997 2000 2003 2006 2009 2012 G_CHN d_G_CHN 10 ADB Economics Working Paper Series No. 622 Figure 4 displays the developments in the Republic of Korea. The per capita GDP growth rates peaked around 1980 and then decreased. Important recessions happened in 1980 and during the Asian Crisis. Altogether, the per capita GDP growth rates display a humped-shaped pattern. The capital– employment ratio rose strongly in the 1960s (the fast-growth phase), and the growth rates were very high until the 1980s, when they experienced a strong cutback and decreased further throughout the Asian Crisis.12 The inverse of the employment rate shows the same behavior as in the PRC. Figure 4: Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse of the Employment Rate, Republic of Korea Note: Values are logarithms of levels and their first differences. Sources: Penn World Table version 9.0; Authors’ calculations. 12 For separate time series of capital and employment for Japan, the PRC, and the Republic of Korea, we refer the interested reader to Figure A1 in the Appendix. –0.02 0.00 0.02 0.04 0.06 0.08 0.10 0.12 0.14 0.16 0 2 4 6 8 10 12 14 1953 1956 1959 1962 1965 1968 1971 1974 1977 1980 1983 1986 1989 1992 1995 1998 2001 2004 2007 2010 2013 K_EMP_KOR d_K_EMP_KOR –0.08 –0.06 –0.04 –0.02 0.00 0.02 0.04 0.06 0.08 0.10 0.12 0.14 0 2 4 6 8 10 12 1953 1955 1957 1959 1961 1963 1965 1967 1969 1971 1973 1975 1977 1979 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 Y_P_KOR d_Y_P_KOR –0.06 –0.04 –0.02 0.00 0.02 0.04 0.06 0.08 0 0.2 0.4 0.6 0.8 1.0 1.2 1.4 1953 1956 1959 1962 1965 1968 1971 1974 1977 1980 1983 1986 1989 1992 1995 1998 2001 2004 2007 2010 G_KOR d_G_KOR 2013 Trajectories to High Income 11 In Japan (see Figure 5), we observe a more or less stable decrease in the per capita growth rates over the whole sample period. Higher per capita growth rates corresponded to fast increases in capital intensity until the oil crisis in 1973. Afterwards, the capital–employment ratio decreased, although we observe a short-lived boom during the second half of the 1980s. The inverse of the employment rate rose until the oil crisis and then seems to have followed the Japanese business cycle. Figure A1 in the Appendix shows the separate time series for capital and employment for the three countries. Figure 5: Gross Domestic Product per Capita, Capital–Employment Ratio, and Inverse of the Employment Rate, Japan Note: Values are logarithms of levels and their first differences. Sources: Penn World Table version 9.0; Authors’ calculations. -0.08 -0.06 -0.04 -0.02 0.00 0.02 0.04 0.06 0.08 0.1 0.12 0.14 0 2 4 6 8 10 12 1950 1953 1956 1959 1962 1965 1968 1971 1974 1977 1980 1983 1986 1989 1992 1995 1998 2001 2004 2007 2010 2013 Y_P_JP d_Y_P_JP 0 0.02 0.04 0.06 0.08 0.10 0.12 0.14 0.16 0.18 0 2 4 6 8 10 12 14 1950 1953 1956 1959 1962 1965 1968 1971 1974 1977 1980 1983 1986 1989 1992 1995 1998 2001 2004 2007 2010 2013 K_EMP_JP d_K_EMP_JP –0.025 –0.020 –0.015 –0.010 –0.005 0.000 0.005 0.010 0.015 0.020 0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 1950 1953 1956 1959 1962 1965 1968 1971 1974 1977 1980 1983 1986 1989 1992 1995 1998 2001 2004 2007 2010 2013 G_JP d_G_JP 12 ADB Economics Working Paper Series No. 622 C. Univariate Properties of the Data For the unit root tests, we analyze the period from 1956 to 2014. In the previous section, we saw that the growth rates decreased or increased over time. We thus assume that the time series under consideration present deterministic trends and hence test especially for unit roots of the most general form, allowing for a constant and a deterministic trend. According to the unit root tests, the per capita GDP can be assumed to be integrated of order one (I(1)) for all three countries. The same is valid for the employment time series. The tests lead to less convincing results for the capital and capital– employment ratio series. We conclude that the levels of the time series are at least integrated of order one (I(1)). The capital and capital–employment ratio time series may even be I(2).13 The I(2) properties of the capital–employment ratio time series could require an I(2) analysis. However, Juselius (2006) points out that the behavior of an I(2) trend can be mimicked with an I(1) stochastic trend around a broken linear deterministic trend. Thus, it is possible to avoid an I(2) analysis by including an appropriate number of deterministic (linear, broken linear, or quadratic) trends in the data. Our estimations will rely on the variables in levels. This is the procedure that we choose for this section. To obtain a better understanding of where deterministic components could be fitted into our models, we analyze the time series for potential structural breaks in the next section. D. Structural Break Tests From a visual inspection of the time series growth rates, we can assume the occurrence of some structural breaks in the time series. These could also contribute to the results of the unit root tests. To obtain further information about the occurrence of structural breaks in the data, we apply the Bai and Perron (2003) structural break tests, which allow us to obtain a number of interesting details about whether, when, and in which way structural breaks are present in the data. We assume that the time series values evolve in the following way: 𝑥=𝛼 +𝛼 𝑥 +𝜀  (6) If 𝛼 is equal to 1, this process characterizes a random walk plus drift, which can be assumed to be the data-generating process of many macroeconomic time series. The test identifies two breaks. The results for the three countries are displayed in Table A1. In the PRC, structural breaks are first present around the high tide of the Cultural Revolution (1966–1968). The second period during which breaks are found is the end of the Cultural Revolution in 1976 and the beginning of the first reform period (1979). Finally, the breaks in the early 1990s could be associated with Deng Xiaoping’s Southern Tour (in 1992). For the Republic of Korea, we observe structural breaks during the high-growth phase from the mid-1960s to the mid-1970s. The beginning of the stabilization phase in 1980 appears to have had permanent effects on employment and output. The final period of structural breaks is related to the Asian Crisis. For Japan, a relatively clear point to take away from the tests is that we observe structural breaks in the early 1970s. These are most likely related to the end of the fast-growth phase in 1972 and the beginning of the oil shock phase in 1973. The second break seems to occur toward the end of the 13 The unit root tests are not displayed but are available from the authors on request. Trajectories to High Income 13 bubble economy phase and at the beginning of the lost decade in the late 1980s. The employed persons’ time series indicates instability during the Asian Crisis in 1997 but could also point toward the lost decade. The tests provide some evidence that the Japanese economy began to stagnate from about the late 1990s onward, as the coefficient of 𝛼 is well below 1. E. Econometric Framework To determine whether the respective countries were following comparable growth paths toward high income, we use the concept of cointegration. Cointegration appears to be especially appropriate for our research question, as the notion of cointegration states that, between two or more nonstationary time series, a linear combination exists that generates residuals that are stationary (see Engle and Granger 1987). The error correction representation of the cointegrated vector autoregressive (CVAR) model in the multivariate case is ∆𝑋=Π𝑋  +∑Γ∆𝑋 +   𝜙𝐷+𝜀  (7) 𝑋 is a vector that contains the variables included in the model. 𝐷 is a vector of the deterministic components of the model. The Γ matrices contain the short-run information of the model, while Π contains the information about long-run relationships (which we are especially interested in) and can be rewritten as a vector product of α∙β'. Here β' comprises the long-run information, while α contains the information on how and how quickly deviations from the long-run relations are corrected. Cointegration emerges when two or several nonstationary time series are driven by the same persistent shocks (Juselius 2006). In our case, we assume that the shocks driving our cointegrating relationships derive from economic policy measures implemented in the corresponding reform periods. Long-run relations can then be interpreted as economic steady-state relations. Extraordinary events can lead to outliers violating the normality assumption, as they lead to excess skewness and kurtosis (see Juselius 2018). Some of these problems can be resolved by including intervention dummies to account for significant political or institutional events. Less feasible seems to be the possibility to split the sample into more homogeneous periods, as only yearly data for our variables are available. A subsample analysis would hence suffer from problems associated with small samples. Even if the residuals of the VAR pass the misspecification tests sufficiently well, this does not rule out the possibility that the model suffers from parameter nonconstancy. For this reason, Juselius (2006) provides a description of several tests to assess the parameter constancy of the CVAR model. In the next section, we will discuss three selected tests for each model that we estimate to identify any remaining signs of parameter instability. First, we are especially interested in the stability of the long-run relationships, that is, the stability of 𝛽 󰆹. 𝛽 󰆹 can be seen as the average of the related coefficients over the sample period. It could be assumed that structural breaks induced by reforms affected not only the TFP growth but also the output elasticity of inputs. The optimal capital intensity could hence have changed (see Stijepic and Wagner 2011). Theoretically this is also proved by the work of Acemoglu (2003). To test this, we perform a test of the ‘known beta’. This test checks for the consistency of 𝛽 󰆹, that is, the stability of the long-run relations. The basic idea is that the model is estimated for a subsample period 1 to 𝑇, with 𝑇<𝑇, and then the recursive sample is extended until 𝑇 is reached (see Juselius 2006). Second, we apply recursive calculated prediction tests for the long-run relation. If the test value is above 1, the model is not able to predict the observation for this period within the 95% 14 ADB Economics Working Paper Series No. 622 confidence bands. This can be a helpful test to diagnose systematic predictive failure of the model (Juselius 2006). Third, we can test the stability of the estimated coefficients over a sample period. This can give us additional insights into the stability of the long-run relations (Juselius 2006). F. Estimation The variable set that we use is relatively small. It is difficult to obtain data in sufficient quality that reach back far enough. Multivariate cointegration analysis allows for multiple equilibrium relationships. Hence, with n variables in the system, up to n-1 cointegrating relations are possible. (Over-) identification of these relationships requires a sufficient amount of theoretical assumptions on what the concrete cointegrating relationships are. With a rising number of variables, it becomes more and more difficult to identify the ‘true’ structure of the data. In any case, cointegrating relationships that have already been identified should remain stable when additional variables are added. Hence, what we detect in the following should remain present in a richer information environment. As mentioned in the previous section, we assume a simple Cobb–Douglas production function as the starting point of our analysis. Our vectors 𝑋 for the PRC, the Republic of Korea, and Japan are 𝑋_ =(𝑌_𝑃_𝑗 ,𝐾_𝐸𝑀𝑃_𝑗,𝐺_𝑗 , 𝐷_𝑗) (8) for each country j. As mentioned earlier 𝑌_𝑃_𝑗 stands for the respective countries per capita GDP, 𝐾_𝐸𝑀𝑃_𝑗 is the capital–employment ratio, 𝐺_𝑗 refers to the inverse of the employment rate. All variables are in logarithms. 𝐷_𝑗 captures deterministic components like dummy variables and (broken) linear trends. We start by specifying the model for the PRC. To test for cointegration, we first have to obtain a well-specified VAR model of the data. As TFP is usually captured by the residuals of the equation, we take into consideration the need to capture technological progress and other variables that are not accounted for by the introduction of deterministic trends. Juselius (2006) generally proposes two approaches to identifying cointegrating relationships: the general-to-specific and the specific-to- general approaches. As we face a relative lack of restrictions and theories, we find it appropriate to start with a narrow model like that motivated by Chow (2015). Imposing restrictions on the model is crucial to decide how well the data fit our theoretical assumptions. We hence rearrange our baseline theoretical model to obtain a representation that allows us to impose overidentifying restrictions. Chow (2015) estimates several specifications of his production function. The most specific is one with a fixed capital–labor ratio (𝑘/𝑙), which corresponds to capital intensity. 1. Analysis for the People’s Republic of China Lag Length Selection and Diagnostic Testing of the Unrestricted Vector Autoregressive As we have yearly data, a lag length of order 1 appears to be appropriate, as it not very likely that information further back than 1 year is included in the investment and employment decisions. Lag reduction tests support this assumption. A lag length of 1 is superior to lag lengths of 2 or more. As the test statistics for the CVAR rely on the assumption of Gaussian residuals, deviation from the normality assumptions may distort the results. We hence must establish the necessary residual properties before we test for possible long-run relationships. From sections III.B to III.D, we already have a certain idea Trajectories to High Income 15 about possible outliers, trends, or (structural) breakpoints in the data. For the PRC, we choose the sample period of 1969–2014. This excludes the ‘abnormal’ years of the Cultural Revolution (see Figure 2 in section II). The Cultural Revolution was officially announced to be complete in 1969. We thus only miss out four values included by Chow (2015), whose sample is 1952–2012, omitting (1958–1969).14 From our theoretical hypotheses in (7), we assume that we will have to include a deterministic trend in the model that allows for economic growth even if the input factors remain constant. We will, however, test the validity of this assumption. Chow (2015) argues that such a trend would catch increases in TFP. Including a deterministic trend in the CVAR model is also the most general specification that one can choose. Hence, we include a deterministic trend over the whole sample period, additionally allowing for a break in this trend that mirrors a structural break induced by reforms or political events. Chow (1993) finds no evidence of TFP prior to 1980 and thus includes a broken linear trend beginning with t = 1 in 1979 (Chow 2015). With our data, a break in the deterministic trend appears in 1976 (often seen as the de facto end of the Cultural Revolution) such that t = 1 in 1977 seems to be more in line with the data. We additionally include intervention dummies for 1976, 1989, 1990, and 1991. We also include a shift dummy for 1990. With this specification, we obtain a relatively well-specified VAR model, as can be seen from the residual diagnostic tests in Table 2 as there are no signs of autocorrelation, and heteroscedasticity. Skewness and kurtosis are also in line with the assumption of normally distributed residuals Table 2: Residual Analysis—Diagnostic Testing of the Unrestricted Vector Autoregression (1) Model, People’s Republic of China Multivariate test Residual autocorrelation: LM (1): ChiSqr(4) = 5.742 [0.219] LM (2): ChiSqr(4) = 0.879 [0.970] Test for ARCH: LM (1): ChiSqr(9) = 11.875 [0.220] LM (2): ChiSqr(18) = 16.430 [0.563] Univariate tests ARCH(1) Normality Skewness Kurtosis Y_P_CHN 0.391 [0.532] 0.840 [0.657] 0.297 2.685 K_EMP_CHN 1.277 [0.258] 0.856 [0.652] –0.160 3.050 ARCH = autoregressive conditional heteroskedasticity, VAR = vector autoregression. Note: p-values in brackets. Source: Authors’ calculations. Rank Determination and Testing Restrictions on the Cointegrated Vector Autoregression A likelihood-ratio (LR) test of long-run exclusion indicates that 𝑔 is not part of the equilibrium relationship and hence we exclude it from the information set. Any external influences that are not accounted for by the variables in the system are caught by either the deterministic components or the residuals. The trace tests propose one cointegrating relationship, which we would also expect from the 14 Using a longer sample including earlier years leads to more dummy variables in our model and to a smaller, though significant, coefficient of the capital–employment ratio. 16 ADB Economics Working Paper Series No. 622 theory. We thus restrict the rank to be equal to 1. This gives the following just-identified model (see Table 3). At this stage of our identification of the long-run relations, we test whether the data is in line with our theoretical hypothesis, that is, the long-run relation between per capita GDP and the capital–employment ratio should be positive, and deterministic trends which account for technological change should also have a positive relation with per capita GDP growth. Table 3 displays the stationary relationship. The first line gives the stationary long-run relationship. Hence, to see how the capital– employment ratio and the deterministic trends are connected with per capita GDP in the PRC, we have to invert the signs of the former. All variables then would have the expected (positive) signs. Besides the long-run relation in the first line, Table 3 (as well as the following tables) also presents the short-run adjustment coefficients in the second row. We refrain from interpreting these in the following, but nonetheless present them for the sake of completeness. Table 3: The Just-Identified Long-Run Cointegration Relations for r = 1, People’s Republic of China Y_P_CHN K_EMP_CHN T(1976:01) Trend β  1.000 –0.453 –0.033 0.013 (.NA) ( –7.189) ( –3.456) (1.445) α   –0.075 0.114 ( –1.494) (8.932) Source: Authors’ calculations. All the variables have the expected sign, and the GDP per capita and the capital–employment ratio commove in a positive long-run relationship. TFP growth, proxied by the broken linear trend, has a positive effect on output and a significant coefficient of 0.033. In the next step, we relax the implicitly imposed restriction that the coefficients of capital and employment sum up to 1 by including capital and employment separately. By including the same dummy variables, we obtain a rather well-specified model without autocorrelation and mild skewness. The trace test indicates one cointegrating relationship. We continue to assume that the rank of Π is equal to 1, which means that there is one single equilibrium relationship among the variables. Again, the trend is insignificant, as displayed in Table 3. We thus exclude the deterministic trend by imposing a 0 on the corresponding coefficient. Additionally, we explicitly test the coefficients of capital and labor to be of the same size. In this way, we impose more restrictions on the data and may be able to obtain overidentified cointegrating relationships. This overidentified cointegrating relation is displayed in Table 4. Table 4: The Overidentified Long-Run Cointegration Relation for r = 1, People’s Republic of China Y_P_CHN K_CHN EMP_CHN T(1976:01) Trend β  1.000 –0.417 0.417 –0.022 0.000 (.NA) ( –7.278) (7.278) ( –4.349 (.NA) α   –0.075 0.104 –0.031 ( –1.534) (8.079) ( –5.951) Source: Authors’ calculations. Trajectories to High Income 17 The overidentifying restrictions are accepted with a p-value of 0.135 and a 𝜒(2) of 4.010.15 This shows that the restrictions we have imposed on the data are not rejected hence the PRC’s growth path in our sample can be represented by a long-run equilibrium relationship of the form: 𝑌_𝑃− 0.417 ∙ (𝐾−𝐸𝑀𝑃 ) − 0.022 ∙ 𝑡𝑟𝑒𝑛𝑑76 ~ 𝐼(0) (9) Our empirical results show that the coefficient of capital intensity, given by the difference between capital stock and employed persons in the PRC has a coefficient of 0.42. Additionally, the data are in line with a positive deterministic trend which could be interpreted as 2.2% annual TFP growth. Figure 6 displays the corresponding cointegrating relationships. The residuals of the concentrated model (Figure 6) show that the relationship is stable. Figure 6: Cointegrating Relationship for the People’s Republic of China Source: Authors’ calculations. Tests of Constancy The test of beta constancy is displayed in Figure 7. As we can see from Table A1, this is a very challenging test, as 1992 (the beginning of the subsample for the test in Figure 7) is actually an important breakpoint in the PRC data. The forward recursive calculated test shows that the model performs fairly well. After the assumed breakpoint in 1992 (Deng Xiaoping’s Southern Tour of 1992), there is some instability, but the test statistic approaches its critical value of 1 relatively quickly. The Asian Crisis of 1997 and the Global Financial Crisis of 2007–2009 seem to have introduced some instability, but altogether we can assume that the coefficients remain stable after the Asian Crisis. 15 Juselius (2006) suggests applying the small-sample Bartlett correction in moderately sized samples of 50–70 observations, which applies in our case (45 observations). The noncorrected values are p = 0.063 and a chi-square (2) of 5.537 (correction factor: 1.381). 0.15 0.10 0.05 0.00 –0.05 –0.10 –0.15 1970 1972 1974 1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 24 ADB Economics Working Paper Series No. 622 Figure 11: Test of Beta Equal to the ‘Known Beta’, Republic of Korea Source: Authors’ representation. Figure 12: One-Step Prediction Test for the Concentrated Model, Republic of Korea Source: Authors’ calculations. 1981 1.00 0.75 0.25 0.50 0.0 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 5% C.V. (9.49 = Index) X = R1 9 8 7 6 5 4 3 2 1 0 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 X(t) = R1(t) Trajectories to High Income 25 The coefficients displayed in Figure 13 of the long-run relations in the case of the Republic of Korea remained relatively stable over the subsample period. There was some instability from around 1987 until 1990. Figure 13: Test of Coefficient Constanc y , Republic of Korea Source: Authors’ calculations. Beta 1 (R1-model) 2.00 Y_P_KOR = 1 1.75 1.50 1.25 1.00 0.75 0.50 0.25 0.00 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 –0.56 –0.59 –0.60 –0.62 –0.64 –0.66 –0.68 –0.70 –0.72 K_KOR –0.016 –0.019 –0.020 –0.022 –0.024 –0.026 –0.029 –0.030 –0.032 T(1980:01) 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 1.00 0.75 0.50 0.25 0.00 –0.25 –0.50 –0.75 –1.00 TREND = -0 1981 1983 1985 1987 1989 1991 1993 1995 1997 1999 2001 2003 2005 2007 2009 2011 2013 0.72 0.70 0.68 0.66 0.64 0.62 0.60 0.59 0.56 EMP_KOR 26 ADB Economics Working Paper Series No. 622 3. Analysis for Japan Lag Length Selection and Diagnostic Testing of the Unrestricted Vector Autoregression For Japan, the lag reduction tests point towards a lag order of 1. Additionally to the deterministic trend, we include one breakpoint (t = 1 in 1973) to account for changes in the light of the oil crisis. Additionally, we include dummies for 1973 and 1997 (shift dummies) as well as an intervention dummy for 2009. We thus obtain a fairly well-specified model, as displayed in Table 9. Table 9: Residual Analysis—Diagnostic Testing of the Unrestricted Vector Autoregression (1) Model, Japan ARCH = autoregressive conditional heteroskedasticity. Note: p-values in brackets. Source: Authors’ calculations. Rank Determination and Testing Restrictions on the Cointegration Vector Autoregression The rank test indicates one or two cointegrating relationships (see Table 10). Additionally, this time, the test of long-run exclusion does not suggest that 𝑔 is not part of the cointegrating relationships. We assume a rank of 1. Table 10: Likelihood-Ratio Trace Test for the Unrestricted Vector Autoregression (1) Model, Japan R p-r Eigenvalue Trace 95% crit. value p-value 3 0 0.828 143.722 58.737 0.000 2 1 0.431 41.084 37.115 0.015 1 2 0.144 8.941 18.800 0.631 Source: Authors’ calculations. Multivariate test Residual autocorrelation: LM (1): ChiSqr(9) = 31.703 [0.000] LM (2): ChiSqr(9) = 20.288 [0.016] Test for ARCH: LM (1): ChiSqr(36) = 52.761 [0.035] LM (2): ChiSqr(72) = 84.896 [0.142] Univariate tests ARCH(1) Normality Skewness Kurtosis Y_P_JP 1.583 [0.208] 1.007 –0.048 3.157 K_EMP_JP 0.245 [0.621] [0.604] –0.021 3.173 DG_JP 0.127 [0.721] 1.057 –0.261 3.107 Trajectories to High Income 27 This model gives the long-run relationship displayed in Table 11. Table 11: The Just-Identified Long-Run Cointegration Relations for r = 1, Japan Y_P_JP K_EMP_JP G_JP T(1972:01) Trend β  1.000 –0.283 0.829 0.037 –0.046 (.NA) ( –17.209) (4.245) (22.066) ( –23.745) α   –0.004 0.422 0.044 ( –0.093) (15.554) (2.694) Source: Authors’ calculations. The capital–employment coefficient is comparably small. The trend points to overall TFP growth over the whole sample. The break in 1973 sets off most of it. Afterward, TFP growth would only be as high as 0.08 on average. We also note that 𝑔 is significant and very close to the theoretical value of 1 that we derived in section III.A. In the model including capital and employment separately, the inclusion of 𝑔 is not rejected as well. Here, the trace test indicates two cointegrating relationships, and we try to impose the theoretical relation developed in (7) on the second relation. This gives the following overidentified model (see Table 12). Table 12: The Overidentified Long-Run Cointegration Relations for r = 1, Japan Y_P_JP K_JP EMP_JP G_JP T(1972:01) Trend β  0.000 –0.048 1.000 0.832 0.009 –0.008 (.NA) ( –22.433) (.NA) (29.930) (40.292) ( –27.272) β  1.000 –0.264 0.264 1.000 0.044 –0.051 (.NA) ( –23.069) (23.069) (.NA) (36.361) ( –33.139) Source: Authors’ calculations. These restrictions are accepted with a p-value of 0.493 and a CHISQR (1) of 0.493. The second cointegrating relationship is our estimate for the production function. 𝑌_𝑃 − 0.264 ∙ (𝐾−𝐸𝑀𝑃 ) − 1.000 ∙ 𝐺− 0.051 ∙ 𝑡𝑟𝑒𝑛𝑑 + 0.044 ∙ 𝑡𝑟𝑒𝑛𝑑72 ~ 𝐼(0)(11) Figure 14 displays the corresponding cointegration graph.20 20 For the sake of completeness, the first cointegration graph is displayed in Figure A2 in the Appendix. 28 ADB Economics Working Paper Series No. 622 Figure 14: Second Cointegrating Relation for Japan Source: Authors’ calculations. Tests of Constancy The test of beta constancy indicates extended periods of instability of the beta coefficients from the mid-1980s onward. Toward the Asian Crisis, instability decreased considerably but remained significant. The revitalization phase fits with the overall stable beta coefficients (see Figure 15). Figure 15: Test of Beta Equal to the ‘Known Beta’, Japan Source: Authors’ representation. 1957 0.100 0.050 –0.000 –0.050 –0.100 1960 1963 1966 1969 1972 1975 1978 1981 1984 1987 1990 1993 1996 1999 2002 2005 2008 2014 2011 1984 3.0 2.5 0.5 2.0 1.5 1.0 0.0 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 5% C.V. (15.5 = Index) X = R1 Trajectories to High Income 29 The one-step prediction tests in Figure 16 give the impression that the model fails to make trustworthy predictions for the sample period. While the two largest prediction errors are related to the Asian Crisis and the Global Financial Crisis, the model also has increasing difficulties in predicting the developments during the bubble economy phase of 1985–1990 and the beginning of the lost decade. The coefficients displayed in Figure 17 of the long-run relations in the case of Japan remained very unstable over the phase of the bubble economy in the 1980s. They give a clear indication that the Japanese economy was not on its equilibrium path during this period.21 Figure 16: One-Step Prediction Test for the Concentrated Model, Japan Source: Authors’ calculations. 21 The corresponding test for the first cointegrating relation is displayed in Figure A3 in the Appendix. X(t) = R1(t) 3.5 3.0 2.5 2.0 1.5 1.0 0.5 0.0 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 30 ADB Economics Working Paper Series No. 622 Figure 17: Test of Coefficient Constanc y , Japan Source: Authors’ calculations. Beta 2 (R1-model) 2.00 Y_P_JP = 1 1.75 1.50 1.25 1.00 0.75 0.50 0.25 0.00 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 0.2 K_JP 0.1 0.0 –0.1 –0.2 –0.3 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 0.3 EMP_JP 0.2 0.1 0.0 0.1 0.2 2.00 G_JP = 1 1.75 1.50 1.25 1.00 0.75 0.50 0.25 0.00 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 0.060 0.055 0.050 0.045 0.040 0.035 0.030 T(1972:01) 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2006 2008 2010 2012 2014 –0.04 –0.05 –0.06 –0.07 –0.08 –0.09 –0.10 TREND Trajectories to High Income 31 IV. CONCLUSION Sustained rapid growth has transformed the PRC from a low-income economy into a middle-income economy in a remarkably short period of time. The next challenge for the PRC is to graduate from middle income to high income. How well and smoothly the PRC tackles this difficult challenge has sizable ramifications not only for the PRC but, given the PRC’s large and growing footprint on the global economy, for the rest of the world. In this paper, we sought to obtain some clues about the future dynamics of the PRC’s economic growth by looking at Japan’s, the Republic of Korea’s, and the PRC’s past patterns of growth. Accordingly, we analyzed and compared the growth experiences of the three countries, which share many similarities. Section II pointed out these similarities. Perhaps the most significant common denominator was capable bureaucracy and a developmental state that prioritized economic growth (favoring export- and investment-led growth) and played the role of a catalyst in the rapid growth and structural transformation of the three countries. As such, structural policies and reform played a major role. Our analysis and comparison of the patterns of economic growth and structural breaks in Japan, the Republic of Korea, and the PRC yielded a number of interesting findings. The descriptive analysis of section II and the more in-depth econometric analysis of section III both support the view that many features of the PRC’s economic development mirror the earlier experiences of Japan and the Republic of Korea. The GDP growth and capital–labor ratio moved together in a positive long-run relationship in all countries, which can be brought into line with the hypotheses derived from section III.A. An interesting finding is hence that the export- and investment-led growth models that all three countries followed for an extensive period are reconcilable with Cobb–Douglas production functions with broken linear trends. However, there is one interesting and significant difference between the three countries in their economic growth trajectory. Specifically, our analysis indicates that the PRC experienced growth based on TFP gains at a much earlier stage in its development path than the Republic of Korea. If the PRC shifted toward TFP growth at a similar stage to the Republic of Korea, the shift would have occurred around 2011. In fact, the shift in the PRC began as early as the late 1970s. In comparison with Japan, there are no visible signs that TFP growth in the PRC slowed down significantly in our sample period. The broader question that we sought to address through our comparative analysis of the growth experiences of the PRC, Japan, and the Republic of Korea is whether the PRC can replicate especially the Republic of Korea’s success in graduating smoothly from middle income to high income in a relatively short period of time. At a broader level, the balance of evidence from our analysis provides cautious grounds for optimism about the PRC’s prospects for a smooth and quick transition to high income. Above all, the fact that the PRC’s growth has been led by TFP growth in addition to factor accumulation suggests that it may be sustainable. However, to continue its enviable track record of rapid TFP growth, the PRC must forcefully implement structural reforms, such as state-owned enterprise reform and reducing the role of the state in the financial system. Structural challenges, such as population aging, and new risks, such as rising global protectionism, further strengthen the case for such TFP-promoting reforms. APPENDIX Table A1: Structural Break Tests Japan Y_P_JP Lower 95% Upper 95% ≤ 1970 ≤ 1987 > 1987 1970 1969 1972 𝜶𝟏–0.103 0.134 1.956 1987 1976 1988 𝜶𝟐1.021 0.990 0.813 K_JP Lower 95% Upper 95% ≤ 1972 ≤ 1988 > 1988 1972 1971 1973 𝜶𝟏–0.333 0.764 2.035 1988 1987 1989 𝜶𝟐1.035 0.956 0.878 EMP_JP Lower 95% Upper 95% ≤ 1988 ≤ 1997 > 1997 1988 1984 1989 𝜶𝟏0.155 0.843 1.546 1997 1991 1998 𝜶𝟐0.964 0.800 0.629 K_EMP_JP Lower 95% Upper 95% ≤ 1972 ≤ 1988 > 1988 1972 1971 1973 𝜶𝟏–0.303 0.602 1.350 1988 1987 1989 𝜶𝟐1.046 0.954 0.893 G_JP Lower 95% Upper 95% ≤ 1988 ≤ 1997 > 1997 1988 1986 1991 𝜶𝟏0.116 0.171 0.220 1997 1994 1998 𝜶𝟐0.823 0.723 0.674 Republic of Korea Y_P_KOR Lower 95% Upper 95% ≤ 1967 ≤ 1982 > 1982 1967 1965 1969 𝜶𝟏–1.103 0.411 0.487 1982 1981 1986 𝜶𝟐1.151 0.958 0.956 K_KOR Lower 95% Upper 95% ≤ 1969 ≤ 1997 > 1997 1969 1968 1970 𝜶𝟏–1.103 0.411 0.487 1997 1996 2008 𝜶𝟐1.151 0.958 0.956 EMP_KOR Lower 95% Upper 95% ≤ 1975 ≤ 1984 > 1984 1975 1974 1976 𝜶𝟏–0.079 0.639 0.196 1984 1983 1985 𝜶𝟐1.052 0.761 0.942 K_EMP_KOR Lower 95% Upper 95% ≤ 1965 ≤ 1998 > 1998 1965 1964 1966 𝜶𝟏0.639 0.097 0.646 1998 1997 2002 𝜶𝟐0.930 0.999 0.950 K_EMP_KOR Lower 95% Upper 95% ≤ 1964 ≤ 1984 > 1984 1964 1962 1965 𝜶𝟏0.214 0.052 0.051 1984 1977 1989 𝜶𝟐0.838 0.943 0.921 continued on next page 34 Appendix Table A1 continued People’s Republic of China Y_P_CHN Lower 95% Upper 95% ≤ 1968 ≤ 1976 > 1976 1968 1968 1990 𝜶𝟏3.645 3.103 –0.046 1976 1975 1978 𝜶𝟐0.473 0.564 1.013 K_CHN Lower 95% Upper 95% ≤ 1969 ≤ 1992 > 1992 1969 1968 1982 𝜶𝟏0.966 0.090 –0.054 1992 1990 2005 𝜶𝟐0.933 0.999 1.010 EMP_CHN Lower 95% Upper 95% ≤ 1963 ≤ 1979 > 1979 1963 1963 1970 𝜶𝟏1.394 0.121 0.430 1979 1977 1980 𝜶𝟐0.751 0.985 0.936 K_EMP_CHN Lower 95% Upper 95% ≤ 1969 ≤ 1991 > 1991 1969 1968 1978 𝜶𝟏0.989 –0.031 –0.061 1991 1990 1996 𝜶𝟐0.881 1.008 1.015 G_CHN Lower 95% Upper 95% ≤ 1964 ≤ 1982 > 1982 1964 1963 1979 𝜶𝟏0.258 –0.080 0.035 1982 1981 1983 𝜶𝟐0.712 1.081 0.933 Source: Authors’ calculations (sample: 1956–2014). ASIAN DEVELOPMENT BANK ASIAN DEVELOPMENT BANK 6 ADB Avenue, Mandaluyong City 1550 Metro Manila, Philippines www.adb.org Trajectories to High Income Growth Dynamics in Japan, the People’s Republic of China, and the Republic of Korea Japan and the Republic of Korea quickly graduated from middle-income to high-income status and thus offer valuable lessons for the People’s Republic of China (PRC). This paper analyzes the economic growth patterns of the PRC, Japan, and the Republic of Korea in the postwar period. The authors use Cobb–Douglas production functions to assess the long-run equilibrium relationships between per capita gross domestic product, capital, and labor by means of cointegrated vector autoregressive models. The analysis uncovers a striking similarity between the growth experiences of the PRC and the Republic of Korea. About the Asian Development Bank ADB is committed to achieving a prosperous, inclusive, resilient, and sustainable Asia and the Pacific, while sustaining its efforts to eradicate extreme poverty. Established in 1966, it is owned by 68 members —49 from the region. Its main instruments for helping its developing member countries are policy dialogue, loans, equity investments, guarantees, grants, and technical assistance. TRAJECTORIES TO HIGH INCOME GROWTH DYNAMICS IN JAPAN, THE PEOPLE’S REPUBLIC OF CHINA, AND THE REPUBLIC OF KOREA Michael Murach, Helmut Wagner, Jungsuk Kim, and Donghyun Park ADB ECONOMICS WORKING PAPER SERIES NO. 622 October 2020