Ours, J.C. van

English

Research Papers in Economics

 
 
 
  The Cyclical Behavior of 
Holdout Durations 
   
 
  Jan C. van Ours 
 
  October1998 
    
 
 
 
 
  ABSTRACT 
 
Holdouts are the negotiation periods between the expiration of an old wage contract and 
the signing of a new one. This paper presents an analysis of the hazard function for 
holdout  durations  using  data  on  wage  negotiations  in  The  Netherlands.  The 
unemployment rate is found to have a significant negative effect on the holdout hazard 
rate: holdout duration is counter-cyclical. 
 
 
 
Keywords: Holdouts, wage bargaining 
 
JEL Code: J30, J50 
 
 
 
 
 
 
 
 
 
 
 
 
 
Address: Jan C. van Ours, Department of Economics, CentER for Economic Research, 
Tilburg University, P.O. Box 90153,  5000 LE Tilburg, The Netherlands. 
 
Phone +31-13 4662316; Fax +31-13 4663042; Email vanours@kub.nl    
 
 
 
1.  Introduction 
 
Holdouts are the negotiation periods between the expiration of an old wage contract and 
the signing of a new one. Holdouts occur if agreements cannot be struck before existing 
contracts expire. During a holdout the terms of the old contract apply. There is not a lot of 
research on holdouts, but they seem to be related to strikes. Cramton and Tracy (1992) 
present a theoretical model in which holdouts are used by unions to obtain information 
from  firms  at  lower  costs  than  strikes.  Van  Ours  and  Van  de  Wijngaert  (1996)  find 
empirical evidence that holdout durations have a negative effect on wage increases. This is 
similar to the way strike durations affect wage increases.  
  This  paper  investigates  the  cyclical  behavior  of  holdout  durations  in  The 
Netherlands. There is little theory on this type of fluctuations. Cramton and Tracy (1994) 
indicate that the decision of a union to opt for a holdout in stead of a strike is sensitive to 
economic changes. Furthermore, a holdout will be longer if there is more uncertainty over 
the firm’s willingness to pay. Then if there is more uncertainty in bad times, holdout 
durations are counter-cyclical. 
  As Kennan (1985) did with strike durations, we investigate the cyclical behavior of 
holdout durations by estimating hazard functions. We use the unemployment rate as an 
indicator of the state of the business cycle. Usually, when such an economy-wide indicator 
is used, strike durations are found to be counter-cyclical (Kennan and Wilson (1993)). We 
find the same counter-cyclical behavior for holdout durations. 
 
 
2. Data  
 
The data are the same as in Van Ours and Van de Wijngaert (1996). Information about 
holdouts is gathered from administrative files and individual collective agreements of  7 
industries  (Metal  industry,  Cigar  industry,  Printing  industry,  Manufacture  of  dairy 
products,  Breweries,  Manufacture  of  printing  ink,  Insurance)  and  8  firms  (Philips, 
Unilever,  Douwe  Egberts,  Heineken,  Akzo,  Chemiefarma,  AMEV,  DSM).  The  15 
bargaining pairs provide 150 data points in a balanced panel dataset covering the period 
1975-1987, with no observations for the years 1976, 1981 and 1984. For the year 1976 this 
is due to government interference, when the government decided that changes in on-going 
contracts were not allowed. In 1981 and 1984 there where no negotiations because of the 
evolving biannual agreements in 1980 and 1983. 
  Figure 1 presents the frequency distribution of the holdout durations in our sample 
showing that there is a wide range in these durations. On the one hand about 50% is 
shorter than 6 months, on the other hand about 10% is longer than 1 year. Holdouts in the 
Netherlands are relatively long as compared to the US. Cramton and Tracy (1992) who 
define a holdout as the time between the expiration of the former contract and the time of a 
principle  agreement  find  an  average holdout  of about one-month.  Our definition of a 
holdout not only covers the negotiation period but also the period between the time of a 
principle agreement and the moment of registration at the government office. The latter 
period may be time consuming when experts have to settle details of the arrangement. This    
 
 
 
period in which the final text is drafted and the contract is signed averages 2-3 months. 
Since new agreements are commonly backdated to the date at which the former contract 
elapsed, there is little pressure to shorten holdouts. However, if holdouts are lengthy only 
because of administrative procedures, they would not be related to the economic cycle. 
 
   
3.  Industrial relations in the Dutch labor market 
 
In The Netherlands collective bargaining covers the employment terms of 70-80 percent of 
the labor force in the private sector. Employer coverage approximates some 90 percent of 
all firms. These figures indicate that collective bargaining has a large impact on wage 
formation. In September of each year the Dutch government presents its budget and its 
macroeconomic forecasts to Parliament. Taking this into account, the managements of 
union  and  employers’  federations  consult  their  associated  trade  unions  and  employer 
associations on the common policy for the national negotiations. Actual wage bargaining 
starts in the new calendar year and is the responsibility of the trade unions, which are 
organized  by  industry.  In  the 1980s about  200 industry  contracts  regulated wages for 
approximately  80  percent  of  the  workers  under  collective  contracts.  The  employment 
terms for the remaining 20 percent of the workers were covered in 600 firm contracts since 
larger firms (multinationals) have their own collective agreements.  
  There do not appear to be many serious conflicts between employers and unions. 
The low strike activity in the Netherlands is reflected in our sample. In none of the firms 
and industries there was a strike in the period of analysis. The situation in the labor market 
in the 1970s differed from the one in the 1980s. As in many other European countries 
economic  conditions  in  the  Netherlands  deteriorated  dramatically  in  the  early  1980s. 
Average unemployment rate in the 1970s was 4.5%, while it was 13.9% in the period 
1980-87. 
 
 
4. Empirical analysis 
 
The evolution of the empirical hazard rate over the duration of the holdout is shown in 
Figure 2. The hazard rate steadily increases until 14 months and then starts fluctuating. 
These fluctuations have to do with the limited number of observations at longer durations. 
  The empirical analysis is based on traditional specifications of the hazard rate which 
is assumed to be of the mixed proportional type. So, the hazard rate is specified as the 
product  of  functions  of  observed  characteristics,  elapsed  duration  and  unobserved 
heterogeneity: 
 
  q(tit;xit,vi) = qo(xit).¦(tit).exp(vi)          (1) 
 
where q is the hazard rate, t is the elapsed duration, t an index of calendar time (1,..,10), i 
an  index  of  bargaining  pair  (1,..,15),  x  a  vector  of  observed  characteristics,  ¦(tit)  the 
baseline hazard and v represents unobserved heterogeneity, which is assumed to come    
 
 
 
from a probability density function h(vi). The survival function is specified as: 
 
  S(tit;xit) =  o,
￿
 exp(-o,
s q(tit|xit,vi)ds) h(vi) dv        (2) 
 
where S(t) is the survival function with g(t)=dS(t)/dt as the probability density function of 
the holdout durations. Since we have do not have right censored durations and have inflow 
samples, the likelihood is straightforward: 
 
  L = -- g(tit;xit)              (3) 
                  
i   t 
With respect to the observed characteristics we assume the following specification: 
 
  qo(xit) = exp(bi +bo.ut)            (4) 
 
where the bi’s represent bargaining pair specific fixed effects and ut is the unemployment 
rate at the start of calendar year t. If the coefficient bo is negative, holdout duration is 
counter-cyclical, if it is positive holdout duration is pro-cyclical. 
  To  investigate  the  consequences  of  particular  specifications  of  the  duration 
dependence and the unobserved heterogeneity we compare the estimation results of two 
different functional forms. The first specification is based on the Burr distribution which is 
a Gamma mixture of Weibull distributions. The hazard rate is specified as: 
 
  q(tit;xit,vi) = qo(xit).a.tit
a-1/(1+s
2.qo(xit).tit
a)        (5) 
 
where  a  is  the  duration  dependence  parameter  of  a  Weibull  distribution  and  s
2  the 
variance of a Gamma-distribution with mean 1.  
  The parameters of the model have been estimated using the method of maximum 
likelihood.  The  first  column  of  Table  1  shows  the  estimation  results  for  the  Burr 
distribution. It appears that a is significantly larger than 1 indicating that there is positive 
duration  dependence.  The  hazard  rate  increases  monotonically.  Since  s
2  is  not 
significantly larger than 0 there is no unobserved heterogeneity. The coefficient of the 
unemployment  rate  is  significantly  smaller  than  0  indicating  that  holdout  duration  is 
counter-cyclical. The results in the second column of Table 1 show that the estimation 
results do not change if we impose that s
2=0. This confirms that there is no unobserved 
heterogeneity,  so  the  Burr distribution is  reduced  to a Weibull distribution. The third 
column shows the estimation results if we impose a=1 implying that there is no duration 
dependence. Since the estimation results deteriorate this restriction is not allowed. The 
fourth column shows the estimation results if we impose s
2=1 thus assuming that the 
hazard rate has a log-logistic specification. Such a log-logistic specification with a >1 
implies that the hazard rate first goes up and then goes down. Comparing the results of the 
first and the fourth column, it is obvious that such a restriction is not allowed. So, the 
Weibull distribution is superior to the log-logistic specification. 
  The second specification of the hazard rate is the piecewise constant hazard rate:    
 
 
 
 
  q(tit;xit,vi) = q0(xit).exp(SkbkIk+vi)          (6) 
 
where the Ik’s are dummy variables for the quarterly duration intervals, k=1,..6 and the 
bk’s are the coefficients representing duration dependence, with b1 normalized to 0. Now, 
unobserved heterogeneity is specified as a discrete distribution with two points of support. 
The estimation results again indicate that unobserved heterogeneity is not present. The 
fifth column of Table 1 shows the estimation results when unobserved heterogeneity is 
ignored. The parameter of the unemployment rate in the piecewise constant hazard rate has 
about the same value as in the Weibull specification.  
  Finally,  we  estimated  the  coefficients  of  the  hazard  rate  using  Cox’s  partial 
likelihood estimator in which the baseline hazard is not specified. As shown in the sixth 
column of Table 1 we again find a significant negative effect of the unemployment rate of 
about the same size as before. Apparently, this result is robust and independent of the 
particular specification of the hazard function. 
 
 
5. Conclusions 
 
In the empirical analysis we find a significant negative effect of the unemployment rate on 
the  hazard  rate  of  holdouts.  This  relationship  may  be  caused  by  the  high  level  of 
uncertainty that unions face when unemployment is high. If the holdouts are intended to 
derive  information  about  the  firm’s  willingness  to  pay,  then  more  uncertainty  means 
longer holdout durations. Therefore, holdout durations are counter-cyclical. In this respect 
holdout durations resemble strike durations.    
 
 
 
Table 1 Estimation results hazard rate models (t-values)
a) 
   
              Flexible
b) 
  Burr   Weibull  Exponential Log-logistic  duration  Cox 
  distribution   distribution  distribution distribution  dependence regression     
   
u  -0.066 (2.3)   -0.064 (2.3) -0.037 (0.8) -0.122 (3.0)  -0.064 (2.0)  -0.065 (2.8) 
a   2.16 (9.6)     2.12 (15.7)   1 (-)    3.18 (16.8)     -       - 
s
2   0.03 (0.1)     0 (-)    0 (-)    1 (-)              -       - 
 
-logL    387.5      387.5    435.6    393.5      388.6   570.1 
 
a) In all the estimates dummy-variables for every bargaining pair are included; the coefficients of 
these dummy-variables are not presented in the Table. 
b) The coefficients for the quarterly duration intervals 2-6 (6 being the open interval of more than 5 
quarters) are: 1.12 (3.6), 1.95 (5.9), 2.24 (6.1), 2.58 (4.1), 2.96 (3.8). 
 
 
 
 
 
 
 
 
 
REFERENCES 
 
Cramton, P.C. and J.C. Tracy (1992), “Strikes and Holdouts in Wage Bargaining, Theory and 
Data”, American Economic Review, 82, 100-121. 
 
Cramton, P.C. van J.C. Tracy (1994), The Determinants of U.S. Labor Disputes, Journal of 
Labor Economics, 12, 180-209. 
 
Kennan,  J.  (1985),  “The  Duration  of  Contract  Strikes  in  U.S.  Manufacturing”,  Journal  of 
Econometrics, 28, 5-28. 
 
Kennan, J. and R. Wilson (1993), “Bargaining with Private Information”, Journal of Economic 
Literature, 31, 45-104. 
 
Ours, J.C. van and R.F. van de Wijngaert (1996), “Holdouts and Wage Bargaining in The 
Netherlands”, Economics Letters, 53, 83-88.     
 
 
 
Figure 1 Frequency distribution of holdout durations 
 
 
 
0
2
4
6
8
10
12
14
1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23
Duration in months
P
e
r
c
e
n
t
a
g
e   
 
 
 
Figure 2 Monthly holdout hazard rates 
 
 
0
0,1
0,2
0,3
0,4
0,5
0,6
0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22
Duration in months
H
a
z
a
r
d
 
r
a
t
e

The cyclical behavior of holdout durations.

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