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Keywords:

  • Informalization;
  • Industrial labor;
  • Labor reforms;
  • Import competition
  • J23;
  • J24

Abstract

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

Since the 1980s, industrial labor in India has been increasingly informalized, manifested in a rising share of unorganized sector employment and the growing use of temporary and contract workers, and subcontracting in organized manufacturing. Using unit-level data from the National Sample Survey employment–unemployment survey for 2004–5, the paper investigates econometrically whether labor market rigidities and import competition have been responsible for the informalization of industrial labor in India. The results of econometric models show that labor market reforms tend to increase the creation of regular jobs, while import competition tends to raise casual employment among workers with education levels above primary.


I. INTRODUCTION

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

Since the 1980s, industrial labor in India has been increasingly informalized. This has resulted in a rising share of the unorganized sector in total manufacturing employment, and informalization of the organized manufacturing sector itself with greater use of subcontracting and increasing employment of contract and temporary workers. The share of unorganized manufacturing in total manufacturing employment has increased from 80.7% in 1983, to 83.2% in 1993–94, and further to 85.0% in 2004–5. Workers employed through contractors (hereafter, contract workers) as a percentage of total workers employed in organized manufacturing has increased from 14% in 1995–96 to 31% in 2007–8.1 There has probably been a similar increase in the share of temporary workers in employment in organized manufacturing, though from the available data it is not possible to make a proper assessment. Maiti and Mitra (2010) have presented estimates of informal employment in manufacturing for 1999–2000 and 2004–5.2 According to their estimates, the share of informal sector in the manufacturing workforce increased from 78% in 1999–2000 to 85% in 2004–5. The implication of these changes is that the bulk of the new jobs created in the formal sector of Indian manufacturing in the period 1995 to 2005 (if not over a longer period) were low quality, informal jobs. Thus, in terms of the creation of decent jobs, India's organized manufacturing has not been doing well. According to the estimates made by NCEUS (2009), employment in the formal sector of Indian industry increased from 20.27 million in 1999–2000 to 25.38 million in 2004–5. The bulk of this increase in employment was the increase in informal workers employed in the formal sector of Indian industry—from 12.13 million to 16.71 million (NCEUS 2009). The increase in formal workers employed in the formal sector of Indian industry was from 8.14 million to 8.67 million. According to the NCEUS estimates, the proportion of informal workers in the formal sector of Indian industry in 2004–5 was about 66%, up from about 60% in 1999–2000.3

The phenomenon of increasing informalization of industrial labor is a serious issue because if industrialization does not create many good jobs for people to shift from low productivity occupations, it cannot make a significant contribution to economic development. Available data show that wages and employment benefits received by casual workers are much lower than those of regular salaried/wage workers4 and the incidence of poverty is much greater among casual workers than regular salaried/wage workers (hereafter regular wage workers). Estimates made from unit-level data from the National Sample Survey (NSS) 61st round employment–unemployment survey reveal that in 2004–5, the average wage earned per day by regular wage workers in organized manufacturing was about Rs169, while that earned by casual workers was only about Rs55.5 In unorganized manufacturing, the average wages earned per day by regular wage workers and casual workers, in 2004–5, were Rs83 and 54, respectively. According to the estimates presented by Sundaram (2008), about 5 to 7% of adult regular wage workers in various categories of manufacturing enterprises belonged to poor households in 2004–5, while the corresponding figure for adult casual workers was in the range of 17 to 27%. Casual workers not only get a significantly lower wage, they are also deprived of various benefits and social security.6Papola (2008) notes that organized sector workers employed on a nonregular and contract basis do not enjoy social security cover under different legislative provisions, such as the Employees' Provident Fund Act 1952, the Employees' State Insurance Act 1948, the Workmen's Compensation Act 1923, and the Maternity Benefit Act 1961. He also observes that “increasing casualisation implies not only increase in vulnerability in terms of employment and earnings, but also means that a larger proportion of workers have neither social protection nor productive resources to fall back upon, as most casual laborers are without assets” (p. 16).

What are the reasons for the growing informalization of industrial labor in India? Two possible causes that come to mind are: labor market rigidities and increasing competition, particularly competition from imports. Much has already been written on the effects of labor market rigidities on industrial employment in India, and there is a view that labor market rigidities are responsible for “jobless growth” in organized manufacturing and hence the increasing use of contract and temporary workers. Many scholars (e.g., Datta 2003; Ramaswamy 2003; Sharma 2006; Gupta, Hasan, and Kumar 2008; and Ahsan and Pagés 2008) maintain that the use of contact workers provides a means of getting around the labor regulations, particularly the Industrial Disputes Act (IDA), and industrial enterprises have actually been adopting this means on a wide scale.7 There is, however, not much econometric evidence in support of the view that labor market rigidities are the prime cause or even an important cause of the increasing employment of contract and temporary workers. Maiti, Saha, and Sen (2009) and Sen, Saha, and Maiti (2010) present econometric evidence that indicate that stringent labor regulations have led to greater use of contract workers. As a measure of the degree of labor market regulation, they use the index of Besley and Burgess (2004). This index is based on the amendments made by different Indian states to the Industrial Disputes Act of 1947 in the postindependence period up to 1992, i.e., the period 1947 to 1992. They consider each amendment made by a state and code it as pro-worker (assigned a value of +1), pro-employer (assigned a value of −1) or neutral (assigned a value of 0). These scores are then cumulated for the period 1947 to 1992, and the cumulative score for each state is taken as an indicator of the degree of labor market regulation. This index has been updated by Ahsan and Pagés (2008) and Purfield (2006). In the two papers of Maity, Saha, and Sen mentioned above, the authors find that the Besley-Burgess index has a significant positive effect on the proportion of contract workers out of total workers. The Besley-Burgess index has, however, come under severe criticism from Bhattacharjea (2006, 2009), and therefore, it seems, one has to be cautious in interpreting econometric evidence based on this index.

There is a strong possibility that increased competition (particularly international competition) leads to the informalization of industrial labor since the lower wages of informal workers and the savings made on the expenditure of worker benefits help in reducing costs and thus improve competitiveness. Papola (2008, p. 1) writes: “Apprehensions have also been raised about the likelihood of an increasing number of workers getting employed in relatively poor conditions of work, on low wages and without social security, as a result of the employers ‘pursuit of cost reduction’ in order to remain competitive. . . .” A similar view on the effect of competition on labor standards has been expressed by other researchers (e.g., Schmidt 2005).8 Econometric evidence for India on the issue under discussion is, however, rather limited, and the scant evidence that is available is mixed. Sen, Saha, and Maiti (2010) have analyzed econometrically the effect of trade on the use of contract labor and have found a significant positive effect of import penetration. They have used state-industry-year panel data for the period 1998–99 to 2004–5. Pradhan (2006), on the other hand, finds a negative effect of import penetration on use of contract labor. He has estimated a multiple regression equation to explain the ratio of contract workers to regular workers in India's organized manufacturing using pooled cross-industry data for three years, 1999–2000 to 2001–2. Clearly, the estimates obtained in these two studies point in opposite directions.

The aim of this paper is to assess whether, and how far, the labor market rigidities and import competition have caused informalization of industrial labor in India. The unit level data of the 61st round NSS employment–unemployment survey, 2004–5, have been used for the analysis supplemented by such data for 1999–2000. From the survey data, it is possible to segregate the workers into: (a) self-employed; (b) regular wage workers; and (c) casual workers. Since individuals with informal jobs and those with formal jobs cannot easily be segregated in the unit records, the analysis focuses on the regular worker–casual worker dichotomy. An econometric model is estimated for the analysis, which aims to assess the influence of individual characteristics such as education and experience, and the influence of import competition and labor market reforms on the probability of an individual being in a casual worker's job rather than a regular wage worker's job.

The rest of the paper is organized as follows. Section II examines the share of casual workers in different segments and industries of Indian manufacturing, followed by a comparative study of the characteristics of casual workers and regular wage workers in organized and unorganized manufacturing. Section III discusses the econometric model and the construction of variables. Section IV presents and discusses the estimates of the model. Section V goes into the robustness of the econometric estimates. Section VI summarizes and concludes the paper.

II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

A detailed analysis of casual labor in manufacturing in 2004–5 shows that while a little over one third (35%) of casual labor is in the organized manufacturing sector, around two-thirds are in the unorganized manufacturing sector. Thus, casual labor may be deemed to be more of an unorganized sector phenomenon. It is also mainly concentrated in rural areas (64%) and among males (74%). Most (71.5%) of the casual workers in manufacturing are also educated only up to primary class as compared to 37% of regular wage workers. This is discussed further later in this section.

The distribution of employment in the manufacturing sector reveals (Table 1) that while self-employed workers constitute barely 5% of the organized manufacturing sector, their share is more than two-thirds in the unorganized manufacturing sector. The reverse is the case for regular wage workers who dominate organized manufacturing. Though the share of casual labor in total employment is relatively higher in organized manufacturing, the absolute number of casual workers employed in the unorganized sector far exceeds that in the organized sector. While casual workers number only 2.9 million in organized manufacturing, there are 5.2 million in unorganized manufacturing. While the number of regular workers is three times that of casual workers in organized manufacturing, the absolute number of regular workers in organized manufacturing is almost equal to that in unorganized manufacturing.

Table 1.  Percentage Distribution of Employment in the Manufacturing Sector in 2004–5
StatusOrganizedUnorganizedTotal (Organized and Unorganized Combined)
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

Self-employed5.5967.6450.52
Regular wage worker70.7616.2331.18
Casual labor23.6416.1318.31
Total100100100

The distribution of casual workers is not uniform across all industries of the manufacturing sector. This is seen in Table 2, which shows the status-wide distribution of workers in various two-digit national industrial classification (NIC) groups of manufacturing, separately for organized and unorganized manufacturing.

Table 2.  Status-Wide Distribution of Workers by Industry in Organized and Unorganized Manufacturing, 2004–5
Two-Digit NIC CodeOrganized ManufacturingUnorganized Manufacturing
Self-EmployedRegular Wage WorkerCasual LaborTotal WorkersSelf-EmployedRegular Wage WorkerCasual LaborTotal Workers
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: NIC codes: 15–Food products and beverages; 16–Tobacco and related products; 17–Textiles products; 18–Wearing apparel, dressing, and dyeing of fur; 19–Leather and related products; 20–Wood and wood products; 21–Paper and paper products; 22–Publishing, printing, and related activities; 23–Coke, petroleum products, and nuclear fuel; 24–Chemicals and chemical products; 25–Rubber and plastic products; 26–Nonmetallic mineral products; 27–Basic metals; 28–Fabricated metal products; 29–Machinery and equipment n.e.c.; 30–Office, accounting, and computing machinery; 31–Electrical machinery, and apparatus, n.e.c.; 32–Radio, television, and communication equipments; 33–Medical, precision, and optical instruments; 34–Motor vehicles, trailers, and semi-trailers; 35–Other transport equipments; 36–Furniture and other manufacturing n.e.c.; and 37–Recycling.

154.2859.3736.3510067.8017.1015.10100
1615.3161.8422.8510090.103.906.00100
175.1473.7621.1010070.2812.7716.94100
185.6584.1210.2410078.4013.757.85100
199.7678.6911.5410055.4128.3516.24100
2017.8051.1231.0810083.084.8912.03100
210.8285.3213.8610067.2623.958.79100
2211.4785.383.1510041.6551.956.40100
230.6292.177.2210048.370.8650.77100
244.0779.1216.8110047.5930.4321.98100
2511.4579.798.7610023.7262.9913.29100
263.5231.1965.2910047.625.9246.46100
271.5183.8114.6810044.1534.1621.70100
284.0673.4422.5110048.2636.5815.16100
295.1990.714.1010046.9438.1714.89100
300.5299.480.0010048.3247.963.72100
3110.5282.966.5210052.0641.286.66100
323.0888.698.231009.0283.967.02100
332.7394.442.8310016.3469.9213.74100
340.0589.4310.5110035.5949.2915.11100
350.7994.404.8110023.9867.038.99100
365.4854.1040.4210065.5419.8614.60100
370.000.0010010076.913.1619.94100
Total5.5970.7623.6410067.6416.2316.13100

In the organized manufacturing sector, casual labor employment is relatively high (more than half of the workers) in only two industry groups, namely, manufacturing of other nonmetallic mineral products (26), and recycling (37). On the other hand, regular workers comprise more than three-fourths of the workers employed in 15 out of 23 two-digit manufacturing groups. These groups are 18, 19, 21–25, 27, and 29–35. But in the unorganized manufacturing sector, self-employed workers dominate in a majority of industry groups; regular wage workers are more than 50% only in five out of 23 two-digit manufacturing groups; and the proportion of casual labor varies mostly in the range of 3% to 22%. There are two industry groups in unorganized manufacturing, 23 (coke, petroleum products, and nuclear fuel) and 26 (nonmetallic mineral products), in which the proportion of casual workers is relatively high, 51% and 46%, respectively.

Attention may be drawn to the fact that estimates of casual workers obtained for organized manufacturing using NSS data may not properly cover contract workers. At the aggregate level, based on ASI data, the proportion of contract workers out of the total number of persons employed in organized manufacturing is 21% for 2004–5. This is lower than the proportion of casual workers, 24%, as shown in Table 2. However, when estimates for two-digit industries are considered, significant differences are found between the proportion of contract workers based on ASI data and the proportion of casual workers based on NSS data. The cross-industry correlation coefficient between the proportion of casual workers and the proportion of contract workers is only 0.20. Evidently, there is considerable dissimilarly between the proportion of casual workers and contract workers across industries.9 Why such dissimilarities arise is unclear. Perhaps definitions adopted by data collection agencies differ. It is also possible that a large number of contract workers getting regular wages is included in the regular wage worker category.

It is evident from the results of 2004–5 that the education level of casual workers is lower than that of regular workers. The same also applies to some extent to experience. The differences with respect to experience (proxied by age) are shown in Table 3. It is clear that more than 70% of the workers—both casual and regular—have relatively less experience (lower age), and as they gain experience there are more chances of being in the organized sector and then as a regular wage worker. Similarly, in Table 4, it can be seen that the level of education is higher for regular wage workers both in the organized and unorganized manufacturing sectors. As the level of education increases the proportion of regular wage workers becomes higher than that of casual labor. Furthermore, 67% of the casual workers in organized manufacturing and 74% of the casual workers in unorganized manufacturing are literate only up to primary level, whereas the corresponding proportions are 30% and 48% for regular salaried workers engaged in organized and unorganized manufacturing, respectively. The implication of the differences in education and experience between casual and regular wage workers is that with growing informalization, the average quality of labor declines, with possible adverse impacts on labor productivity.

Table 3.  Percentage Distribution of Regular and Casual Employment in Organized and Unorganized Manufacturing Sector by Age Categories
Age (Years)Organized ManufacturingUnorganized Manufacturing
Regular Wage WorkersCasual LaborRegular Wage WorkersCasual Labor
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

15–2526.9342.0140.8138.09
26–3534.6627.0833.1332.46
36–4522.3820.9717.1817.19
46–6015.298.837.4710.27
Above 600.741.111.411.99
Total100100100100
Table 4.  Percentage Distribution of Regular and Casual Employment in Organized and Unorganized Manufacturing Sector by Education Categories
Education LevelOrganized ManufacturingUnorganized Manufacturing
Regular Wage WorkersCasual LaborRegular Wage WorkersCasual Labor
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

Not literate10.6135.5516.2636.44
Literate below primary6.3411.6411.5716.24
Literate–primary13.1819.7020.1121.38
Literate–middle19.5522.7724.7116.56
Literate–secondary16.137.4114.116.60
Literate–higher secondary8.921.485.011.66
Literate–diploma/certificate course8.041.002.510.74
Literate–graduate14.340.434.770.38
Literate–postgraduate and above2.890.010.960.00
Total100100100100

III. MODEL AND VARIABLES

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

The econometric model used for the analysis explains the status of a worker: regular wage worker versus casual worker. The third category, namely self-employed worker, is excluded from the analysis (to keep the analysis and interpretation of results simple). The dependent variable SW takes two values: 0 for regular worker and 1 for casual worker. All workers (other than self-employed) engaged in manufacturing activities are considered in the sample. The unit level data from the 61st round employment–unemployment survey of the NSS for 2004–5 are used.10

Since the variable to be explained is dichotomous, the logit model is applied for the econometric analysis and it is estimated by the maximum likelihood method.11 The descriptive statistics for the dependent and explanatory variables are reported in Appendix Tables 5 and 6. The observations are for the individuals surveyed, confined to those in manufacturing. The observation-wise multipliers are used as weights in the estimation.12 The model is specified as:

  • image(1)

In this equation, education, age,13 and sex refer to the individual in question. For capturing education, a set of dummy variables is used, and is explained later. A dummy variable for the sex of the respondent is used taking a value of 1 for male and 0 for female. In the same way, the dummy variable for Scheduled Caste/Scheduled Tribe (SC/ST) is used; if the individual belongs to SC/ST (socially disadvantaged groups), the variable takes a value one of 1, and 0 otherwise. ORG is a dummy variable which takes a value of 1 if the individual is working in the organized sector enterprises, and 0 otherwise. The aim is to see if the probability of being in a regular wage job is higher if one is employed in the organized sector than in the unorganized sector. The rural–urban dummy variable is constructed in a similar way. It takes a value of 1 if the individual is residing in an urban area and 0 otherwise. In this case, the purpose is to assess if the probability of being in a regular wage job is greater if one is residing in an urban area or a rural area.

MPR is the import penetration ratio of the industry to which the individual belongs. For different input–output sectors, the import penetration ratios have been computed from the input–output table for 2003–4 prepared by the Central Statistics Office (CSO), Government of India. MPR for each input–output sector is obtained as: Imports/(domestic production + imports – exports). These have been mapped into three-digit National Industrial Classifications (NIC). Then, for each observation (i.e., each individual), the industry affiliation is used to set the values of MPR for that individual.14

LMR stands for labor market reforms index. This has been taken from Dougherty (2008). This index was formed by Dougherty on the basis of a state-level survey undertaken sometime between the mid- and later 2000s. The survey covered eight major areas of labor law, identifying 50 specific subjects of possible reform, many of which could be implemented by administrative procedure rather than through formal amendments to laws. The eight areas covered in the index are the Industrial Disputes Act (IDA), Factories Act, State Shops and Commercial Establishments Acts, Contract Labor Act, the role of inspectors, the maintenance of registers, the filing of returns, and union representation. State government officials were the respondents. Each state has been given a score reflecting the extent of reforms undertaken. Some examples of the reforms undertaken by states that are included in the index are: (1) contract worker allowed to work in noncore activities; (2) prior authorization needed for surprise inspection; (3) provision of self-certification by the firm in filing of returns to government agencies; (4) renewal of license under Factories Act allowed to be done for more than one year at a time; and (5) more than 21 days notice has to be given under Section 9A of the Industrial Disputes Act (this section relates to the issue of how labor is utilized within the enterprise and the purpose of the reform is curb outside interference in this matter). Further details are provided in Dougherty (2008).15 The index is available for 21 different states.16 Information on the state in which the individual is residing has been used to set the LMR index value for each individual.

To capture the effect of education on worker status, a set of dummy variables have been used for different levels of education. Nine levels of education have been considered. These are: not literate; literate below primary; literate–primary; literate–middle; literate–secondary; literate–higher secondary; literate–diploma/certificate course; literate–graduate; and literate–postgraduate and above. The base category is “not literate.” An alternative specification of the model employed involves dividing the sample into three subsamples according to education level and estimating the model separately for the three subsamples after dropping the education variable. The three levels corresponding to the three subsamples are: (1) education up to primary; (2) education beyond primary and up to higher secondary; and (3) education beyond higher secondary. This specification has the advantage that the effects of import competition, labor market reforms, and other explanatory variables are allowed to differ according to level of education.

The main hypotheses to be tested are that import competition as reflected in import penetration ratio induces greater informalization of industrial labor, and labor market reforms help in generating more regular wage employment. Another hypothesis to be tested is that the probability of being in a regular wage job goes up with the level of one's education and experience. Such a relationship is expected because the substitution possibilities between a regular worker and a casual worker are likely to be greater for simple jobs than for jobs requiring a higher level of skill/education/training. A somewhat similar argument can be made for experience.

IV. MODEL RESULTS

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

Table 5 presents the estimates of the model described in Section III above. A significant negative coefficient is found for labor market reforms index, as hypothesized. As regards import penetration ratio, the coefficient is positive as hypothesized, but not statistically significant. The results indicate that labor market reforms encourage industrial enterprises to provide regular jobs rather than temporary or contractual jobs. However, there is insufficient evidence to infer confidently that import competition tends to raise casual labor employment.

Table 5.  Model Estimates, Explaining Dichotomy between Regular and Casual Workers
Explanatory VariableCoefficientt-ratioMarginal Effect
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: * and *** represent statistical significance at the 10% and 1% level, respectively.

Import penetration ratio0.1450.700.0302
Labor market reforms index−0.023***−6.64−0.0049
Organized sector dummy−0.853***−12.24−0.177
Literate below primary−0.469***−3.97−0.090
Literate–primary−0.718***−7.13−0.134
Literate–middle−1.052***−9.89−0.188
Literate–secondary−1.595***−12.92−0.243
Literate–higher secondary−2.151***−11.73−0.267
Literate–diploma/certificate course−2.576***−10.25−0.283
Literate–graduate−3.735***−9.38−0.338
Literate–postgraduate and above−6.731***−9.26−0.313
Age−0.015***−5.03−0.0031
Sex (1–male, 0–female)−0.407***−4.45−0.088
SC/ST dummy0.145*1.900.031
Urban area dummy−1.155***−16.67−0.243
Constant3.13714.49 
No. of observations10,923  
Pseudo R-squared0.222  
LR chi-squared (15)1,084.9  
Prob. = 0.000

As hypothesized, education is found to have a negative relationship with the probability of being in a casual job (rather than a regular job), i.e., the higher the level of education of an individual, the lower the probability of his/her being in a casual job. The probability of being in a casual job for individuals in different education categories is depicted in Figure 1. This has been prepared by taking all other explanatory variables at their sample mean. The negative relationship between education level and the probability of being in a casual job is clearly visible. While for an illiterate person the probability of being in a casual job is about 65%, for category 8 (literate–graduate) and category 9 (literate–postgraduate and above) the probability is only about 4% and 0.2%, respectively.

image

Figure 1. Probability of Being in a Casual Job Rather Than a Regular Job, by Education Category. Source: Prepared on the basis of results reported in Table 5. The education categories are: 1–Not literate; 2–Literate below primary; 3–Literate–primary; 4–Literate–middle; 5–Literate–secondary; 6–Literate–higher secondary; 7–Literate–diploma/certificate course; 8–Literate–graduate; and 9–Literate–postgraduate and above.

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Turning to the results obtained for other explanatory variables, the estimated coefficient is significantly negative for age and for organized sector, urban areas and sex dummy variables, and positive for the SC/ST dummy variable. It may accordingly be inferred that the probability of being in a casual job is relatively lower if an individual is employed in the organized sector or is residing in an urban area or both; and the probability of being in a casual job is higher if the individual belongs to the SC/ST categories.

The negative coefficient of age implies that the probability of being in a regular job goes up with age. This is arguably also the effect of experience.

The coefficient of the dummy variable for the sex of the respondent indicates that, other things remaining the same, males are more likely to be in a regular job than females. The difference in probability is about 9 percentage points.

Although the coefficient of the import penetration variable is positive in Table 5, interindustry differences (at three-digit industry level) in import penetration ratio and the proportion of casual workers presented in Figure 2 show a negative correlation (especially when an outlier is removed). By contrast, interstate differences in labor market reforms index does bear a negative relationship with the proportion of casual workers in manufacturing,17 consistent with results of the model presented in Table 5. The correlation coefficient is −0.29. The scatter plot is shown in Figure 3.

image

Figure 2. Import Penetration Ratio and the Use of Casual Labor in Different Three-Digit Industries, 2004–5. Source: Import penetration ratios have been computed from input–output tables for 2003–4. The share of casual workers in various three-digit manufacturing industries has been computed from unit-level NSS 61st round survey data.

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image

Figure 3. State's Labor Reform Score and Share of Casual Workers in Total Casual and Regular Workers. Source: Labor market reforms index for states has been taken from Dougherty (2008) (available for 21 states). The share of casual workers in various states has been computed from unit-level NSS 61st round survey data.

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The finding of a negative correlation between the level of import penetration and the use of casual labor across various three-digit industries as depicted in Figure 2 is at variance with the estimated positive coefficient of the import penetration variable in results of the logit model estimation presented in Table 5. Therefore, based on these two pieces of empirical evidence, it is not possible to reach any definite conclusion on the effect of import competition.18 The estimates of the logit model for different educational categories thus come in handy for gaining a better understanding of the effect of import competition. The results are presented in Table 6 and suggest that the effect of import competition on workers with low education differs from that on workers with relatively higher education. It appears from the results that, for workers with education beyond primary level, import competition tends to reduce the probability of their being in a regular job. Such effect is, however, not found for workers with education up to primary level. Why import competition causes informalization among relatively more educated industrial workers is a pertinent question. The explanation possibly lies in the fact that cost saving through casualization is relatively greater for workers with a relatively higher education. The wage gap between regular worker and casual worker is relatively greater for workers with education higher than primary than that for worker with education up to primary.

Table 6.  Model Estimates, Explaining Dichotomy between Regular and Casual Workers, by Educational Category
Explanatory VariableEducational Categories
Up to PrimaryBeyond Primary and up to Higher SecondaryBeyond Higher Secondary
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: t-ratios in parentheses.

  3. *, **, and *** represent statistical significance at the 10%, 5%, and 1% level, respectively.

Import penetration ratio−0.592**0.809***0.979
(−2.41)(2.59)(0.73)
Labor market reforms index−0.021***−0.023***−0.042*
(−5.05)(−3.53)(−1.84)
Organized sector dummy−0.867***−0.859***−1.771***
(−9.94)(−7.63)(−3.66)
Age−0.004−0.029***−0.071**
(−1.37)(−4.77)(−2.48)
Sex (1–male, 0–female)−0.356***−0.690***−0.157
(−3.68)(−3.65)(−0.29)
SC/ST dummy0.1110.312**0.594
(1.20)(2.41)(0.85)
Urban area dummy−1.283***−1.094***−0.500
(−14.73)(−9.56)(−1.20)
Constant2.4582.2822.125
(10.41)(6.39)(1.88)
No. of observations5,2944,3041,325
Pseudo R-squared0.1090.1110.165
LR chi-squared (7)355.3205.265.1
Prob. = 0.000Prob. = 0.000Prob. = 0.000

It is observed that in the model estimates for workers with education above higher secondary, the coefficient of import penetration is positive but not statistically significant. This category includes workers with high levels of education, for example, those with engineering degrees who obviously cannot be replaced by casual workers. This seems to be the reason for the statistical insignificance of the coefficient. It appears, therefore, that casualization of labor caused by import competition is relatively greater for workers belonging to the education category “above primary and up to higher secondary” than for workers with education above higher secondary, especially those with high levels of education.

The results for explanatory variables other than import penetration ratio are similar to that presented in Table 5, which utilizes the entire sample. In particular, it may be noted that the coefficient of labor market reforms index is significantly negative in the estimates for all three subsamples. However, some interesting patterns are observed in Table 6. The coefficient of the SC/ST dummy variable in the model estimate for the education category “beyond higher secondary” is over five times the coefficient in the model estimate for the education category “up to primary.” The opposite pattern is observed for the coefficient of urbanization dummy variable. It should be pointed out here that the estimated marginal effect for SC/ST dummy is not much different between the education categories “up to primary” and “beyond higher secondary” (Figure 4). By contrast, the rural–urban difference in the probability of finding a regular job is far greater for individuals with education up to primary level than for individuals with education above higher secondary. Similarly, the male–female difference in the probability of finding a regular job declines with the level of education.

image

Figure 4. Marginal Effects, by Education Level. Source: Same as for Table 6.

Download figure to PowerPoint

It may be mentioned in passing that the use of a multinomial logit model instead of a logit model for the analysis presented in Table 6 does not change the results qualitatively (these results are reported in Appendix Tables 2–4). The labor market reforms index has a negative and statistically significant coefficient for all three educational groups. The coefficient of the import penetration variable is positive and statistically significant only in the model estimate for individuals with education above primary and up to higher secondary.

V. ROBUSTNESS OF ECONOMETRIC ESTIMATES

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

To check the robustness of the econometric estimates, the following changes have been made. First, instead of using data from only one survey, data from two surveys, 1999–2000 and 2004–5 have been pooled.19 It is not possible to pool data for earlier years, as data on some of the explanatory variables are not available. Since data from two surveys are used, a dummy variable for the year of survey is introduced, taking value 1 for observations belonging to the 2004–5 survey and 0 for observations belonging to the 1999–2000 survey. MPR for observations for 1999–2000 is based on the input–output table for 1998–99. Second, two new explanatory variables have been introduced in the model. These are (a) the growth of the industry to which the respondent belongs; and (b) an index of economic freedom for the state to which the respondent belongs. Growth rate in real output in different input–output sectors belonging to manufacturing has been computed for the periods 1993–94 to 1998–99 and 1998–99 to 2003–4. The computed growth rates for input–output sectors are then mapped into three-digit industry groups, as in the case of MPR. The growth rates computed for the period 1993–94 to 1998–99 are used for the observations for 1999–2000 and the growth rates computed for the period 1998–99 to 2003–4 are used for the observations for 2004–5. However, for LMR, only one set of values is available. Hence, the same set of values is used for the observations for 1999–2000 as well as 2004–5. The same applies to the index of economic freedom which relates to the year 2005.

While LMR is expected to capture in the estimated econometric model the effect of labor market reforms on work status, LMR being the only variable varying across states may capture additionally the influence of other state-specific factors (e.g., policies, institutions, etc.) which are correlated both with the explained variable and with LMR. The introduction of the index of economic freedom is intended to control for interstate differences in policies, institutions, etc. The indices of economic freedom for different Indian states in 2005 were prepared by Debroy, Bhandari, and Aiyar (2011). This index is based on three parameters: size of government; legal structure and security of property rights; and regulation of credit, labor, and business. Under each of these, there are subthemes. The methodology applied by Debroy and associates follows the methodology developed by the Fraser Institute for “Economic Freedom of the World.” This has been modified to suit the conditions in, and data availability for, Indian states. The methodological details are available in the report of Debroy et al., and hence not discussed here, except to point out that the labor-related indicators included in the index are: (a) the average wage of unskilled workers as a ratio of the minimum notified agricultural wage; and (b) the inverse of man-days lost in strikes and lockouts as a ratio of the total number of industrial workers.

A similar motive is behind the introduction of the growth of industries as an explanatory variable. The aim is to control for other variables which are correlated both with the explained variable and with MPR. Besides adding the growth of industries as an explanatory variable, an interactive term involving the output growth rate and LMR has been included in the econometric model. The aim is to assess if the impact of labor market reforms on work status differs between fast growing and slow growing industries.

The results are reported in Table 7. The first column shows the estimates of the Logit model. The second column shows the estimates of the Probit model. The estimates based on the Probit model are very similar to those based on the Logit model. In columns (3) and (4), the estimates of the Logit model are presented with a different option taken for the estimation of the variance–covariance matrix. The parameter estimates in columns (3) and (4) are the same as those in column (1), but the estimated standard errors and hence t-ratios are different. The estimates of the variance–covariance matrix in columns (3) and (4) are based on clusters. While the estimates in column (1) are based on the assumption that the observations are independent, the estimates in columns (3) and (4) allow for intra-group correlation. Column (3) uses state-wise clustering, while column (4) uses industry-wise clustering.

Table 7.  Model Estimates, Explaining Dichotomy between Regular and Casual Workers, Pooled Data from 1999–2000 and 2004–5 Surveys
Explanatory VariableLOGITPROBITLOGIT with CLUSTER (STATE)LOGIT with CLUSTER (INDUSTRY)
Coefficientt-ratioCoefficientt-ratioCoefficientt-ratioCoefficientt-ratio
  1. *, **, and *** represent statistical significance at the 10%, 5%, and 1% level, respectively.

Import penetration ratio0.0960.660.0690.800.0960.180.0960.27
Labor market reforms index (LMR)−0.022***−7.49−0.013***−7.58−0.022*−1.69−0.022**−2.03
Organized sector dummy−0.877***−19.08−0.525***−19.44−0.877***−9.34−0.877***−6.08
Growth rate of industry's output (GRI)−5.782***−4.86−3.476***−4.90−5.782**−2.36−5.782*−1.94
GRI*LMR0.134***4.780.080***4.820.134**2.190.134*1.83
Economic freedom score−0.556**−2.08−0.326**−2.06−0.556−0.44−0.556−0.62
Time (0 = 1999, 1 = 2004)−0.004−0.09−0.002−0.09−0.004−0.03−0.004−0.07
Literate below primary−0.408***−5.26−0.253***−5.41−0.409***−4.01−0.409***−2.90
Literate–primary−0.656***−9.79−0.403***−9.93−0.657***−5.65−0.657***−3.73
Literate–middle−0.925***−13.79−0.563***−13.92−0.925***−5.73−0.925***−3.96
Literate–secondary−1.496***−19.05−0.911***−19.70−1.496***−11.67−1.496***−5.82
Literate–higher secondary−2.188***−16.10−1.275***−17.88−2.188***−14.08−2.188***−7.20
Literate–above higher secondary−3.072***−19.60−1.707***−22.70−3.072***−20.72−3.072***−12.05
Age−0.013***−7.40−0.008***−7.65−0.014***−3.84−0.014***−5.37
Sex (0–female, 1–male)−0.371***−6.23−0.223***−6.23−0.371*−1.76−0.371−1.51
SC/ST dummy0.313***5.810.191***5.940.313*1.730.313*1.74
Urban area dummy−1.112***−24.71−0.667***−24.95−1.117***−8.42−1.117***−6.66
Constant3.14216.881.89117.193.1423.603.1426.17
No. of observations26,557 26,557 26,557 26,557 
Pseudo R20.213 0.214 0.213 0.213 
LR χ22,344.31 2,747.06 4,927.00 1,391.42 
 Prob. = 0.00 Prob. = 0.00 Prob. = 0.00 Prob. = 0.00 

It can be seen from Table 7 that the results are broadly in line with those presented in Table 5, which is based on data for one year and uses a smaller number of explanatory variables. In particular, it may be noted that the coefficient of the labor market reforms variable is negative and statistically significant even after adding new explanatory variables, as well as adopting a different method of computing the variance–covariance matrix. The impact of age, education, male dummy, urban dummy, organized sector dummy, and SC/ST dummy remains, by and large, the same in Table 7 as in Table 5.

When the sample is divided into three parts according to the level of education of the respondent and the model is estimated separately, the coefficient of the import penetration variable is found to be positive and statistically significant for respondents with education above primary but up to higher secondary. For respondents with a higher level of education, the coefficient is positive but not statistically significant. Thus, these results, shown in Table 8, are quite similar to those in Table 6 even though the growth rate of industries has been included as an explanatory variable in the model for controlling for some of the industry-specific factors.

Table 8.  Model Estimates, Explaining Dichotomy between Regular and Casual Workers, by Educational Category, Pooled Data from 1999–2000 and 2004–5 Surveys
Explanatory VariableUp to PrimaryBeyond Primary and up to Higher SecondaryBeyond Higher Secondary
Coefficientt-ratioCoefficientt-ratioCoefficientt-ratio
  1. Note: * and *** represent statistical significance at the 10% and 1% level, respectively.

Import penetration ratio−0.504***−2.730.587***2.720.8980.91
Labor market reforms index (LMR)−0.021***−6.03−0.018***−3.57−0.030−1.49
Organized sector dummy−0.855***−14.46−0.929***−12.60−1.407***−4.01
Growth rate of industry's ouput (GRI)−6.534***−4.59−5.532***−2.67−10.867−1.39
GRI*LMR0.172***5.260.085*1.770.1951.16
Economic freedom score−1.539***−4.370.1880.45−1.197−0.63
Time (0 = 1999, 1 = 2004)0.0270.48−0.039−0.55−0.241−0.76
Age−0.003−1.20−0.035***−8.98−0.065***−2.70
Sex (0–female, 1–male)−0.436***−6.82−0.519***−4.150.0230.04
SC/ST dummy0.335***5.120.329***3.421.220***2.53
Urban area dummy−1.19***−20.43−1.046***−14.37−1.227***−3.68
Constant2.85412.842.2226.882.3941.78
No. of observations12,600 11,108 2,849 
Pseudo R20.107 0.1207 0.1808 
LR χ2741.34 524.76 96.38 
Prob.0.00 0.00 0.00 

The negative coefficient of the output growth variable observed in all the model estimates suggests that a rapid growth in industries helps create regular jobs rather than casual jobs. The positive coefficient of the interactive term involving output growth rate and LMR possibly means that if the growth rate of an industry is not high, labor reforms can help in creating regular jobs.

VI. CONCLUDING REMARKS

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

The results of the econometric analysis presented above indicate that intensification of import competition in the post-reform period has been responsible, to some extent, for the increasing informalization of labor in Indian manufacturing. This supports, to some extent, a view that globalization is leading to a “race to the bottom” in labor standards in developing countries. The results of economic analysis also indicate that labor market reforms tend to counter the tendency towards informalization and help in the creation of regular salaried/wage employment opportunities in manufacturing. One may be tempted to treat this as empirical evidence in support of the view that the labor market rigidities, particularly the Industrial Disputes Act is the prime cause, or an important cause, of increasing employment of contract and temporary workers. Such inference is, however, hasty since it is important to recognize that the sample used for the econometric analysis is dominated by hired workers engaged in unorganized manufacturing, and the Industrial Disputes Act does not apply to them. The same is possibly true for several other regulations which are held responsible for rigidities in the labor market. Why labor market reforms encourage unorganized manufacturing enterprises to substitute casual labor by regular wage labor is therefore a puzzle. One possibility is the index of labor market reforms constructed by Dougherty (2008) includes a large number of reform measures that can be taken by the state governments, and some of these impact both organized sector manufacturing enterprises and the unorganized sector manufacturing enterprises that have hired workers. Another possibility is that reform measures taken by state governments reflect the changes in the attitude of the state governments towards enterprises and the workers employed in the enterprises. The changes in attitude of the state governments have a significant bearing on the way business is done in the states, and in particular it encourages small enterprises to hire workers on a more regular basis. Needless to say, these are speculations on what could be the reasons for the observed, robust negative relationship between labor market reforms and the proportion of casual workers, and a more thorough investigation is needed. This is a matter that may be taken up in future research.

REFERENCES

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices
  • Ahsan, Ahmad, and Carmen Pagés. 2008. “Are All Labor Regulations Equal? Evidence from Indian Manufacturing.” IZA Discussion Paper no. 3394. Bonn: Institute for the Study of Labor.
  • Besley, Timothy J., and Robin Burgess. 2004. “Can Labor Regulation Hinder Economic Performance? Evidence from India. Quarterly Journal of Economics 119, no. 1: 91134.
  • Bhandari, Amit K., and Almas Heshmati. 2005. “Labour Use and Its Adjustment in Indian Manufacturing Industries. Global Economic Review 34, no. 3: 26190.
  • Bhandari, Amit K., and Almas Heshmati. 2006. “Wage Inequality and Job Insecurity among Permanent and Contract Workers in India: Evidence from Organized Manufacturing Industries.” IZA Discussion Paper no. 2097. Bonn: Institute for the Study of Labor.
  • Bhattacharjea, Aditya. 2006. “Labour Market Regulation and Industrial Performance in India: A Critical Review of the Empirical Evidence. Indian Journal of Labour Economics 49, no. 2: 21132.
  • Bhattacharjea, Aditya. 2009. “The Effects of Employment Protection Legislation on Indian Manufacturing.” CDDRL Working Paper no. 96. Stanford, Calif.: Center on Democracy, Development, and the Rule of Law, Freeman Spogli Institute for International Studies, Stanford University.
  • Datta, Ramesh C. 2003. “Labour Market—Social Institutions, Economic Reforms and Social Cost.” In Labour Market and Institution in India 1990s and Beyond, ed. Shuji Uchikawa. New Delhi: Manohar.
  • Debroy, Bibek; Laveesh Bhandari; and Swaminathan S. Anklesaria Aiyar. 2011. Economic Freedom of the States of India 2011. New Delhi: Academic Foundation. http://www.freetheworld.com/pdf/EconomicFreedomIndia-2011.pdf (accessed October 28, 2011).
  • Dougherty, Sean M. 2008. “Labour Regulation and Employment Dynamics at the State Level in India.” OECD Economics Department Working Paper. no. 624. Paris: OECD Publishing.
  • Gang, Ira N.; Kunal Sen; and Myeong-Su Yun. 2008. “Poverty in Rural India: Caste and Tribe. Review of Income and Wealth 54, no. 1: 5070.
  • Gupta, Poonam; Rana Hasan; and Utsav Kumar. 2008. “What Constrains Indian Manufacturing?” ERD Working Paper no. 119. Manila: Asian Development Bank.
  • Maiti, Dibyendu, and Arup Mitra. 2010. “Skills, Informality, and Development.” Working Paper no. 306. Delhi: Institute of Economic Growth, University Enclave.
  • Maiti, Dibyendu; Bibhas Saha; and Kunal Sen. 2009. “Liberalisation, Labour Legislation and Flexibility—Theory and Evidences in India.” Paper presented at International Conference on the Informal Sector in South Asia: Organizational Dynamics, Institutional Determinants, Interlinkages, and Development, Institute of Economic Growth, Delhi, July 27–28.
  • National Commission for Enterprises in the Unorganised Sector (NCEUS). 2009. “The Challenge of Employment in India: An Informal Economy Perspective.” I and II. New Delhi: National Commission for Enterprises in the Unorganised Sector.
  • National Sample Survey Organisation (NSSO). 2006. “Employment and Unemployment Situation in India 2004–05: NSS 61st Round.” Report no. 515, part I. New Delhi: National Sample Survey Organisation.
  • Papola, Trilok S. 2008. “Employment Challenge and Strategies in India.” ILO Asia Pacific Working Paper Series. New Delhi: Subregional Office for South Asia International Labour Organization.
  • Pradhan, Jaya Prakash. 2006. “How Does Trade, Foreign Investment, and Technology Affect Employment Patterns in Organized Indian Manufacturing?” Indian Journal of Labour Economics 49, no. 2: 24972.
  • Purfield, Catriona. 2006. “Mind the Gap—Is Economic Growth in India Leaving Some States Behind?” IMF Working Paper no. 06/103. Washington, D.C.: International Monetary Fund.
  • Ramaswamy, Krishnarajapet Vidyaranya 2003. “Liberalization, Outsourcing and Industrial Labour Markets in India: Some Preliminary Results.” In Labour Market and Institution in India: 1990s and Beyond, ed. Shuji Uchikawa. New Delhi: Manohar.
  • Schmidt, Johannes Dragsbaek. 2005. “Flexicurity, Casualisation and Informalisation of Global Labour Markets.” Working Paper no. 133. Denmark, Aalborg: Research Center on Development and International Relations (DIR), Aalborg University.
  • Sen, Kunal; Bibhas Saha; and Dibyendu Maiti. 2010. “Trade Openness, Labour Institutions and Flexibilisation: Theory and Evidence from India.” BWPI Working Paper no. 123. Manchester: Brooks World Poverty Institute, University of Manchester.
  • Sen, Sunanda, and Byasdeb Dasgupta. 2009. Unfreedom and Waged Work: Labour in India's Manufacturing Industry. New Delhi: Sage Publications India.
  • Sharma, Alakh N. 2006. “Flexibility, Employment and Labour Market Reforms in India. Economic and Political Weekly 41, no. 21: 207885.
  • Sundaram, Krishnamurthy 2008. “Employment, Wages and Poverty in the Non-Agricultural Sector: All-India, 2000–05. Economic and Political Weekly 43, no. 22: 919.
Footnotes
  • 1

    This estimate is based on Annual Survey of Industries (ASI) data.

  • 2

    They first estimate the total number of UPSS (usual principal and subsidiary status) workers in different activities for 1999–2000 and 2004–5 using NSSO (National Sample Survey Organisation) employment–unemployment surveys for these two years along with estimates of population, and then split the estimates of workers into formal and informal workers using NSSO results of the survey on informal sector workers in nonagricultural activities.

  • 3

    Note that these estimates are for the industrial sector, which has greater coverage than manufacturing. In particular, it seems, the construction industry is included in the industrial sector.

  • 4

    National Sample Survey Organisation (NSSO) (2006, p. 13) defines casual wage labor as “A person who was casually engaged in others' farm or non-farm enterprises (both household and non-household) and, in return, received wages according to the terms of the daily or periodic work contract.” A regular salaried/wage worker, on the other hand, is defined as the one who “receives salary or wages on a regular basis (i.e., not on the basis of daily or periodic renewal of work contract). This category includes not only persons getting time wage but also persons receiving piece wage or salary and paid apprentices, both full time and part-time.”

  • 5

    Sen and Dasgupta (2009) have undertaken a survey of industrial units in a large number of clusters in different parts of India (during 2004–5). The wages of casual workers were found to be significantly lower than those of permanent workers. In NOIDA, for instance, the permanent workers' average wage was about Rs4,760 per month, while that of casual workers was Rs2,480 per month. In Kolkata, the relevant figures were Rs4,820 and Rs1,970, respectively.

  • 6

    That the working conditions of casual workers in organized manufacturing are inferior to that of regular workers has been noted by Sundaram (2008, Table 4, p. 94).

  • 7

    Sharma (2006, p. 2081) writes: “contract labour has been one of the principal methods used by the employers to gain flexibility in the labour market. Thus, employers have been able to find ways to reduce the workforce even with the ‘restrictive’ provisions in place.” Similarly, Gupta, Hasan, and Kumar (2008, p. 7) write, “hiring contract workers can enable firms to get around many of the regulatory restrictions on adjusting employment levels, productions tasks, and others. . . .”Ahsan and Pagés (2008) note that contract labor has become a common way to deal with the problems posed by the labor regulation arising from the Industrial Disputes Act. The use of contract labor is found to have a favorable effect on employment in the econometric analysis undertaken by them. However, from the results obtained, they conclude that contract labor may be more effective at ameliorating the effects of regulations on output than on employment. At the same time, Ahsan and Pagés (2008) point out that while firms hire contract labor as a way to reduce wage and adjustment costs, the fact that contract workers are not covered by industrial dispute laws is probably an additional source of interest for employers. See also Bhattacharjea (2009) and Bhandari and Heshmati (2005, 2006) in this context.

  • 8

    Sen, Saha, and Maiti (2010) observe that while import competition may force firms to seek short-run efficiency by resorting to low-wage employment, firms mindful of long-run efficiency or concerned about quality improvement (a key issue for exporting firms) may invest in productivity improvement of the regular workers.

  • 9

    The implication is that conclusions drawn from an econometric study that utilizes inter-industry differences in the use of contract labor may be at variance with the conclusions drawn from a study that utilizes inter-industry differences in the use of casual labor.

  • 10

    NSSO in the 61st round employment–unemployment survey for 2004–5 collected data for a sample of 602,833 at all-India level. It calculated a multiplier for each sample observation, which it represents. These multipliers are used as weights to estimate an all India characteristic. Out of the total sample, only 244,849 belong to those who are UPSS workers and of this only 26,990 sample observations belong to manufacturing at all-India level. The sample actually used is lower than 26,990 because the analysis had to be confined to the states for which a measure of labor market reforms could be obtained. For the 21 states for which the labor market reforms index is available, the number of sample observations is 24,631.

  • 11

    As a check on the results obtained by the dichotomous model (logit), a multinomial logit model has been estimated in which regular worker, casual worker, and self-employed worker are taken as three categories. This is obviously a more general specification of the model. The results are reported in the Appendix Tables 1–4. The estimated equation for casual worker in the multinomial logit model matches well the estimated logit model corroborating the findings.

  • 12

    The model is estimated by the STATA statistical package. The “pweights” option is used for weighting, which uses sampling weights. Several earlier econometric studies using NSS unit level data have used the multipliers as weights for estimation of econometric models (e.g., Gang, Sen, and Yun 2008).

  • 13

    Age is taken as a proxy for experience.

  • 14

    Another variable which could have been included in the model is the export orientation of different industries (to reflect the competition faced by Indian firms in export markets). This, however, could not be done since the export intensity of different input–output sectors belonging to manufacturing is strongly positively correlated with the import penetration ratio. The correlation coefficient across 68 input–output sectors belonging to manufacturing is 0.63 (Input–Output Table for 2003–4). In a sense, therefore, the import penetration variable captures the level of trade competition faced by different industries.

  • 15

    The index is based on 50 specific subjects of possible reform. Each state is given a score reflecting the extent of reforms undertaken; the maximum possible score is 50 and the average score across states is 21. For the analysis presented here, the relative scores reported by Dougherty are used.

  • 16

    Since the index is not available for a number of states and union territories, these are excluded. However, all major states are included in the analysis.

  • 17

    There is a possibility that the industries with a relatively high share of regular wage workers had a large share of industrial employment in states for which the Dougherty index is high. This will lead to a high positive correlation between the share of regular wage worker and Dougherty index across states even if labor reforms have no significant effect on work status. To investigate this aspect, the states have been divided into two groups: those with LMR above average and those with LMR below average, and then the ratio of casual worker to (casual + regular) workers have been computed for each three-digit industry in the two groups of states. For a majority of industries (63% of cases), the proportion of casual workers (ratio mentioned above) is found to be higher in the groups of states that have low LMR, i.e., those in which there has been relatively less labor market reform. The paired sample t-test for comparing the means shows that the difference is statistically significant at the 5 % level.

  • 18

    Attention may be drawn here to the distinction between partial and total correlation. Thus, even if trade liberalization bears a positive relationship with the employment of casual workers, a simple correlation coefficient between the two variables may turn out to be negative (since the influence of other explanatory factors is not controlled for). It may be noted further that the coefficient of the import penetration variable is positive in the multinomial logit model (shown in the Appendix Table 1) as in the results of the logit model in Table 5.

  • 19

    An employment–unemployment survey with a large sample is conducted by the NSSO once every five years or so. Prior to 2004–5, such surveys were conducted in 1999–2000, 1993–94, 1987–88, 1983, and 1978–79.

Appendices

  1. Top of page
  2. Abstract
  3. I. INTRODUCTION
  4. II. CASUAL LABOR IN MANUFACTURING: RELATIVE SHARE AND CHARACTERISTICS
  5. III. MODEL AND VARIABLES
  6. IV. MODEL RESULTS
  7. V. ROBUSTNESS OF ECONOMETRIC ESTIMATES
  8. VI. CONCLUDING REMARKS
  9. REFERENCES
  10. Appendices

APPENDIX I RESULTS OF THE MULTINOMIAL LOGIT MODEL

Table APPENDIX TABLE1.  Model Estimates, Explaining Worker Status: Regular Workers, Casual Workers, and Self-Employed Workers
Explanatory VariableEquation for Casual WorkersEquation for Self-Employed Workers
Coefficientt-ratioCoefficientt-ratio
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: * and *** represent statistical significance at the 10% and 1% level, respectively.

Import penetration ratio0.2441.250.339*1.95
Labor market reforms index−0.019***−6.070.00090.30
Organized sector dummy−0.753***−11.08−3.951***−43.40
Literate below primary−0.416***−3.66−0.142−1.33
Literate–primary−0.648***−6.65−0.284***−2.97
Literate–middle−0.987***−9.81−0.390***−4.16
Literate–secondary−1.473***−12.09−0.489***−4.20
Literate–higher secondary−2.058***−11.24−0.171−1.13
Literate–diploma/certificate course−2.510***−10.05−1.26***−6.07
Literate–graduate−3.551***−8.970.0480.21
Literate–postgraduate and above−6.808***−9.36−0.599*−1.79
Age−0.008***−3.160.025***10.22
Sex (1–male, 0–female)−0.326***−3.84−1.535***−20.19
SC/ST dummy0.1151.57−0.335***−4.72
Urban area dummy−1.041***−15.47−1.169***−18.80
Constant2.50713.572.58214.67
No. of observations24,627   
Pseudo R20.285   
LR χ2 (30)3,062.5   
Prob. = 0.000
Table APPENDIX TABLE2.  Model Estimates, Explaining Worker Status: Regular Workers, Casual Workers, and Self-Employed Workers, Education up to Primary
Explanatory VariableEquation for Casual WorkersEquation for Self-Employed Workers
Coefficientt-ratioCoefficientt-ratio
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: * and *** represent statistical significance at the 10% and 1% level, respectively.

Import penetration ratio−0.463*−1.79−0.163−0.62
Labor market reforms index−0.015***−4.070.0020.56
Organized sector dummy−0.853***−9.85−4.463***−31.70
Age−0.002−0.540.023***7.61
Sex (1–male, 0–female)−0.380***−3.98−1.624***−17.27
SC/ST dummy0.1011.12−0.348***−3.84
Urban area dummy−1.252***−14.27−1.496***−17.31
Constant2.11910.202.79513.92
No. of observations14,185   
Pseudo R20.237   
LR χ2 (14)1,546.1   
Prob. = 0.000
Table APPENDIX TABLE3.  Model Estimates, Explaining Worker Status: Regular Workers, Casual Workers, and Self-Employed Workers, Education beyond Primary and up to Higher Secondary
Explanatory VariableEquation for Casual WorkersEquation for Self-Employed Workers
Coefficientt-ratioCoefficientt-ratio
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: **, and *** represent statistical significance at the 5% and 1% level, respectively.

Import penetration ratio0.829***2.850.604**2.48
Labor market reforms index−0.022***−3.650.0071.43
Organized sector dummy−0.808***−7.31−3.975***−29.16
Age−0.025***−4.370.030***6.87
Sex (1–male, 0–female)−0.619***−3.53−1.625***−12.15
SC/ST dummy0.273**2.19−0.328***−2.65
Urban area dummy−1.050***−9.25−0.987***−10.02
Constant2.0146.161.6996.07
No. of observations8,466   
Pseudo R20.257   
LR χ2 (14)1,069.2   
Prob. = 0.000
Table APPENDIX TABLE4.  Model Estimates, Explaining Worker Status: Regular Workers, Casual Workers, and Self-Employed Workers, Education beyond Higher Secondary
Explanatory variableEquation for Casual WorkersEquation for Self-Employed Workers
Coefficientt-ratioCoefficientt-ratio
  1. Source: Authors' computation based on unit records of NSS 61st round employment–unemployment survey data.

  2. Note: *, **, and *** represent statistical significance at the 10%, 5%, and 1% level, respectively.

Import penetration ratio0.5100.440.942**2.05
Labor market reforms index−0.041*−1.67−0.011−0.86
Organized sector dummy−1.709***−3.44−2.967***−12.85
Age−0.077***−2.620.048***5.13
Sex (1–male, 0–female)−0.105−0.20−0.310−1.13
SC/ST dummy0.5760.87−0.338−0.96
Urban area dummy−0.280−0.66−0.087−0.46
Constant2.0951.72−0.371−0.55
No. of observations1,976   
Pseudo R20.267   
LR χ2 (14)229.9   
Prob. = 0.000

APPENDIX II DESCRIPTIVE STATISTICS

Table APPENDIX TABLE5.  Descriptive Statistics, Data for 2004–5
VariableMeanStandard DeviationMinimumMaximum
SW (=0 if regular worker; =1 if casual)0.3750.48401
Import penetration ratio0.1280.20100.859
Labor market reforms index (LMR)42.36610.3112458
Organized sector dummy (=1 if working in organized sector, =0 otherwise)0.2450.43001
Growth rate of industry's output (GRI) (annual rate of growth)0.0470.099−0.1260.460
GRI*LMR2.0204.317−7.35126.699
Economic freedom score0.3990.0820.250.57
Age33.90413.041690
Sex (1–male, 0–female)0.7040.45701
SC/ST dummy (=1 if SC/ST, =0 if otherwise)0.2140.41001
Urban area dummy (=0 if rural and =1 if urban)0.5120.50001
Table APPENDIX TABLE6.  Descriptive Statistics for Pooled Data, 1999–2000 and 2004–5
VariableMeanStandard DeviationMinimumMaximum
SW(=0 if regular worker, =1 if casual)0.3550.47801
Import penetration ratio0.1200.16400.858
Labor market reforms index (LMR)42.73210.3492458
Organized sector dummy (=1 if working in organized sector, =0 otherwise)0.2500.43301
Growth rate of industry's output (GRI) (annual rate of growth)0.0600.087−0.1260.460
GRI*LMR2.5783.900−7.35126.699
Economic freedom score0.3970.0850.250.57
Age34.03613.165696
Sex (1–male, 0–female)0.7200.44901
SC/ST dummy (=1 if SC/ST, =0 if otherwise)0.1750.38001
Urban area dummy (=0 if rural and =1 if urban)0.5600.49601