strike frequency and industrial wage differentials: an econometric study of british production...

STRIKE FREQUENCY AND INDUSTRIAL WAGE DIFFERENTIALS :

AN ECONOMETRIC STUDY OF BRITISH PRODUCTION INDUSTRIES*

I , INTRODUCTION Several econometric studies of strike incidence have been conducted recently for the United Kingdom. Most of this work has been based upon the theoretical bargaining model developed by Ashenfelter and Johnson (1969). This model relates the incidence of strikes to both the degree of worker discontent and the resistance displayed by employers to demands made by workers.

All the previous studies using this type of model have emphasized the importance of the sluggish rate of advance in real income as the most important cause of greater worker discontent leading to increased strike frequency. Other important economic influences on strike frequency that have been emphasized are profits, productivity and unemployment. A drawback of this work is that many of these variables do not have a theoretically unahbiguous impact on strike incidence, because they affect both the degree of employer resistance and worker agressiveness with offsetting effects. Despite this there is a surprising degree of unanimity in the empirical work with the most robust result being the negative association of gross or net real income growth with strike frequency. The majority of previous studies for the U.K. have been conducted at the macroeconomic level. However, Pencavel ( I 970) has tested the Ashenfelter and Johnson model for six British manufacturing industries with much less success than with aggregate data. Poor results have also been obtained by Armstrong et al (1977) in an attempt to establish an association between industry level strike incidence and wage changes.

Strikes of course are the result of very complex social and political factors, as well as economic ones, however the aggregate results do suggest that an economic model can provide an important part of the explanation. As we consider more disaggregated data however it becomes more difficult to dis- entangle even the economic influences. One reason for poor results achieved with disaggregated data may be that the definition of industry groups and the actual impact of economic variables on strikes may not match. Furthermore what may be true of the whole is not necessarily true of the parts. However, it is the contention of this paper that when respecified to more accurately reflect the realities of the bargaining situation this type of model can provide a useful explanation of strike incidence at the industry level. For example, most strikes in Great Britain in the industries considered by Pencavel occur at factory level

* We would like to thank the Centre for Industrial, Economic and Business Research at Warwick University and the Institute of Manpower Studies at Sussex University for their support in the early stages of this research. Andrew Woods, Diane Elwood and Milly Casey gave us invaluable assistance.

66 STRIKE FREQUENCY AND WAGE DIFFERENTIALS

and in so far as wage issues are a reason for dispute, earnings rather than wage rates may be the more relevant independent variable. A further consideration is the importance of RELATIVE wage effects at the disaggregated level. At the industry level Shorey’s (1976) cross-section study indicates a positive association between the degree of deterioration in an industry’s differentials with other comparable industries and strike frequency in that industry. The models developed by Bean and Peel (1974) and by Taylor (1975) simply introduce a measure of the change in the dispersion of wages over time into an otherwise standard strike equation. They also obtain a significant relationship between strike incidence1 and a wage structure variable. However, the rationale for the inclusion of a variable which measures whether the industrial wage structure is being compressed or stretched overall is not clear.

There are many good reasons for expecting relative wages to be important in explaining industry level strike incidence. A strike signals the breakdown of negotiations between employer and worker. The likelihood of breakdown will be governed by the degree of resistance to worker demands by the employer and the aggressiveness of workers at the bargaining table. From the employer’s point of view, deterioration of differentials will reduce the flow of labour to an industry at a given level of hiring standards. Quits may also increase. In these circumstances the employer’s willingness to concede will increase and the probability of a strike will decrease. The opposite is likely to be the case as far as workers are concerned as the existing literature (Bean and Peel (1974), Taylor ( I 975) and Shorey (1976) ) has emphasized. Deterioration in differentials from the worker’s point of view represents not only access to an inferior income stream but may also signify a decline in status. This effect has long been recognized in the economics literature. Keynes ( 1 9 3 6 ) ~ for example, saw resistance to money wage cuts as the result not of the threat to real wages, but to the existing relative wage structure posed by such cuts. We would expect, therefore, that the deterioration of differentials would, by increasing the aggressiveness of workers, increase the probability of a strike occurring. The problem with the existing literature is that it simply acknowledges the latter effect without also taking account of the employer’s reaction to deteriorating differentials, which may decrease the probability of a strike occurring.

2. THE MODEL The model used in the current study is based on that developed by one of the

authors in a previous paper (Knight, 1972). This model is a development of the Ashenfelter and Johnson model modified to take account of the prevalence of plant level bargaining in Great Britain. A modified version of the estimation equation derived by Knight is given here as equation ( I ) . For a full discussion of the theory underlying this equation the reader is referred to the earlier paper

Strike incidence in the form of strike frequency in Bean and Peel (1974) and working days lost in the case of Taylor (1975). The former measure is preferred in the present study being less sensitive than measures such as working days lost to such events as token national stoppages in the engineering industry.

a Page 14.

STRIKE FREQUENCY AND WAGE DIFFERENTIALS 67

which also provides justification for the use of the change in the number of strikes as the dependent variable.’

+ $ u ~ ~ L A D : ~ - I - A + ~ 5 i Ut + v A=1

where 5‘i = number of strikes in the ith sector W’i = hourly wage rates or hourly earnings in the ith sector P = retail prices U = aggregate unemployment q is an error term

The change in the number of strikes is related to the change in money wages, the change in retail prices and an unemployment term. Theory and evidence from Knight and other studies suggest that the expected coefficient signs are positive for a2 and negative for a1 ( ie . slow rates of advance in real wages increase worker discontent and hence strike incidence). The expected sign of a5 is more ambiguous but has generally been found to be negative (i.e. tight labour markets increase employee aggressiveness and lower employer resistance and vice versa). In Knight’s original paper the strike equation was combined with an equation explaining changes in earnings as a function of changes in strikes. The results of estimating this model suggested that a recursive model was most appropriate with strike incidence being determined by lagged changes in earnings and prices. Similar results have been obtained in other studies. I n the present paper we therefore concentrate upon the strikes equation in isola- tion. Following Ashenfelter and Johnson (1969) we experimented with the use of Almon lag techniques (obtaining the best results with A set equal to 6 quarters).

The novel element in equation ( I ) is the measure of deterioration in relative wages. As we have noted above, changes in wage differentials may be expected to affect the incidence of strikes via their impact on both employers and workers. It is important however to clarify the meaning of the term deterioration in differentials. As noted in the previous section, most previous studies have used a general measure of the variation in the wage structure to monitor the effects of a general stretching or compression of differentials over time. This type of measure fails to recognize that for a particular industry a deterioration of differentials implies compressions of differentials relative to industries below it in the wage structure but stretching of the wage differentials relative to industries above it in the wage structure. Conventional measures such as the standard deviation fail to recognize that both stretching and compression of differentials

1 Re-estimation in levels might improve the explanatory power and fit but would not increase our understanding of the causes of strike incidence. Previous work in levels has generally found a very significant time trend, simply proving that the incidence has increased, e.g. Pencavel (1970). Use of first differenccs reduces the multicollinearity problems that plague such an approach and enables the effect of different economic factors to be distinguished. For further discussion see Knight ( I 972).


are consistent with a deterioration in wage differentials for any particular industry.

In equation ( I ) the measures of deterioration in relative wages are given as AD:, and AD;,. These are calculated in the following manner:

ADit = Dit - Dtt-l Di-t = 2, (Wit - Wjt) Dt‘t = z k (Wkt - wtt)

(2)

(3) (34

where Wt = wage rateslearnings in the ith industry Wj = wage rateslearnings in the j t h industry BELOW the ith

w k = wage rateslearnings in the kth industry ABOVE the ith industry

A deterioration in the relative position of any given industry can take one or both of two forms. Firstly, if industries below the given industry in the pay ranking are catching up with that industry, DT, will fall in value and AD;, will be negative. Secondly, the relative position of a given industry could deteriorate because industries above it in the pay ranking are pulling away. In this case D t , will increase in value and ADt, will be positive. A general stretching and a general compression of the pay structure will cause one (and only one) of these forms of deterioration although both could occur in a specific instance of industry decline. Measuring changes in relative advantage in this way concen- trates attention on changes in absolute differentials. This also contrasts with the measures used in previous studies’ but we would claim several advantages for this approach. Firstly, it is in accordance with the findings of both Behrend (1974) and Daniel (1975) that workers are strongly influenced in their bargaining behaviour by absolute rather than relative differentials. Secondly, measures employing the relative differential may imply no deterioration though the absolute differential may be widening. This may lead to bias in measuring the impact of differentials on strike incidence. Thirdly, this division of differentials permits consideration of differing reaction to deterioration caused by industries pulling ahead and by industries catching up. This differing reaction may result from the employee’s view of which industries he should be comparable with or to the employer’s view of which industries he competes with for labour. Fourthly, our measures permit us to distinguish the interest in relative wages of both employers and employees and to distinguish the impact of their interest. The employer’s willingness to concede will increase if the relative position of an industry has deteriorated for reasons stated in the introductory section of this paper. In this case, the coefficients on ADr, will be positive and on AD:, will be negative. This will reduce the probability of a strike occurring and therefore our expectation will be a3 > o and a4 < 0. On the other hand, in these circumstances the aggressiveness of workers is likely to increase thus

industry in the ranking of pay

in the industry ranking of pay.

1 Shorey (1976) and Taylor (1975) employ measures which utilise differences in the rates of wage increase by industry. Bean and Peel (1974) employ a relative wage level variable.


enhancing the probability of a breakdown in negotiations (and a strike occurring). If this effect is paramount we expect as < o and a4 > 0. As in previous studies in this area we are therefore faced by a variable whose sign is dependent on the relative size of its impact on employer and employee at the bargaining table. I t might be argued that this type of measure implies too simple a view of the effect of relativities on bargaining. In particular it does not take into account the role of ‘key bargains’ or ‘leading sectors’. On the other hand we would argue that it is an improvement on previously used measures. I t is difficult to identify a key bargain by statistical methods and the selection is inevitably arbitrary. Furthermore the key bargain may change through each bargaining round. In any event the key bargain will disrupt the existing wage structure and this is picked up by this method of measuring differentials.

3. THE RESULTS The basic model of equation ( I ) was estimated using both hourly wage rate

and hourly earnings data’ for production industries in Great Britain. Data sources are described in the Appendix. The model was fitted to six-monthly data for the period 1959 11 to 1973 11. In common with Pencavel (1970)~ and Ashenfelter and Johnson (1969) the Almon technique was used to identify the effect of lamed values of

on strike incidence.2 Significantly superior results were obtained with hourly earnings rather than wage rates as the real wage variable which supports the contention that it is earnings not basic rates that is the crucial issue when wage disputes occur. The results using the hourly earnings data are included in Table I . ~ The equations are generally well determined with no evidence of positive autocorrelation, although there are limited indications of negative autocorrelation. Explanatory power is high for an equation with the change in strikes as the dependent variable. Coefficients of the lagged values of the change in hourly earnings have the conventionally expected sign in I I out of 14 cases and are statistically significant in 8 cases including the largest industry, engineering and shipbuilding. This is in contrast to the poor results obtained by Pencavel (1970) and Armstrong et af. (1977) using wage rate data. I t seems

An extended version of the model was estimated. This was to test the impact of further measures of employers resistance to claims. Data limitations prevented the use of a profits variable used in aggregative studies but a capital utilization variable was constructed. Our expectation was that the sign on this variable would be negative. In fact the inclusion of this variable left the coefficients of all other variables almost unaffected in sign and significance. The utilization variable itself had a positive sign in eight cases (only one significant) and a negative sign in six cases (one significant). We do not report these results but they are available on request from the authors.

* Each equation was estimated for each industry assuming a linear and a quadratic form of the Almon lag structure. Table 1 indicates in the case of each industry the preferred equation. In all cases I , = 6 and we present the values of the individual coefficients on the lagged variables. In Table I vehicles is consistently the highest paid sector and therefore 0: = o for all periods.

* The results using wage rates are available from the authors on request. Using wage rates, the coefficients on lagged wages, prices, unemployment are significant with expected signs in only two, one and zero cases respectively. The coefficients on the relative wage variables are significant in five cases.

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clear that hourly earnings is the appropriate variable to employ in industry level strike equations.

In the case of the price change variable coefficients had the anticipated positive sign in 13 out of 14 cases. These coefficients were significant in only 4 equations, this being due primarily to multicollinearity between the wage and price variables. A noteworthy feature of these coefficients is that in most industries they are significantly different from the money wage coefficients, indicating differential response to differential causes of real wage erosion. In addition, there are indications of an inverse association between the absolute size of the two coefficients suggesting that industries where there is a large strike response to high price increases react to low money wage increases to only a minor extent, and vice versa.

The performance of the labour market variable (unemployment) was a good deal less satisfactory than in previous studies. Although in 10 out of the 14 cases the sign was as anticipated (negative), it was statistically significant in only 4. One explanation for these results may be the mid-sixties break in the unemployment/vacancies relationship which may have rendered the former variable inadequate as a labour market indicator for the period covered by this study. Another is that although the prime impact of greater unemployment might be expected to operate in its impact on reducing union aggressiveness, it might also have various secondary effects. For example, reduction in the overall level of activity may make employers take a tougher stance. Furthermore, higher unemployment may with a largish lag eventually lead to frustration and discontent amongst workers if changes of job mobility and promotion are threatened thus eventually outweighing the initial effect on union aggressiveness.

Relative wage variables were significant in 7 out of 14 industries.1 In all of these cases the ‘pull ahead’ ( AD;,) variable had a significant negative coefficient. This indicates that periods in which an industry experiences a deterioration of absolute differentials compared with higher paying industries are followed by increased willingness on the part of employers to concede a pay demand. Increased quits, the greater difficulty of recruitment, and/or the lower standard of new recruits appears to result in a less intransigent attitude at the bargaining table by employers with the result that fewer strikes occur. The counter effect on employee militancy appears to exert a less powerful influence on strike incidence. This result lends support to those who argue that if a spill-over effect occurs it is likely to be the result of competitive pressures on employers encouraging agreement to pay claims based on comparability with leading sectors. In this case institutional and market pressures will coincide to produce the same result in terms of wage increases secured in the non-leading sectors.

The ‘catch up’ variable (AD;,) was significant in 4 out of 14 cases. In 3 of these industries the coefficient was significantly negative. These industries were the engineering, shipbuilding, vehicles and printing industries, all of which have a higher than average level of unionization. In these cases employee

We estimated equation ( I ) excluding the relative wage variables. From these results it is clear that while their inclusion has virtually no effect on the other coefficients (in terms of signs or significance) it significantly improves the overall fit, with Rz being increased in all cases.


pressure to restore absolute differentials eroded by less well paid industries catching up appears to result in a significant increase in strike incidence. An important factor in this is likely to be that the employer in these relatively higher paid industries is under no competitive pressure to concede such increases so a more intransigent bargaining stance results. The pay position of these industries will ensure an advantage in hiring over lower paid industries, even in the presence of substantial narrowing of the absolute differential. Another interesting feature of the results is the fact that the significance of the relative wage variables is clearly related to the position of the industry in the wage structure. For high paying industries the 0: variable refers to very few industries and similarly for the DF variable and low paying industries. There- fore, we would not expect them to be of great importance either from employee or employer viewpoints. Thus, not surprisingly, the four industries for which the Dc variable is most significant are amongst the highest paid and vice versa,

4. CONCLUSIONS

While we would not want to read too much into the present set of results, we feel that they provide evidence for some tentative conclusions. First, and in support of Knight's earlier results a t the aggregate level, the use of hourly earnings rather than wage rates significantly improves the results from industry level strike equations. Second, further improvement is obtained by the use of relative wage variables. As far as the earnings equations are concerned, the results suggest that relative wage effects are important because of the impact of a deterioration of absolute differentials (particularly in higher paying industries) on the perceived potential ease with which labour is retained and hired. In this case union aggressiveness is conceded by employers and fewer strikes result. Employee pressure leading to strike incidence appears to derive its greatest impetus from slow advance in real pay rather than from deterioration in inter- industry differentials although in a few high paying, highly unionized, industries both are important. Clearly we have not captured the full effect of movements in differentials. Occupational, regional, skill and intra-industrylinter-plant differentials may exert an equally great or greater impact.' However, it is not possible to test such hypotheses at the level of aggregation used in this study due to inadequate data. K. G. KNIGHT

R. A. WILSON UNIVERSITY OF WARWICK MANPOWER RESEARCH GROUP UNIVERSITY OF WARWICK

Date of receipt ofJim1 typescript: March 1980

* The importance of comparisons of this type in wage determination has been emphasized in the industrial relations literature, e.g. W. Brown and K. Sissons (1975).


APPENDIX DATA SOURCES AND METHODS

All rates of change are calculated as follows:

X xt-1 t = year

All data was obtained from the D.0.E. (Ministry of Labour) Gazette. S

each year P = retail price index U = percentage unemployed in all industries, Great Britain.

Hourly earnings data was obtained from the six-monthly earnings and hours enquiry and the New Earnings Survey and relates to males over 2 1 . Earnings figures were not adjusted for overtime. It might be argued that some adjust- ment is necessary to purge this variable of any cyclical effect. On the other hand a significant amount of overtime is institutionalized and any attempt to adjust the data for the purely cyclical element would necessarily be ad hoc. Our results using wage rates as opposed to earnings indicate the superiority of the latter variable, however the impact of this variable on strikes must be regarded as including the effects of disputes about normal hours as well as in rates of pay. Since our dependent variable measures all disputes including those concerned with hours worked this procedure does not seem unreasonable.

REFERENCES ARMSTRONG, K., BOWERS, D. and BURKITT, B. (1g77), ‘The Measurement of Trade

Union Bargaining Power’, British Journal of Industrial Relations, March. Vol. xv,

ASHENPELTER, 0. and JOHNSON, G. E. (1969), ‘Bargaining Theory, Trade Unions and Industrial Strike Activity’, American Econoniic Review, March, Vol. 59, pp. 35-49.

BEAN, R. and PEEL, D. A. (1g74), ‘A Quantitative Analysis of Wage Strikes in Four U.K. Industries 1962-1970’, Journal of Economic Studies, November, New Series,

BEHREND, H. (1g74), ‘The Impact of Inflation on Pay Increase Expectations and Ideas of Fair Play’, Industrial RelationsJournal, Vol. 5, No. I , pp. 5-10.

BROWN, W. and SISSONS, K. (1g75), ‘The Use of Comparisons in Workplace Wage Determination’, British Journal of Industrial Relations, March, Vol. XIII, No. I , PP. 23-53.

DANIEL, W. W. (1g75), The P E P Survey on InJlation, PEP. JACKSON, D., TURNER, H. A. and WILKINSON, F. (I972), Do Trade Unionr Cause Inflation?,

Cambridge University Press. KEYNES, J. M. (1936), General Theory of Emfiloyment, Interest and Moncv, London:

Macmillan. KNIGHT, K. G. (1g72), ‘Strikes and Wage Inflation in British Manufacturing Industry,

1950-68’, Bulletin qf Oxford University Institute of Economics and Statistics, August, Vol. 34, No. 3, pp. 281-94.

PENCAVEL, J. (1970), ‘An Investigation into Industrial Strike Activity in Britain’, Economica, August, Vol. XXXVII, No. 147, pp. 239-56.

SHOREY, J. ( 1976), ‘An Inter-Industry Analysis of Strike Frequency’, Economica, Novem- ber, Vol. 43, No. 172,. pp. 349-65.

SHOREY, J. ( 1977), ‘Time Series Analysis of Strike Frequency’, British Journal o j Industrial Relations, MaIch, Vol. xv, No. I , pp. 63-75.

TAYLOR, J. (1g7=j), ‘The Organised Pressure for Higher Wages’, in &I. Parkin and A. Nobay, Contemporary Issues in Economics, University of Manchester Press.

= number of strikes beginning in each industry in each half of

NO. I, pp. 91-100.

Vol. I , NO. 2, pp. 88-97.

strike frequency and industrial wage differentials: an econometric study of british production...

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