Abstract
This paper extends previous work on the effect of the Bologna reform on university enrolment in Italy. The analysis considers more recent data and also attempts to disentangle the effect of the reform from the influence on enrolment exerted by time-varying confounding factors. The empirical findings consistently show that the “Bologna Process” had a positive impact on university participation, though the magnitude of this impact is smaller than previously concluded. One main reason for our lower estimates lies in the use of a difference-in-differences methodology, which allows us to control for the influence on enrolment exerted by unobserved factors that could have changed coincidentally at the same time as the reform.
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Notes
Jacobs and van der Ploeg (2006) discuss the potential benefits associated with this Anglos-Saxon system.
There is still relatively little information about the proportion of first-cycle degree holders who are willing to pursue a second-cycle degree. Crosier et al. (2007), analyzing data from many higher education institutions in Europe, find that only 22% of them reported that most of their students will enter into the labour market after successfully completing the first-cycle degree.
This is likely to be due to the fact that the reform has only been recently implemented.
Not being able to control for the secular cohort trend in enrolment before the reform may yield an imprecise estimate of the effect of the “Bologna Process” on university participation.
Although widening the time window has some advantages, it also increases the likelihood of introducing confounding factors in the analysis. However, several robustness and sensitivity checks are performed at the end of Sect. 5 to address this concern.
This issue is also raised by Cappellari and Lucifora (2009). They state that their estimates are preliminary and will need to be refined as more data become available (p. 647).
There are no admission standards, except for specialized disciplines such as medicine and architecture.
Given the low direct cost of university education in Italy (compared to other countries such as, for instance, the United States), tuition fees are unlikely to exert an influence on the enrolment decision even for individuals from less advantaged backgrounds. Previous studies (see for, instance, Cappellari and Lucifora, 2009) have not considered the impact of this factor.
Three geographical areas are considered: North, South and Centre. Unfortunately, we are unable to define areas of residence at a more disaggregated level (i.e. region) given that this information is unavailable in the 2004 and 2007 waves.
The datasets on 1995, 1998, 2001 and 2004 cohorts of high school leavers are matched with information on the average unemployment rate for the 1995–1997, 1998–2000, 2001–2003 and 2004–2006 periods, respectively. These data are taken from ISTAT Labour Force Survey.
Hence the conclusions of this study rely on the assumption that unobserved factors associated with both family background and university enrolment are constant over time.
See Cappellari and Lucifora (2009) for reasons for this exclusion.
The original sample sizes are: 18,443 (1995 cohort), 23,261 (1998 cohort), 20,407 (2001 cohort) and 25,880 (2004 cohort).
We report estimates in which standard errors are not clustered at cohort level. Nichols and Schaffer (2007) argue that with a small number of clusters (less than 50) inference using the cluster-robust estimator may be incorrect more often than not.
Results from a logit estimation are consistent with those reported in Table 2. The logit model indicates that, whilst individuals from poorer family backgrounds are 14.74% more likely to enroll at university in the post-reform period relative to the pre-reform period, the reform appears to have no statistically significant effect on the enrolment of individuals from richer family backgrounds.
It is worth to observe that our labour market indicator turns out to have a differential impact on people’s decision to attend university depending on their family background. Whilst unemployment rate is not found to be a significant determinant of university enrolment among high school leavers from more advantaged backgrounds, this variable is a strong predictor of university participation among those from poorer backgrounds.
Unfortunately, the survey does not provide information on parental income. Thus it is not possible to use this factor to define family background.
Managers, entrepreneurs, high-level civil servants and people with a professional occupation are included in this category.
Results from the new pre-program test indicate that there are no statistically significant differences in the enrolment trend among individuals from more and less advantaged backgrounds during the pre-reform period. These results are available from the author upon request.
The Laws 448/2001 and 9/1999 may have a different impact across our treatment and control groups. Whilst the first measure is likely to act as an incentive to enrol at university especially for people from less advantaged backgrounds, the second provision may encourage particularly individuals from richer backgrounds to attend university.
In an attempt to make the specification closer to that used by Cappellari and Lucifora (2009), the linear cohort trend and interactions between this and area of residence dummies have been excluded.
Although our before–after point estimates are significantly smaller than those reported by Cappellari and Lucifora (2009) in Columns (5) and (6) of Table 2, this can be mainly attributed to the different set of labour market controls employed. Our before–after point estimates (i.e. 0.086 and 0.089) from a basic specification (that includes only gender) are very similar to that reported by Cappellari and Lucifora (2009) in Column (1) of Table 2 (i.e. 0.09).
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I am grateful to two anonymous referees for their helpful comments. However, any errors or omissions remain my own responsibility.
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Di Pietro, G. The Bologna Process and widening participation in university education: new evidence from Italy. Empirica 39, 357–374 (2012). https://doi.org/10.1007/s10663-011-9172-5
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DOI: https://doi.org/10.1007/s10663-011-9172-5