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The German labour market and the unification shock
Economic Modelling 17 Ž2000. 439᎐454
The German labour market and the unification shock Gerd Hansen Institute of Statistics and Economics, Christian Albrechts Uni¨ ersity of Kiel, Olshausenstrasse 40, D-24098 Kiel, Germany
Abstract The paper analyses labour demand and wage setting in Germany before and after the unification by means of vector error correction models. We conclude that restrictions in traditional labour demand and wage setting equations can be accepted and that there is no evidence of insider wage setting in West Germany either before or after the unification. In contrast, after the unification there are higher effects of the unemployment rate on real wages as before the unification. Finally it is shown that that model can also explain data for the unified Germany after 1990:3. 䊚 2000 Elsevier Science B.V. All rights reserved.
1. Introduction The last recession has pushed the West-German unemployment rate to a new top level of 12%. This is at least partly due to wage policies after the German unification in 1990, when East-German wages adjusted rather quickly towards West-German wages despite the low labour productivity in the East. Wage differentials as well as high unemployment in the East led to high labour inflows into the West. This situation is quite different from that in other European countries. The paper analyses if there are major structural changes in the labour market equations for West Germany and if these equations can also be applied to data of the unified Germany. Economists have given many explanations as to why unemployment is persistent after a new increase during a recession. For example, unemployment persistence can be described as a failure of the dynamic system of demand and supply for labour to adjust towards equilibrium. This failure may be due to the fact E-mail address: hansen@stat ᎐ econ.uni ᎐ kiel.de ŽG. Hansen. 0264-9993r00r$ - see front matter 䊚 2000 Elsevier Science B.V. All rights reserved. PII: S 0 2 6 4 - 9 9 9 3 Ž 9 9 . 0 0 0 4 7 - 4
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G. Hansen r Economic Modelling 17 (2000) 439᎐454
that shocks to supply or demand are persistent or because even temporary shocks cannot be removed Žcompare Andersen and Hylleberg, 1993.. Among models of unemployment persistence, the insider᎐outsider hypothesis is an important candidate to explain the specific situation before and after the German unification. It focuses on wage bargaining of insiders as observed in most European countries. Within wage bargaining, expected shocks to labour demand are taken into account by the insider and do therefore not affect employment, whereas unexpected shocks are fully reflected in changes in employment ŽAlgoskoufis and Manning, 1988; Andersen and Hylleberg, 1993.. The German economic and monetary union on 1 July 1990 was associated with different types of shocks to the West-German labour market: 1. There was a positive shock to labour supply due to high labour inflow from East Germany. 2. There was a short-run positive labour demand shock due to strong preferences in East Germany for West-German products like cars. 3. Finally the high unemployment in the East was partly financed by higher contributions to the social security system in the West and therefore by increasing gross wages in the West. Due to a lack of trade unions and independent employers in East Germany wage negotiations in the East and the West took place between West-German trade unions and employers after the unification. This raises the question of whether the unification shock were well understood in these negotiations. Trade unions could be forced in such a situation to increase wages with respect to the short-run effects by neglecting the long-run perspectives for German unemployment. If this is the case, one would expect a lower impact of unemployment on wage setting after the unification and therefore a more insider-oriented wage policy. This results in a negative structural break in the unemployment parameter. On the other hand, if wage setters take into account the long-run shock to labour supply, the parameter of the unemployment rate should increase. Finally, the third shock actually led to an increase in the wage wedge Žthe difference between gross and net wages.. If wage setting is oriented at net real wages, one would expect no specific effect of this variable on net wages but solely an increase in gross wages and the respective employment effects. To avoid these employment effects, net wages should be reduced at least partly, because of the increasing wage wedge. We will test if there are such effects after the unification and, therefore, if trade unions anticipated the long-run consequences of the increasing labour supply and wage wedge correctly. Therefore the paper analyses: 1. if there is evidence for insider behaviour in the West-German labour market before the unification; 2. how the West-German system reacted to the unification shock; and 3. if there is a stable structure of the labour market for the unified Germany.
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The paper is organised as follows. Section 2 gives a brief review of the structure of the insider᎐outsider model. In Section 3 we discuss the data, the econometric methodology and the empirical results. Section 4 contains the concluding remarks.
2. The insider–outsider hypothesis The standard model of insider wage bargaining assumes that the real product wage is set at the beginning of each period so that full employment of all currently employed is ensured Že.g. Algoskoufis and Manning, 1988.. This means that wt follows from expectations with respect to the labour demand function wEq. Ž1.x as in Eq. Ž2.: l t s ␣ o l ty1 y ␣ 1wt q ␣ 2 z td q 1 t ;
0 - ␣ o - 1,
␣ i ) 0 for
i s 1,2
E Ž l t . s ␣ o l ty1 y ␣ 1wt q ␣ 2 E Ž z td < Ity1 .
Ž1. Ž2.
In Eq. Ž1. l, w and z d denote Žlog. employment, Žlog. gross real wages and exogenous effects on labour demand like those of Žlog. productivity. 1t denotes white noise errors. By solving Eq. Ž2. for wt we get under insider behaviour EŽ l t . s l ty1 Ž ␣ o y 1 . l ty1 q ␣ 2 E Ž z td < Ity1 . . wt s ␣y1 1
Ž3.
Inserting Eq. Ž3. into Eq. Ž1. gives the insider employment function wEq. Ž4.x: l t s l ty1 q ␣ 2 z td y E Ž z td < Ity1 . q 1 t
Ž4.
In addition to the unit root the employment function wEq. Ž4.x contains unanticipated shocks in the exogenous variables z d affecting labour demand. But Eq. Ž4. still implies insider wage setting due to Eq. Ž5.: E Ž l t . s l ty1
Ž5.
If one assumes that 1 t as well as the unanticipated shocks are white noise, employment becomes a random walk, which can be tested against some type of cointegration relation for l. The insider model wEqs. Ž3. and Ž4.x depends on the assumption that trade unions care only about the currently employed. If trade unions will at least try to reduce part of the excess rate of unemployment u y uU , we get Eq. Ž5a. instead of Eq. Ž5., with uU as the natural rate of unemployment Žcompare Nickell and Andrews, 1983, p. 511. E Ž l t . s l ty1 q Ž u ty1 y uUty1 .
with
G 0.
Ž 5a .
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G. Hansen r Economic Modelling 17 (2000) 439᎐454
In this case expected employment in Eqs. Ž2. and Ž5a. must be identical: E Ž l t . s ␣ o l ty1 y ␣ 1wt q ␣ 2 E Ž z td < Ity1 . s l ty1 q Ž u ty1 y uUty1 .
Ž 2a .
Solving this model for real wages gives Eq. Ž3a. and inserting Eq. Ž3a. into Eq. Ž1. gives Eq. Ž4a.: Ž ␣ o y 1 . l ty1 q ␣ 2 E Ž z td < Ity1 . y Ž u ty1 y uUty1 . wt s ␣y1 1
Ž 3a .
l t s l ty1 q Ž u ty1 y uUty1 . q ␣ 2 z td y E Ž z td < Ity1 . .
Ž 4a .
Eq. Ž4a. can be rewritten as Eq. Ž4b. because the rate of unemployment is defined as log labour force lf minus log employment l. l t s Ž 1 y . l ty1 q lf ty1 y uUty1 q ␣ 2 z td y E Ž z td < Ity1 . .
Ž 4b .
Therefore has to be zero for pure insider wage setting. If w in Eq. Ž3a. depends also on unexpected demand shocks and if l depends also on expected demand shocks Žsee Andersen and Hylleberg, 1993., a test is not a test of insider behaviour, because it does not test for effects of unemployment in the wage setting. As long as there is an effect of unemployment on wages, the insider hypothesis has to be rejected. But strong effects of w EŽ z d < Ity1 .x on employment are further evidence against insider behaviour. Assuming rational expectad tions we can test this within a VEC by means of effects of z ty1 on employment. Such shocks to the labour market happened especially after the German unification.
3. Data and empirical results Within this analysis we use seasonally adjusted data for West Germany from 1966:1 to 1994:4 and a data set which contains West-German data up to 1990:2 and data for the unified Germany from 1990:3 to 1995:4 taken from the quarterly national accounts as published by the DIW ŽDeutsches Institut fur ¨ Wirtschaftsforschung, Berlin, 1995.. The first recession after the post-war boom in West Germany was characterised by 1966 with a notable unemployment rate of approximately 2.5%. The years before 1966 are characterised by an unemployment rate less than 1%. Employment Ž l . is measured as number of employees, gross real product wages Ž wp . are taken as gross nominal wage bill devided by the number of employees and the price level of GNP. Wage settings are most often assumed to target net real wages in consumer prices. We therefore include the difference between gross and net real product wages and test this assumption. Net nominal wages are calculated as gross wage bill minus direct taxes and contributions to the social security system per employee. As a measure of output we use log GNP Žgnp. and as a measure of unemployment the unemployment rate Ž u. defined as
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difference between log labour force lf and log employment l. Dickey᎐Fuller and Perron tests Žsee Perron, 1989. for unit roots ᎏ the latter for time series with structural break ᎏ are given in Table A1 in Appendix A. These tests show that all variables except the real wage rates wp and wnp in dataset I ŽWest-German data. are IŽ1., whereas the latter two are trend stationary. In dataset II Ždata for East and West Germany since 1990:3. the Perron-test accepts the null hypothesis of a unit root in all variables. The cointegration analysis is done by means of Johansen’s Ž1988, 1995. maximum-likelihood estimation of a vector-error-correction model ŽVEC. for the variables labour productivity Ž y y l ., gross real product wage per employee Ž wp ., net real consumer wage per employee Ž wnp., the rate of unemployment Ž u. and the number of employees Ž l .. We expect two cointegration relations, namely a labour demand function corresponding to a CES-technology l s  0 q  1 y y  2 wp y  3 t
Ž6.
with time trend as measure of the effects of technical progress and a wage setting equation wnp s  0U q  1U Ž y y l . q  2U l q  3U u.
Ž7.
If technology has constant returns to scale  1 s 1 will hold in Eq. Ž6.. As cointegration relations found in subsets of variables will also be stationary in a larger set of variables, we first analysed the subsets of variables in Eq. Ž6. vs. Eq. Ž7. in order to identify cointegration relations. The results of the subset analysis of Eq. Ž6. are given in Tables A2 in Appendix A. They show that each subset includes one cointegration relation which can be understood as Eq. Ž6. or Eq. Ž7., respectively. For Eq. Ž6. we accept the hypothesis that there are constant returns to scale. After having specified reasonable restrictions for the labour demand and the wage equation, we analyse if these restrictions are also accepted within a VEC of the five variables: y y l, wp, wnp, u and l. Again, we specify a system with unrestricted constant and trend in the cointegration space, but test if we can restrict the constants to zero in addition to the above restrictions. In order to test for breaks in the cointegration parameters due to the German unification, we include some dummies as exogenous variables in the cointegration relation. We assume that these exogenous dummy variables do not change the distribution of the rank statistics. Johansen and Nielsen Ž1993. provide a program ‘Disco’, which computes critical values of the rank statistics if dummy variables are included in the set of differenced variables. In the case of the German unification it seems more reasonable to assume structural breaks in the cointegration relations. By means of some Monte-Carlo experiments we found that exogenous dummy variables work quite well within the Johansen procedure. We analysed three different data sets, namely: 1. West-German data before the unification Ž1966:1᎐1990:2.;
444 G. Hansen r Economic Modelling 17 (2000) 439᎐454
Fig. 1. Data for West Germany Ž1966:1᎐1990:2. and unified Germany Ž1990:3᎐1995:4..
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445
2. West-German data before and after the unification Ž1966:1᎐1994:4.; and 3. West-German data until 1990:2 and data for the unified Germany from 1990:3᎐1995:5. These data are given in Fig. 1 and show the structural break in 1990:3. 3.1. Results for West Germany before the unification We first analysed the subsets of variables in the labour demand and wage functions. Table A2 in Appendix A shows the result for the labour demand function. There is only one cointegration relation in this subset. The coefficients fit well to what they should be in a labour demand function. The output coefficient can be restricted to one so that we can use labour productivity as one variable in what follows. For the subset of variables in the wage equation we also got one cointegration relation with similar coefficients as in Tables 1a and 1b. The results with respect to the cointegration rank are given in Table 1a. We accept that there are two cointegration relations in this set. From Table 1b it follows that the above restrictions are accepted with marginal significance level SL s 0.09. Coefficients  o ,  1 ,  2 Žand ␣ o , ␣ 1 , ␣ 2 . in Table 1b refer to different sets of restrictions. Due to constant returns to scale we include log productivity y y l instead of including y and l separately. The parameters in the restricted vectors  1 are plausible in terms of Eqs. Ž6. and Ž7. and are mainly the same as in the subset models. It has to be recognised that coefficients of wp and t have the opposite sign, because y y l enters the equation instead of l y y as normalised variable. It is shown in Hansen et al. Ž1998. that in finite samples classical 2-tests reject the null too often due to fat tails of the distribution of ˆ. Mittnik et al. Ž1996. emphasise modified critical values c 2 for significance level ␥ according to ln c 2 Ž d.f.. f 1.5725ln 2 Ž d.f.. y ln Ž 1 y ␥ . 0.6317 2 Ž d.f.. 2
y0.0434 Ž ln 2 Ž d.f... , where 2 Žd.f.. is the classical 2-statistic.
Table 1a Test for cointegration rank r
0
1
2
3
4
m U Trace tr U
0.3758 43.36 24.35 105.23 86.96
0.2883 31.28 20.41 61.87 62.61
0.1583 15.86 16.73 30.58 42.20
0.0944 9.13 13.08 14.72 25.47
0.0590 5.60 12.39 5.60 12.39
446
Table 1b Restricted cointegration vectors  and loading coefficients ␣ a yyl
wp
wnp
1.0 y0.123
0.410 y0.610
1
1.0
y0.579 Ž13.79. 0
y1.208 Ž21.96.
2
1.0 y0.293 Ž2.63.
y61.388 Ž8.48. 0
y1.498 1.0 0 1.0 0 1.0
y1.787 0.784 0 1.925 Ž13.85. 0
l
Time
Const.
y0.529 0.416
y0.001 0.000
0.000 0.000
y0.002 Ž7.0. 0
0
y0.035 Ž0.78. 0
0
0 0.957 Ž8.78. 0
1.098 Ž5.25.
0.358 Ž2.63.
␣o
0.131 Ž2.32. y0.009 Žy0.10.
0.230 Ž5.17. 0.146 Ž2.19.
0.350 Ž7.36. y0.059 Žy0.83.
y0.008 Žy1.05. 0.023 Ž1.95.
0.031 Ž2.21. 0.009 Ž0.43.
␣1
0.011 Ž0.18. y0.131 Žy1.57.
0.130 Ž2.69. y0.099 Žy1.50.
y0.038 Žy0.71. y0.379 Žy5.19.
0.011 Ž1.20. 0.025 Ž2.07.
0.003 Ž0.19. y0.042 Žy2.04.
␣2
y0.001 Žy0.89. y0.134 Žy1.11.
y0.001 Žy1.47. y0.189 Žy1.95.
y0.005 Žy5.66. y0.624 Žy6.04.
ARCH Ž6. Norm. Ž10. a
0 0
0.000 Ž0.06. y0.003 Žy0.11.
5.09
5.33
5.95
8.46
4.40
6.66
1.07
0.49
6.35
1.92
SLwLMŽ1.x s 0.94, SLŽnorm. s 0.08.
0
0
2 Žd.f.. SL Ž 2 . wSL Žc 2 .x
9.62 Ž5. 0.09 w0.57x 10.12 Ž7. 0.18 w0.78x
G. Hansen r Economic Modelling 17 (2000) 439᎐454
o
u
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447
Hansen et al. show that this modified test has the correct size. The c 2-statistic and its marginal significance level are given in brackets in Table 1b᎐3b. ARCH, norm, SLwLMŽ1.x and SLŽnorm. are the test-statistics for ARCH effects, normality and the significance level of the LM-test for first order autocorrelation and normality, respectively. Degrees of freedom of these test statistics are given in parentheses. Although the 2-test accepts the restriction that the unemployment rate is weakly exogenous, this restriction changes the parameters of the labour demand function Ž  2 . towards implausible values. 3.2. Results for West Germany including the unification period Our next topic is to analyse if and how this labour market system changes due to the German economic and monetary union on 1 July 1990. First we estimate the model specified so far for the period 1966:1᎐1994:4. We got the following results not reported here: 1. the cointegration rank is now likely to be 1; 2. if we analyse the variables of the labour demand function and impose one cointegration vector as well as constant returns to scale, this is strongly rejected, although the parameters are mainly unchanged; 3. the wage setting restrictions are accepted with marginal significance level SL s 0.06, but the coefficients of employment are now doubled with other coefficients mainly unchanged; and 4. if we still assume two cointegration vectors, the coefficients change tremendously compared to those for 1966:1᎐1990:2. Taken together there is evidence for a change in the model due to the unification. We therefore test if there is a break 1. in the unemployment cointegration parameter by defining a dummy variable d s 0 for 66:1 to 90:2,u for 90:3 to 94:44 2. andror in net wages due to the increasing wage wedge after unification by defining a second dummy variable d1 s 0 for 66:1 to 90:2,wp y wnp for 90:3 to 94:44 These results in Table 2 show that the cointegration rank for the model with dummy variables is again 2. The two cointegration vectors in  1 as well as in  2 give the same parameters as in Table 1b. The dummy variables can be restricted to appear only in the wage equation. With respect to the two dummy variables in  1 there are two opposite effects. Firstly, the coefficient of the unemployment rate is much higher after 1990:2 and therefore unemployment is taken more seriously after the unification. But secondly, the increasing wage wedge dummy has a positive effect on net real wages after 1990:2, which contradicts the expectation that the increasing wage wedge should reduce net real wages in order to avoid partly the negative employment effects of higher gross real wages. But this result
G. Hansen r Economic Modelling 17 (2000) 439᎐454
448
Table 2 Labour market equations with two dummy variables for 1990:3᎐1994:4 r
0
1
2
3
4
Ža. Rank test-statistics Ž5% critical values for trace-test . 0.3767 0.2574 0.1577 0.1257 0.0619 m 52.00 32.74 18.88 14.77 7.03 U 24.35 20.41 16.73 13.08 12.39 Trace 125.42 73.43 40.68 21.80 7.03 tr U 86.96 62.61 42.20 25.47 12.39 Žb. Restricted cointegration vectors  and loading coefficients ␣a yyl wp wnp u l d1 d
1
1.0
2
y1.176 Ž18.70. 1.0 y1.196 Ž20.27.
3
1 y1.196 Ž20.27.
␣2
y0.018 Ž0.41. y0.146 Ž2.03. ARCH 11.51 Ž6.norm. 3.72 Ž10. a
y0.610 Ž10.34. 0 y0.594 Ž10.61. 0
y0.593 Ž10.59. 0
0 1.0 0 1.0
0 1.0
0.092 y0.058 Ž2.44. Ž1.46. y0.082 y0.369 Ž1.37. Ž5.82. 1.75 8.56 0.09 0.53
0
0
1.938 Ž12.42. 0
0.967 Ž8.48. 0
1.953 Ž13.29.
0.974 Ž8.77.
0 1.953 Ž13.20.
0 0.975 Ž8.78.
0.011 y0.006 Ž1.70. Ž0.55. 0.024 y0.046 Ž2.36. Ž2.67. 12.22 5.37 8.24 4.32
2 Žd.f.. SLŽ 2 . wSL Žc 2 .x 0 0 y0.003 0.395 8.64 Ž5. Ž7.5. Ž2.628. 0.12 w0.60x y3.262 28.826 0 1.278 Ž1.60. Ž1.66. Ž2.057. 0 0 y0.003 0 11.0 Ž8. Ž1.5. 0.20 w0.85x 0 1.110 0 0 Ž7.76. 0 0.130 Ž7.65.
0 0
Time
y0.003 Ž7.5. 0
Const.
0 0
0
0
0
0
0
0
0
0
11.2 Ž8. 0.19 w0.84x
SLwLMŽ1.x s 0.91, SLŽnorm. s 0.03.
depends on the multicollinearity of the two dummy variables. The results for  2 and  3 show that one of the dummy variables can be dropped. If we drop the wage gap dummy Ž d1., the negative coefficient of the unemployment rate increases from 1.953 for 1966:1᎐1990:2 to 3.063 for 1990:3᎐1994:4, reflecting the higher weight put on the unemployment problem after the unification. In the same way we get in  3 a negative influence of the wage gap dummy if the unemployment dummy is dropped, reflecting the lower increase in net real wages due to the increasing wage wedge after the unification. This is evidence that wage setters reacted to unification shocks with less support for insiders as before the unification.
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449
3.3. Stable labour-market model for the unified Germany In our last exercise we take West-German data from 1966:1 up to the unification in 1990:2 and data for the unified Germany from 1990:3 to 1995:4 in order to analyse if the long-run relations in this set of data can also be described by means of a few dummy breaks. This data set is characterised by strong structural breaks in output, employment and the labour force in 1990:3 because of a 12% increase in GNP and a 29% increase in the employment Žsee Fig. 1.. After the unification employment in East Germany decreased rather fast and productivity increased, adjusting to the new real wages and a convertible currency. Therefore we tried to model the data by means of breaks in the parameter of labour productivity Ž y y l ., output Ž y ., employment Ž l . or unemployment Ž u.. The most reasonable results are obtained with a break in the output coefficient in the labour demand function Žvariable d2. rejecting the null of constant returns to scale after 1990:2. From the point of economic theory this result seems to be plausible, because by means of the unification the German capital᎐labour ratio is heavily distorted due to: 1. a lower capital᎐labour ratio in East Germany; and 2. the fact that a large part of the East-German capital stock turns out to be idle capital under conditions of a competitive open economy. The results are improved by adding a 0,1-dummy Ž d3. for 1989:4, the period when the wall came down and approximately half a million people left East Germany for the West. For this set of data the trace test gives again two cointegration vectors according to Table 3a. The restricted cointegration vectors of Table 3b give similar cointegration parameters as in the West-German model, except a much smaller employment coefficient in the wage equation. This reduction means that insider power in wage setting is substantially reduced in the labour market for the unified country. Tests for autocorrelation, ARCH-effects and normality accept the null hypothesis, except ARCH-effects in the ‘net wage’ equation.
Table 3a Rank statistics of the labour market-equations for unified Germany Ž5% critical values. r
0
1
2
3
4
m m) Trace tr U
0.3145 43.05 24.35 113.77 86.96
0.2271 29.37 20.41 70.72 62.61
0.1763 22.12 16.73 41.35 42.20
0.0927 11.10 13.08 19.24 25.47
0.0689 8.14 12.39 8.14 12.39
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450
Table 3b Restricted cointegration vectors  and loading coefficients ␣ a yyl
2
1
y1.260 Ž15.2.
␣2
ARCH Ž6. norm. Ž10. a
wp
y0.498 Ž33.2. 0
wnp
0
1
u
l
0
1.647 Ž5.07.
0
0.190 Ž3.7.
0.217 Ž1.5.
0.262 Ž2.1.
y0.149 Ž1.0.
y0.084 Ž4.7.
0.69 Ž2.4.
0.051 Ž2.0.
0.090 Ž4.0.
0.045 Ž1.6.
y0.009 Ž2.7.
0.030 Ž5.5.
4.3 4.1
6.6 0.71
10.53 0.51
8.8 2.9
18.4 0.41
d2
d3
Time
0.005 Ž5.0.
y0.563 Ž2.8.
y0.002 Ž5.0.
0
16.16 Ž1.8.
0
2 Žd.f.. wSL Ž 2 .x 2.71 Ž4. 0.61 w0.80x
SLwLMŽ1.x s 0.27, SLwLMŽ4.x s 0.17, SLŽnorm. s 0.09.
With dummy variables at zero, recursive ML-estimation of a VEC is not possible. Therefore we choose normally distributed dummy variables with mean value of zero and very small standard deviation ŽS.D.s 0.0005.. Fig. 2 in Appendix A shows the test for constancy of the cointegration space. It tests if the identified cointegration parameters ˜ in Table 3b is in the cointegration space in all sub-samples H 0 : ˜ g spŽ  ., s T0 , . . . ,T. The test statistic is asymptotically 2 wŽ p y r . r x distributed with p as dimension of  . The test statistic has been scaled by the 95% quantile of the 2-distribution such that unity corresponds to a 5% significance level. Beta ᎏ Z uses recursive estimates of the short-run parameters and the cointegration parameters, whereas Beta ᎏ R uses fixed short-run parameters from the complete sample. It can be seen that the test for parameter stability of the cointegration parameters, Beta ᎏ R, is accepted. Plots of these cointegration parameters in Fig. 3 show no break at 1990:2, although there seems to be evidence for an opposite break in the productivity and employment coefficient in 1988:3 and 1988:4. We tried to eliminate this break by an additional dummy variable for 1988:3 without success for the overall stability of the model. In total we conclude that the German social and monetary union in 1990:3 did not destroy the long-run labour market equations but transfer them to East Germany. Furthermore, it did reduce the power of insiders in wage setting.
4. Conclusions Within this paper we analysed the West German labour market by means of
G. Hansen r Economic Modelling 17 (2000) 439᎐454
451
Fig. 2. Test of known beta eq. to betaŽ t . wLABOUR MARKETx.
Johansen’s maximum-likelihood method for vector error correction ŽVEC. models. We tested for insider wage setting before and after the German unification in 1990. Our results show that the West German labour market can be described by means of a five-variable VEC with two cointegration relations which satisfy standard labour demand and wage setting restrictions. We found significant negative effect of the rate of unemployment on wages and therefore strong evidence against pure insider behaviour. For the period 1966:1᎐1994:4, which includes the unification, an increasing coefficient of the unemployment rate was found, which means that the shock to the labour force from the unification is well understood in wage bargaining, although the speed of adjustment towards equilibrium may still be too slow. Finally the model has been successfully extended to explain a data set which includes East Germany after the unification. If the model is estimated using data for West Germany until 1990:2 and for the unified Germany from 1990:2 to 1995:4, a single dummy variable for a break in returns to scale reinstalls the previous results.
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452
Fig. 3. Recursive estimates of cointegration parameters for unified Germany.
Acknowledgements Comments by an anonymous referee are gratefully acknowledged.
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Appendix A Table A1. Augmented Dickey᎐Fuller test-statistics ŽADF. and t-statistics ŽPP. a Dataset I
Variable
Variable
Max lag
ADF
l wp y wnp u yyl
5 5 4 4 5 4
0.876 y4.50b y1.09 y4.25b y1.95 y1.66
⌬l ⌬p ⌬y ⌬ wnp ⌬u ⌬Ž y y l .
Max lag
ADF
5 4 3 4 5 3
y3.70 y3.19 y7.39 y7.167 y3.41 y8.78
Dataset II Variable
Max lag
ADF
l wp y wnp u yyl
4 4 4 4 4 4
y2.52 y1.27 y3.23 y1.54 y3.13 y3.46
a Dataset I: West-German data, 1966:1 to 1994:4; Dataset II: Data for West-German from 1966:1 to 1990:2 and for unified Germany 1990:3 to 1995:4. b With trend.
Table A2. Results for the subset of variables in the labour demand equation wEq. Ž6.x R
ma x
Trace
U max
tr U
0 1 2
0.3062 0.1281 0.0491
33.63 12.61 4.63
50.88 17.25 22.95
16.13 12.39 10.56
42.20 25.47 12.39
Table A2b. Labour demand-function cointegration vectors Ž  . and loading coefficients Ž ␣ . wARCH Žp. and norm Ž2. are test-statistics for ARCH-effects and normality, respectivelyx l

␣
ARCH Ž6. norm Ž2.
y
wp
Time
1.0
y0.995 Ž9.48. y1.0
0.514 Ž8.16. 0.514 Ž21.54.
y0.086 Žy4.85.
y0.235 Žy2.91.
y0.292 Žy4.75.
1.0
y8.10 0.16
6.07 1.04
0.0023 Ž5.10. 0.0023 Ž10.2.
2
d.f.
SL
0.00
1
0.96
3.34 0.41
References Algoskoufis, G., Manning, A., 1988. Wage setting and unemployment persistence in Europe, Japan and the USA. Eur. Econ. Rev. 32, 698᎐706. Andersen, T., Hylleberg, S., 1993. Testing for Insider᎐Outsider Effects. Institute of Economics, University of Aarhus, Denmark.
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Hansen, G., Kim, J.R., Mittnik, S., 1998. Testing cointegration coefficients in vector autoregressive error correction models. Economics Letters, 58, 1᎐5. Johansen, S., 1988. Statistical analysis of cointegration vectors. J. Econ. Dynam. Control 12, 231᎐254. Johansen, S., 1995. Likelihood-Based Inference in Cointegrated Vector Autoregressive Models. Oxford. Johansen, S., Nielsen, B., 1993. Manual for the Simulation Program Disco. Institute of Mathematical Statistics, University of Copenhagen. Mittnik, S., Kim, J.-R., Rachev, S.T., 1996. Chi-square-type distributions for heavy-tailed variates. Discussion Paper 94. Institute of Statistics and Econometrics, Christian-Albrechts-University at Kiel. Nickell, S.J., Andrews, M., 1983. Unions, real wages and employment in Britain 1951᎐79. Oxford Econ. Pap. 35 ŽSuppl.., 183᎐206. Perron, P., 1989. The great crash, the oil price shock, and the unit root hypothesis. Econometrica 57, 1361᎐1401.
The German labour market and the unification shock Gerd Hansen Institute of Statistics and Economics, Christian Albrechts Uni¨ ersity of Kiel, Olshausenstrasse 40, D-24098 Kiel, Germany
Abstract The paper analyses labour demand and wage setting in Germany before and after the unification by means of vector error correction models. We conclude that restrictions in traditional labour demand and wage setting equations can be accepted and that there is no evidence of insider wage setting in West Germany either before or after the unification. In contrast, after the unification there are higher effects of the unemployment rate on real wages as before the unification. Finally it is shown that that model can also explain data for the unified Germany after 1990:3. 䊚 2000 Elsevier Science B.V. All rights reserved.
1. Introduction The last recession has pushed the West-German unemployment rate to a new top level of 12%. This is at least partly due to wage policies after the German unification in 1990, when East-German wages adjusted rather quickly towards West-German wages despite the low labour productivity in the East. Wage differentials as well as high unemployment in the East led to high labour inflows into the West. This situation is quite different from that in other European countries. The paper analyses if there are major structural changes in the labour market equations for West Germany and if these equations can also be applied to data of the unified Germany. Economists have given many explanations as to why unemployment is persistent after a new increase during a recession. For example, unemployment persistence can be described as a failure of the dynamic system of demand and supply for labour to adjust towards equilibrium. This failure may be due to the fact E-mail address: hansen@stat ᎐ econ.uni ᎐ kiel.de ŽG. Hansen. 0264-9993r00r$ - see front matter 䊚 2000 Elsevier Science B.V. All rights reserved. PII: S 0 2 6 4 - 9 9 9 3 Ž 9 9 . 0 0 0 4 7 - 4
440
G. Hansen r Economic Modelling 17 (2000) 439᎐454
that shocks to supply or demand are persistent or because even temporary shocks cannot be removed Žcompare Andersen and Hylleberg, 1993.. Among models of unemployment persistence, the insider᎐outsider hypothesis is an important candidate to explain the specific situation before and after the German unification. It focuses on wage bargaining of insiders as observed in most European countries. Within wage bargaining, expected shocks to labour demand are taken into account by the insider and do therefore not affect employment, whereas unexpected shocks are fully reflected in changes in employment ŽAlgoskoufis and Manning, 1988; Andersen and Hylleberg, 1993.. The German economic and monetary union on 1 July 1990 was associated with different types of shocks to the West-German labour market: 1. There was a positive shock to labour supply due to high labour inflow from East Germany. 2. There was a short-run positive labour demand shock due to strong preferences in East Germany for West-German products like cars. 3. Finally the high unemployment in the East was partly financed by higher contributions to the social security system in the West and therefore by increasing gross wages in the West. Due to a lack of trade unions and independent employers in East Germany wage negotiations in the East and the West took place between West-German trade unions and employers after the unification. This raises the question of whether the unification shock were well understood in these negotiations. Trade unions could be forced in such a situation to increase wages with respect to the short-run effects by neglecting the long-run perspectives for German unemployment. If this is the case, one would expect a lower impact of unemployment on wage setting after the unification and therefore a more insider-oriented wage policy. This results in a negative structural break in the unemployment parameter. On the other hand, if wage setters take into account the long-run shock to labour supply, the parameter of the unemployment rate should increase. Finally, the third shock actually led to an increase in the wage wedge Žthe difference between gross and net wages.. If wage setting is oriented at net real wages, one would expect no specific effect of this variable on net wages but solely an increase in gross wages and the respective employment effects. To avoid these employment effects, net wages should be reduced at least partly, because of the increasing wage wedge. We will test if there are such effects after the unification and, therefore, if trade unions anticipated the long-run consequences of the increasing labour supply and wage wedge correctly. Therefore the paper analyses: 1. if there is evidence for insider behaviour in the West-German labour market before the unification; 2. how the West-German system reacted to the unification shock; and 3. if there is a stable structure of the labour market for the unified Germany.
G. Hansen r Economic Modelling 17 (2000) 439᎐454
441
The paper is organised as follows. Section 2 gives a brief review of the structure of the insider᎐outsider model. In Section 3 we discuss the data, the econometric methodology and the empirical results. Section 4 contains the concluding remarks.
2. The insider–outsider hypothesis The standard model of insider wage bargaining assumes that the real product wage is set at the beginning of each period so that full employment of all currently employed is ensured Že.g. Algoskoufis and Manning, 1988.. This means that wt follows from expectations with respect to the labour demand function wEq. Ž1.x as in Eq. Ž2.: l t s ␣ o l ty1 y ␣ 1wt q ␣ 2 z td q 1 t ;
0 - ␣ o - 1,
␣ i ) 0 for
i s 1,2
E Ž l t . s ␣ o l ty1 y ␣ 1wt q ␣ 2 E Ž z td < Ity1 .
Ž1. Ž2.
In Eq. Ž1. l, w and z d denote Žlog. employment, Žlog. gross real wages and exogenous effects on labour demand like those of Žlog. productivity. 1t denotes white noise errors. By solving Eq. Ž2. for wt we get under insider behaviour EŽ l t . s l ty1 Ž ␣ o y 1 . l ty1 q ␣ 2 E Ž z td < Ity1 . . wt s ␣y1 1
Ž3.
Inserting Eq. Ž3. into Eq. Ž1. gives the insider employment function wEq. Ž4.x: l t s l ty1 q ␣ 2 z td y E Ž z td < Ity1 . q 1 t
Ž4.
In addition to the unit root the employment function wEq. Ž4.x contains unanticipated shocks in the exogenous variables z d affecting labour demand. But Eq. Ž4. still implies insider wage setting due to Eq. Ž5.: E Ž l t . s l ty1
Ž5.
If one assumes that 1 t as well as the unanticipated shocks are white noise, employment becomes a random walk, which can be tested against some type of cointegration relation for l. The insider model wEqs. Ž3. and Ž4.x depends on the assumption that trade unions care only about the currently employed. If trade unions will at least try to reduce part of the excess rate of unemployment u y uU , we get Eq. Ž5a. instead of Eq. Ž5., with uU as the natural rate of unemployment Žcompare Nickell and Andrews, 1983, p. 511. E Ž l t . s l ty1 q Ž u ty1 y uUty1 .
with
G 0.
Ž 5a .
442
G. Hansen r Economic Modelling 17 (2000) 439᎐454
In this case expected employment in Eqs. Ž2. and Ž5a. must be identical: E Ž l t . s ␣ o l ty1 y ␣ 1wt q ␣ 2 E Ž z td < Ity1 . s l ty1 q Ž u ty1 y uUty1 .
Ž 2a .
Solving this model for real wages gives Eq. Ž3a. and inserting Eq. Ž3a. into Eq. Ž1. gives Eq. Ž4a.: Ž ␣ o y 1 . l ty1 q ␣ 2 E Ž z td < Ity1 . y Ž u ty1 y uUty1 . wt s ␣y1 1
Ž 3a .
l t s l ty1 q Ž u ty1 y uUty1 . q ␣ 2 z td y E Ž z td < Ity1 . .
Ž 4a .
Eq. Ž4a. can be rewritten as Eq. Ž4b. because the rate of unemployment is defined as log labour force lf minus log employment l. l t s Ž 1 y . l ty1 q lf ty1 y uUty1 q ␣ 2 z td y E Ž z td < Ity1 . .
Ž 4b .
Therefore has to be zero for pure insider wage setting. If w in Eq. Ž3a. depends also on unexpected demand shocks and if l depends also on expected demand shocks Žsee Andersen and Hylleberg, 1993., a test is not a test of insider behaviour, because it does not test for effects of unemployment in the wage setting. As long as there is an effect of unemployment on wages, the insider hypothesis has to be rejected. But strong effects of w EŽ z d < Ity1 .x on employment are further evidence against insider behaviour. Assuming rational expectad tions we can test this within a VEC by means of effects of z ty1 on employment. Such shocks to the labour market happened especially after the German unification.
3. Data and empirical results Within this analysis we use seasonally adjusted data for West Germany from 1966:1 to 1994:4 and a data set which contains West-German data up to 1990:2 and data for the unified Germany from 1990:3 to 1995:4 taken from the quarterly national accounts as published by the DIW ŽDeutsches Institut fur ¨ Wirtschaftsforschung, Berlin, 1995.. The first recession after the post-war boom in West Germany was characterised by 1966 with a notable unemployment rate of approximately 2.5%. The years before 1966 are characterised by an unemployment rate less than 1%. Employment Ž l . is measured as number of employees, gross real product wages Ž wp . are taken as gross nominal wage bill devided by the number of employees and the price level of GNP. Wage settings are most often assumed to target net real wages in consumer prices. We therefore include the difference between gross and net real product wages and test this assumption. Net nominal wages are calculated as gross wage bill minus direct taxes and contributions to the social security system per employee. As a measure of output we use log GNP Žgnp. and as a measure of unemployment the unemployment rate Ž u. defined as
G. Hansen r Economic Modelling 17 (2000) 439᎐454
443
difference between log labour force lf and log employment l. Dickey᎐Fuller and Perron tests Žsee Perron, 1989. for unit roots ᎏ the latter for time series with structural break ᎏ are given in Table A1 in Appendix A. These tests show that all variables except the real wage rates wp and wnp in dataset I ŽWest-German data. are IŽ1., whereas the latter two are trend stationary. In dataset II Ždata for East and West Germany since 1990:3. the Perron-test accepts the null hypothesis of a unit root in all variables. The cointegration analysis is done by means of Johansen’s Ž1988, 1995. maximum-likelihood estimation of a vector-error-correction model ŽVEC. for the variables labour productivity Ž y y l ., gross real product wage per employee Ž wp ., net real consumer wage per employee Ž wnp., the rate of unemployment Ž u. and the number of employees Ž l .. We expect two cointegration relations, namely a labour demand function corresponding to a CES-technology l s  0 q  1 y y  2 wp y  3 t
Ž6.
with time trend as measure of the effects of technical progress and a wage setting equation wnp s  0U q  1U Ž y y l . q  2U l q  3U u.
Ž7.
If technology has constant returns to scale  1 s 1 will hold in Eq. Ž6.. As cointegration relations found in subsets of variables will also be stationary in a larger set of variables, we first analysed the subsets of variables in Eq. Ž6. vs. Eq. Ž7. in order to identify cointegration relations. The results of the subset analysis of Eq. Ž6. are given in Tables A2 in Appendix A. They show that each subset includes one cointegration relation which can be understood as Eq. Ž6. or Eq. Ž7., respectively. For Eq. Ž6. we accept the hypothesis that there are constant returns to scale. After having specified reasonable restrictions for the labour demand and the wage equation, we analyse if these restrictions are also accepted within a VEC of the five variables: y y l, wp, wnp, u and l. Again, we specify a system with unrestricted constant and trend in the cointegration space, but test if we can restrict the constants to zero in addition to the above restrictions. In order to test for breaks in the cointegration parameters due to the German unification, we include some dummies as exogenous variables in the cointegration relation. We assume that these exogenous dummy variables do not change the distribution of the rank statistics. Johansen and Nielsen Ž1993. provide a program ‘Disco’, which computes critical values of the rank statistics if dummy variables are included in the set of differenced variables. In the case of the German unification it seems more reasonable to assume structural breaks in the cointegration relations. By means of some Monte-Carlo experiments we found that exogenous dummy variables work quite well within the Johansen procedure. We analysed three different data sets, namely: 1. West-German data before the unification Ž1966:1᎐1990:2.;
444 G. Hansen r Economic Modelling 17 (2000) 439᎐454
Fig. 1. Data for West Germany Ž1966:1᎐1990:2. and unified Germany Ž1990:3᎐1995:4..
G. Hansen r Economic Modelling 17 (2000) 439᎐454
445
2. West-German data before and after the unification Ž1966:1᎐1994:4.; and 3. West-German data until 1990:2 and data for the unified Germany from 1990:3᎐1995:5. These data are given in Fig. 1 and show the structural break in 1990:3. 3.1. Results for West Germany before the unification We first analysed the subsets of variables in the labour demand and wage functions. Table A2 in Appendix A shows the result for the labour demand function. There is only one cointegration relation in this subset. The coefficients fit well to what they should be in a labour demand function. The output coefficient can be restricted to one so that we can use labour productivity as one variable in what follows. For the subset of variables in the wage equation we also got one cointegration relation with similar coefficients as in Tables 1a and 1b. The results with respect to the cointegration rank are given in Table 1a. We accept that there are two cointegration relations in this set. From Table 1b it follows that the above restrictions are accepted with marginal significance level SL s 0.09. Coefficients  o ,  1 ,  2 Žand ␣ o , ␣ 1 , ␣ 2 . in Table 1b refer to different sets of restrictions. Due to constant returns to scale we include log productivity y y l instead of including y and l separately. The parameters in the restricted vectors  1 are plausible in terms of Eqs. Ž6. and Ž7. and are mainly the same as in the subset models. It has to be recognised that coefficients of wp and t have the opposite sign, because y y l enters the equation instead of l y y as normalised variable. It is shown in Hansen et al. Ž1998. that in finite samples classical 2-tests reject the null too often due to fat tails of the distribution of ˆ. Mittnik et al. Ž1996. emphasise modified critical values c 2 for significance level ␥ according to ln c 2 Ž d.f.. f 1.5725ln 2 Ž d.f.. y ln Ž 1 y ␥ . 0.6317 2 Ž d.f.. 2
y0.0434 Ž ln 2 Ž d.f... , where 2 Žd.f.. is the classical 2-statistic.
Table 1a Test for cointegration rank r
0
1
2
3
4
m U Trace tr U
0.3758 43.36 24.35 105.23 86.96
0.2883 31.28 20.41 61.87 62.61
0.1583 15.86 16.73 30.58 42.20
0.0944 9.13 13.08 14.72 25.47
0.0590 5.60 12.39 5.60 12.39
446
Table 1b Restricted cointegration vectors  and loading coefficients ␣ a yyl
wp
wnp
1.0 y0.123
0.410 y0.610
1
1.0
y0.579 Ž13.79. 0
y1.208 Ž21.96.
2
1.0 y0.293 Ž2.63.
y61.388 Ž8.48. 0
y1.498 1.0 0 1.0 0 1.0
y1.787 0.784 0 1.925 Ž13.85. 0
l
Time
Const.
y0.529 0.416
y0.001 0.000
0.000 0.000
y0.002 Ž7.0. 0
0
y0.035 Ž0.78. 0
0
0 0.957 Ž8.78. 0
1.098 Ž5.25.
0.358 Ž2.63.
␣o
0.131 Ž2.32. y0.009 Žy0.10.
0.230 Ž5.17. 0.146 Ž2.19.
0.350 Ž7.36. y0.059 Žy0.83.
y0.008 Žy1.05. 0.023 Ž1.95.
0.031 Ž2.21. 0.009 Ž0.43.
␣1
0.011 Ž0.18. y0.131 Žy1.57.
0.130 Ž2.69. y0.099 Žy1.50.
y0.038 Žy0.71. y0.379 Žy5.19.
0.011 Ž1.20. 0.025 Ž2.07.
0.003 Ž0.19. y0.042 Žy2.04.
␣2
y0.001 Žy0.89. y0.134 Žy1.11.
y0.001 Žy1.47. y0.189 Žy1.95.
y0.005 Žy5.66. y0.624 Žy6.04.
ARCH Ž6. Norm. Ž10. a
0 0
0.000 Ž0.06. y0.003 Žy0.11.
5.09
5.33
5.95
8.46
4.40
6.66
1.07
0.49
6.35
1.92
SLwLMŽ1.x s 0.94, SLŽnorm. s 0.08.
0
0
2 Žd.f.. SL Ž 2 . wSL Žc 2 .x
9.62 Ž5. 0.09 w0.57x 10.12 Ž7. 0.18 w0.78x
G. Hansen r Economic Modelling 17 (2000) 439᎐454
o
u
G. Hansen r Economic Modelling 17 (2000) 439᎐454
447
Hansen et al. show that this modified test has the correct size. The c 2-statistic and its marginal significance level are given in brackets in Table 1b᎐3b. ARCH, norm, SLwLMŽ1.x and SLŽnorm. are the test-statistics for ARCH effects, normality and the significance level of the LM-test for first order autocorrelation and normality, respectively. Degrees of freedom of these test statistics are given in parentheses. Although the 2-test accepts the restriction that the unemployment rate is weakly exogenous, this restriction changes the parameters of the labour demand function Ž  2 . towards implausible values. 3.2. Results for West Germany including the unification period Our next topic is to analyse if and how this labour market system changes due to the German economic and monetary union on 1 July 1990. First we estimate the model specified so far for the period 1966:1᎐1994:4. We got the following results not reported here: 1. the cointegration rank is now likely to be 1; 2. if we analyse the variables of the labour demand function and impose one cointegration vector as well as constant returns to scale, this is strongly rejected, although the parameters are mainly unchanged; 3. the wage setting restrictions are accepted with marginal significance level SL s 0.06, but the coefficients of employment are now doubled with other coefficients mainly unchanged; and 4. if we still assume two cointegration vectors, the coefficients change tremendously compared to those for 1966:1᎐1990:2. Taken together there is evidence for a change in the model due to the unification. We therefore test if there is a break 1. in the unemployment cointegration parameter by defining a dummy variable d s 0 for 66:1 to 90:2,u for 90:3 to 94:44 2. andror in net wages due to the increasing wage wedge after unification by defining a second dummy variable d1 s 0 for 66:1 to 90:2,wp y wnp for 90:3 to 94:44 These results in Table 2 show that the cointegration rank for the model with dummy variables is again 2. The two cointegration vectors in  1 as well as in  2 give the same parameters as in Table 1b. The dummy variables can be restricted to appear only in the wage equation. With respect to the two dummy variables in  1 there are two opposite effects. Firstly, the coefficient of the unemployment rate is much higher after 1990:2 and therefore unemployment is taken more seriously after the unification. But secondly, the increasing wage wedge dummy has a positive effect on net real wages after 1990:2, which contradicts the expectation that the increasing wage wedge should reduce net real wages in order to avoid partly the negative employment effects of higher gross real wages. But this result
G. Hansen r Economic Modelling 17 (2000) 439᎐454
448
Table 2 Labour market equations with two dummy variables for 1990:3᎐1994:4 r
0
1
2
3
4
Ža. Rank test-statistics Ž5% critical values for trace-test . 0.3767 0.2574 0.1577 0.1257 0.0619 m 52.00 32.74 18.88 14.77 7.03 U 24.35 20.41 16.73 13.08 12.39 Trace 125.42 73.43 40.68 21.80 7.03 tr U 86.96 62.61 42.20 25.47 12.39 Žb. Restricted cointegration vectors  and loading coefficients ␣a yyl wp wnp u l d1 d
1
1.0
2
y1.176 Ž18.70. 1.0 y1.196 Ž20.27.
3
1 y1.196 Ž20.27.
␣2
y0.018 Ž0.41. y0.146 Ž2.03. ARCH 11.51 Ž6.norm. 3.72 Ž10. a
y0.610 Ž10.34. 0 y0.594 Ž10.61. 0
y0.593 Ž10.59. 0
0 1.0 0 1.0
0 1.0
0.092 y0.058 Ž2.44. Ž1.46. y0.082 y0.369 Ž1.37. Ž5.82. 1.75 8.56 0.09 0.53
0
0
1.938 Ž12.42. 0
0.967 Ž8.48. 0
1.953 Ž13.29.
0.974 Ž8.77.
0 1.953 Ž13.20.
0 0.975 Ž8.78.
0.011 y0.006 Ž1.70. Ž0.55. 0.024 y0.046 Ž2.36. Ž2.67. 12.22 5.37 8.24 4.32
2 Žd.f.. SLŽ 2 . wSL Žc 2 .x 0 0 y0.003 0.395 8.64 Ž5. Ž7.5. Ž2.628. 0.12 w0.60x y3.262 28.826 0 1.278 Ž1.60. Ž1.66. Ž2.057. 0 0 y0.003 0 11.0 Ž8. Ž1.5. 0.20 w0.85x 0 1.110 0 0 Ž7.76. 0 0.130 Ž7.65.
0 0
Time
y0.003 Ž7.5. 0
Const.
0 0
0
0
0
0
0
0
0
0
11.2 Ž8. 0.19 w0.84x
SLwLMŽ1.x s 0.91, SLŽnorm. s 0.03.
depends on the multicollinearity of the two dummy variables. The results for  2 and  3 show that one of the dummy variables can be dropped. If we drop the wage gap dummy Ž d1., the negative coefficient of the unemployment rate increases from 1.953 for 1966:1᎐1990:2 to 3.063 for 1990:3᎐1994:4, reflecting the higher weight put on the unemployment problem after the unification. In the same way we get in  3 a negative influence of the wage gap dummy if the unemployment dummy is dropped, reflecting the lower increase in net real wages due to the increasing wage wedge after the unification. This is evidence that wage setters reacted to unification shocks with less support for insiders as before the unification.
G. Hansen r Economic Modelling 17 (2000) 439᎐454
449
3.3. Stable labour-market model for the unified Germany In our last exercise we take West-German data from 1966:1 up to the unification in 1990:2 and data for the unified Germany from 1990:3 to 1995:4 in order to analyse if the long-run relations in this set of data can also be described by means of a few dummy breaks. This data set is characterised by strong structural breaks in output, employment and the labour force in 1990:3 because of a 12% increase in GNP and a 29% increase in the employment Žsee Fig. 1.. After the unification employment in East Germany decreased rather fast and productivity increased, adjusting to the new real wages and a convertible currency. Therefore we tried to model the data by means of breaks in the parameter of labour productivity Ž y y l ., output Ž y ., employment Ž l . or unemployment Ž u.. The most reasonable results are obtained with a break in the output coefficient in the labour demand function Žvariable d2. rejecting the null of constant returns to scale after 1990:2. From the point of economic theory this result seems to be plausible, because by means of the unification the German capital᎐labour ratio is heavily distorted due to: 1. a lower capital᎐labour ratio in East Germany; and 2. the fact that a large part of the East-German capital stock turns out to be idle capital under conditions of a competitive open economy. The results are improved by adding a 0,1-dummy Ž d3. for 1989:4, the period when the wall came down and approximately half a million people left East Germany for the West. For this set of data the trace test gives again two cointegration vectors according to Table 3a. The restricted cointegration vectors of Table 3b give similar cointegration parameters as in the West-German model, except a much smaller employment coefficient in the wage equation. This reduction means that insider power in wage setting is substantially reduced in the labour market for the unified country. Tests for autocorrelation, ARCH-effects and normality accept the null hypothesis, except ARCH-effects in the ‘net wage’ equation.
Table 3a Rank statistics of the labour market-equations for unified Germany Ž5% critical values. r
0
1
2
3
4
m m) Trace tr U
0.3145 43.05 24.35 113.77 86.96
0.2271 29.37 20.41 70.72 62.61
0.1763 22.12 16.73 41.35 42.20
0.0927 11.10 13.08 19.24 25.47
0.0689 8.14 12.39 8.14 12.39
G. Hansen r Economic Modelling 17 (2000) 439᎐454
450
Table 3b Restricted cointegration vectors  and loading coefficients ␣ a yyl
2
1
y1.260 Ž15.2.
␣2
ARCH Ž6. norm. Ž10. a
wp
y0.498 Ž33.2. 0
wnp
0
1
u
l
0
1.647 Ž5.07.
0
0.190 Ž3.7.
0.217 Ž1.5.
0.262 Ž2.1.
y0.149 Ž1.0.
y0.084 Ž4.7.
0.69 Ž2.4.
0.051 Ž2.0.
0.090 Ž4.0.
0.045 Ž1.6.
y0.009 Ž2.7.
0.030 Ž5.5.
4.3 4.1
6.6 0.71
10.53 0.51
8.8 2.9
18.4 0.41
d2
d3
Time
0.005 Ž5.0.
y0.563 Ž2.8.
y0.002 Ž5.0.
0
16.16 Ž1.8.
0
2 Žd.f.. wSL Ž 2 .x 2.71 Ž4. 0.61 w0.80x
SLwLMŽ1.x s 0.27, SLwLMŽ4.x s 0.17, SLŽnorm. s 0.09.
With dummy variables at zero, recursive ML-estimation of a VEC is not possible. Therefore we choose normally distributed dummy variables with mean value of zero and very small standard deviation ŽS.D.s 0.0005.. Fig. 2 in Appendix A shows the test for constancy of the cointegration space. It tests if the identified cointegration parameters ˜ in Table 3b is in the cointegration space in all sub-samples H 0 : ˜ g spŽ  ., s T0 , . . . ,T. The test statistic is asymptotically 2 wŽ p y r . r x distributed with p as dimension of  . The test statistic has been scaled by the 95% quantile of the 2-distribution such that unity corresponds to a 5% significance level. Beta ᎏ Z uses recursive estimates of the short-run parameters and the cointegration parameters, whereas Beta ᎏ R uses fixed short-run parameters from the complete sample. It can be seen that the test for parameter stability of the cointegration parameters, Beta ᎏ R, is accepted. Plots of these cointegration parameters in Fig. 3 show no break at 1990:2, although there seems to be evidence for an opposite break in the productivity and employment coefficient in 1988:3 and 1988:4. We tried to eliminate this break by an additional dummy variable for 1988:3 without success for the overall stability of the model. In total we conclude that the German social and monetary union in 1990:3 did not destroy the long-run labour market equations but transfer them to East Germany. Furthermore, it did reduce the power of insiders in wage setting.
4. Conclusions Within this paper we analysed the West German labour market by means of
G. Hansen r Economic Modelling 17 (2000) 439᎐454
451
Fig. 2. Test of known beta eq. to betaŽ t . wLABOUR MARKETx.
Johansen’s maximum-likelihood method for vector error correction ŽVEC. models. We tested for insider wage setting before and after the German unification in 1990. Our results show that the West German labour market can be described by means of a five-variable VEC with two cointegration relations which satisfy standard labour demand and wage setting restrictions. We found significant negative effect of the rate of unemployment on wages and therefore strong evidence against pure insider behaviour. For the period 1966:1᎐1994:4, which includes the unification, an increasing coefficient of the unemployment rate was found, which means that the shock to the labour force from the unification is well understood in wage bargaining, although the speed of adjustment towards equilibrium may still be too slow. Finally the model has been successfully extended to explain a data set which includes East Germany after the unification. If the model is estimated using data for West Germany until 1990:2 and for the unified Germany from 1990:2 to 1995:4, a single dummy variable for a break in returns to scale reinstalls the previous results.
G. Hansen r Economic Modelling 17 (2000) 439᎐454
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Fig. 3. Recursive estimates of cointegration parameters for unified Germany.
Acknowledgements Comments by an anonymous referee are gratefully acknowledged.
G. Hansen r Economic Modelling 17 (2000) 439᎐454
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Appendix A Table A1. Augmented Dickey᎐Fuller test-statistics ŽADF. and t-statistics ŽPP. a Dataset I
Variable
Variable
Max lag
ADF
l wp y wnp u yyl
5 5 4 4 5 4
0.876 y4.50b y1.09 y4.25b y1.95 y1.66
⌬l ⌬p ⌬y ⌬ wnp ⌬u ⌬Ž y y l .
Max lag
ADF
5 4 3 4 5 3
y3.70 y3.19 y7.39 y7.167 y3.41 y8.78
Dataset II Variable
Max lag
ADF
l wp y wnp u yyl
4 4 4 4 4 4
y2.52 y1.27 y3.23 y1.54 y3.13 y3.46
a Dataset I: West-German data, 1966:1 to 1994:4; Dataset II: Data for West-German from 1966:1 to 1990:2 and for unified Germany 1990:3 to 1995:4. b With trend.
Table A2. Results for the subset of variables in the labour demand equation wEq. Ž6.x R
ma x
Trace
U max
tr U
0 1 2
0.3062 0.1281 0.0491
33.63 12.61 4.63
50.88 17.25 22.95
16.13 12.39 10.56
42.20 25.47 12.39
Table A2b. Labour demand-function cointegration vectors Ž  . and loading coefficients Ž ␣ . wARCH Žp. and norm Ž2. are test-statistics for ARCH-effects and normality, respectivelyx l

␣
ARCH Ž6. norm Ž2.
y
wp
Time
1.0
y0.995 Ž9.48. y1.0
0.514 Ž8.16. 0.514 Ž21.54.
y0.086 Žy4.85.
y0.235 Žy2.91.
y0.292 Žy4.75.
1.0
y8.10 0.16
6.07 1.04
0.0023 Ž5.10. 0.0023 Ž10.2.
2
d.f.
SL
0.00
1
0.96
3.34 0.41
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