· WEEK 1 Databases and SecurityLesson· Databases and Security.docx

· WEEK 1: Databases and SecurityLesson · Databases and Security Databases are in just about everything we use today. When you are performing any task, think to yourself, Does this involve a database in some way? As a daily process, communication occurs between people by many mediums, but there is no other medium more utilized than the large internetwork of computer systems we know as the Internet. When we look at some of the transactions that are performed on a daily basis, it is highly likely that there is a database involved. For example, if you open a web page to www.google.com and type a keyword in the textbox to search for, this process starts a series of searches through multiple databases. Another example is when searching for a book in the APUS library, this search is conducted using a database of books known as a catalog. so databases play an integral part in our daily lives; they store millions of pieces of data and more is collected each day (Basta, 2012). In recent years, we find that technology has expanded to the reaches of utilities and production environments. Many of the utilities we come to rely on so heavily, such as gas, oil and electric, have been tied into the networks we use today. This interconnection allows for many new innovations in keeping everything in working order, but at the same time it also presents some very real threats to security. In reality, an intruder could take down an entire electrical grid which would remove power to millions of customers. An article in CIO Insight gives a great perspective on this and other issues in security where databases play such an important role (CIOInsight, 2011). With the importance of securing the database infrastructure, we need to look at a multilayered approach to security. As can be seen in many security programs, multiple layers allow for strong security because it adds another roadblock that an

intruder has to bypass to get to these systems. This same approach leads us to begin with the foundation of security; the CIA Triad. It all begins with the most basic approach, computer security and moves forward from that point on. Below is a detailed description of the components of the CIA Triad from (Basta, 2012): · Confidentiality: For a system to provide confidentiality, it needs to do two things: ensure that information maintains its privacy by limiting authorized access to resources; block unauthorized access to resources. · Integrity: This refers to the efforts taken through policy, procedure, and design in order to create and maintain reliable, consistent, and complete information and systems. · Availability: This refers to the efforts taken through policy, procedures, and design to maintain the accessibility of resources on a network or within a database. These resources include, but are not limited to, data, applications, other databases, computers, servers, applications, files, drives, shares, and network access. Database Structure, Models and Management A database alone is just a single collection of data that has some organization based on the various groupings necessary for that data. In most cases, the database follows some model of organization, but by itself it is merely just a file filled with information. So what is needed is a way to bring all of these together, which allows us to introduce the database management system (DBMS). Quite simply, a DBMS is an application that is used to combine databases and allow the addition and modification of data held within a database. A DBMS also allows for added functionality to manipulate data in many ways. For example, going back to the Google search example, we can search for specific criteria, which then creates a search of databases within that criteria; this is also known as a query. A database is made up of several components that aid in the organization of data. The highest level is the table then moves on to records, columns, rows, fields, etc. You can see an

example in the text by reviewing figure 2.1. Database records usually need a starting point, such as an identifier to make the record unique from others. This can be done in many ways, but what this is known as in a database is a key. There are two major categories of a database key; the primary key and the foreign key but there are other keys that could be used. Below is the definition given by (Basta, 2012): · Primary Key: It is a best practice, but not necessary, to use keys that are meaningful to the data being stored. Examples of primary keys are employee ID numbers, student IDs, ISBNs, and Social Security numbers. · Foreign Key: A foreign key is a field within a table that contains a label that is used to build a relationship between two tables. Use Figure 2-2 to aid the discussion. · Other Keys: The use of the following keys depends on the administrator, the DBMS, and the database model within an environment: secondary or alternative key, candidate key, sort or control key, and alternate key. Keys help to map relationships in databases and make it easier to create queries that group data that is being searched. There are many database models to consider. A database model is a representation of the way data is stored. Note that the model for which a database is constructed also determines the way the data can be retrieved and manipulated. Below are examples of database models that are used. · Hierarchical Model · Network Model · Relational Database · Object-Oriented Databases References/Works Sited: Basta, A. and Zgola, M. (2012). Database Security, 1st Edition. Florence, KY. Delmar Cengage Learning. ISBN-10/13: 1435453905/9781435453906 CIOInsight (2011). Stronger Database Security Needed, Cyber Attacks Show. CIO Insight.

URL: /access/content/group/science-and-technology- common/ISSC/ISSC431/Reading-Materials/Stronger-Database- Security-Needed-Cyber-Attacks-Show.pdf Full Terms & Conditions of access and use can be found at https://www.tandfonline.com/action/journalInformation?journal Code=rfec20 Feminist Economics ISSN: 1354-5701 (Print) 1466-4372 (Online) Journal homepage: https://www.tandfonline.com/loi/rfec20 Gender Disparity in Education and the International Competition for Foreign Direct Investment Matthias Busse & Peter Nunnenkamp To cite this article: Matthias Busse & Peter Nunnenkamp (2009) Gender Disparity in Education and the International Competition for Foreign Direct Investment, Feminist Economics, 15:3, 61-90, DOI: 10.1080/13545700802528315 To link to this article: https://doi.org/10.1080/13545700802528315 Published online: 23 Jul 2009. Submit your article to this journal Article views: 481

View related articles Citing articles: 9 View citing articles https://www.tandfonline.com/action/journalInformation?journal Code=rfec20 https://www.tandfonline.com/loi/rfec20 https://www.tandfonline.com/action/showCitFormats?doi=10.10 80/13545700802528315 https://doi.org/10.1080/13545700802528315 https://www.tandfonline.com/action/authorSubmission?journalC ode=rfec20&show=instructions https://www.tandfonline.com/action/authorSubmission?journalC ode=rfec20&show=instructions https://www.tandfonline.com/doi/mlt/10.1080/13545700802528 315 https://www.tandfonline.com/doi/mlt/10.1080/13545700802528 315 https://www.tandfonline.com/doi/citedby/10.1080/13545700802 528315#tabModule https://www.tandfonline.com/doi/citedby/10.1080/13545700802 528315#tabModule G E N D E R D I S P A R I T Y I N E D U C A T I O N A N D T H E I N T E R N A T I O N A L C O M P E T I T I O N F O R F O R E I G N D I R E C T I N V E S T M E N T Matthias Busse and Peter Nunnenkamp A B S T R A C T With few exceptions, the empirical literature on foreign direct

investment (FDI) continues to be gender blind. This paper contributes to filling this gap by assessing the importance of gender inequality in education as a determinant of FDI. The authors estimate a standard gravity model on bilateral FDI flows that is augmented by educational variables, including different measures of gender inequality in education. The analysis covers an unprecedented number of both host and source countries of FDI, thereby reducing the risk of distorted results because of a sample selection bias. The results support the view that foreign investors are more likely to favor locations where education- related gender disparities are small. However, the discouraging effects of gender disparity on FDI are restricted to middle-income (rather than low-income) developing host countries and to investors from developed (rather than developing) countries. K E Y W O R D S Foreign direct investment, gender inequality, education JEL Codes: F23, I21, J16 I N T R O D U C T I O N The question of whether gender inequality hinders or helps the integration of countries into the international division of labor has received only scant attention in the empirical literature. Some evidence exists on

the links between gender inequality and trade. Matthias Busse and Christian Spielmann (2006) find that wage inequality is positively associated with comparative advantage in labor-intensive exports, whereas inequality in terms of labor-market participation and education is negatively related with such exports. According to Stephanie Seguino (1997), wage inequality may have contributed to the export success of countries such as South Korea.1 However, the role of gender inequality has been largely ignored in studies about the countries’ attractiveness for foreign direct investment (FDI). Feminist Economics 15(3), July 2009, 61–90 Feminist Economics ISSN 1354-5701 print/ISSN 1466-4372 online � 2009 IAFFE http://www.tandf.co.uk/journals DOI: 10.1080/13545700802528315 This is fairly surprising in light of the fierce international competition for FDI. Policy-makers are falling over themselves to entice foreign investors, for example, by offering tax breaks and outright subsidies, in the hope that FDI inflows would induce higher growth and employment. Yet,

it is still open to debate what actually drives FDI inflows.2 In particular, the sizeable literature on FDI determinants has generally been gender blind (Elissa Braunstein 2006). This paper attempts to fill this gap by assessing the role of gender disparity with respect to host countries’ attractiveness for FDI. The focus will be on education-related gender disparity and its effects on FDI flows to developing countries, for which the linkage is of particular concern.3 Opposing hypotheses in this regard call for empirical analyses. On the one hand, gender disparity in education may stimulate FDI by offering cost advantages if it leads to lower average wages at a given level of labor productivity. On the other hand, FDI may be discouraged if foreign investors increasingly rely on the local availability of skilled labor, which gender disparity in education is likely to constrain. We estimate a gravity model on bilateral FDI flows, covering as many (developing) host countries of FDI as possible to avoid a sample selection bias. The standard gravity model is augmented by educational variables, including different measures of gender inequality in education. We chose

this rather indirect approach of assessing the FDI effects of gender disparity as disparity measures directly capturing wage costs, labor productivity, and the qualification of the workforce by gender are unavailable or subject to serious data constraints. Our results clearly reject the view that foreign investors favor locations where education-related gender disparities may offer cost advantages. Rather, we find that gender disparity discourages FDI inflows. However, the strength of this relation depends on the level of education, being most pronounced with respect to secondary and tertiary education. Additional robustness tests reveal that the discouraging effect of gender disparity becomes statistically insignificant when considering only low- income host countries and developing source countries. P R E V I O U S S T U D I E S A N D G E N D E R D I S P A R I T Y M E A S U R E S Even though the literature on FDI determinants does not address gender issues, a strand of this literature on social factors and FDI relates to the analysis in this paper. Several studies raise the question of whether FDI tends to go where social standards are low and worker rights are repressed to save costs, or rather where social and political conditions are

similar to those prevailing in the home country.4 Howard J. Shatz (2003) focuses on education as a determinant of FDI but does not consider gender gaps in education. Shatz finds that better educated workers attract more FDI. The A R T I C L E S 62 counter-hypothesis is rejected, according to which FDI is undertaken ‘‘in countries with low levels of education to escape the high compensation costs with which higher levels of education and skill are associated’’ (Shatz 2003: 188). The question addressed in the following analysis – that is, whether gender inequality attracts or rather discourages FDI inflows – resembles this strand of the literature on FDI determinants in that there are two opposing hypotheses. On the one hand, gender disparity in education could be associated with higher FDI inflows. In the process of economic globalization, multinational companies appear to face mounting cost pressure. They increasingly refer to vertical types of FDI (also labeled efficiency-seeking FDI), which provides a

means to allocate specific steps of the production process to where the relevant comparative advantages can be utilized. Consequently, this type of FDI tends to be sensitive to international cost differentials. In particular, vertical FDI is often associated with the relocation of labor- intensive parts of the value chain to lower-wage locations. This may strengthen the incentives of multinational companies to exploit less-skilled, low-wage female labor. The movement of FDI in so-called footloose industries, such as textiles and clothing, to countries with segmented labor markets may provide cases in point.5 On the other hand, multinational companies may be more interested in drawing on sufficiently qualified labor rather than just cheap labor. David Kucera (2002) refers to survey results in which the managers of multinational companies rated the quality of labor in the host country to be more important than the cost of labor. Indeed, empirical evidence suggests that the labor demand of multinational companies is biased toward relatively skilled workers in developing host countries (Overseas Develop- ment Institute 2002). Furthermore, multinational companies are increasingly under pressure, notably from nongovernmental organizations (NGOs), to show good corporate behavior (Matthias Busse 2004). As a

consequence, they may shy away from host countries with pervasive social injustice in general and gender inequality in particular.6 It follows that the impact of gender inequality in education on FDI is theoretically ambiguous. Unit labor costs tend to decline to the extent that gender inequality in education involves lower average wages at a given level of labor productivity, with less educated women entering the labor force.7 Locations where education-related gender inequality is more pronounced might then have a competitive edge in attracting cost-oriented FDI of the vertical type. However, gender inequality may also be associated with higher unit labor costs, and thus less vertical FDI, if it is mainly associated with lower average labor productivity. In other words, gender inequality in education has opposing effects on unit labor costs. Moreover, the impact of gender inequality in education on FDI inflows would still be indeterminate even if unit labor costs declined on balance. In contrast to vertical FDI, the horizontal type of FDI (also labeled market-seeking FDI) may be unaffected G E N D E R D I S P A R I T Y A N D F D I 63

by changes in unit labor costs. This type of FDI essentially duplicates the parent company’s production at home in the host countries. Market access motivations dominate over cost considerations, and factor intensities of production in the host countries largely resemble those at home. Hence, the importance of unit labor costs for overall FDI inflows is likely to depend on the composition of FDI, which – though difficult to measure exactly – tends to vary considerably across host countries (see also Kucera [2002]). Ideally, we would like to cover several aspects of gender disparity and their effects on FDI inflows, including gender wage gaps, differences in labor-force participation rates between males and females, and education- related differences. The focus on education-related disparity measures implies some limitations. Inferences about the FDI effects of gender disparity in general remain indirect and incomplete. First, education- related measures tend to capture the net effects of two transmission mechanisms running through wages and labor productivity, without being able to disentangle them. Second, any positive FDI effects of less disparity in education may be associated with gender wage disparity to the

extent that an improved qualification of female workers does not lead to a corresponding pay rise. Consequently, concerns about gender equity and fairness would not necessarily be overcome if only less gender disparity in education resulted in more FDI. However, wage disparity measures and differences in labor- force participation rates are not particularly useful as possible determinants of FDI in the present context of a large panel of host countries and a time span of about twenty-five years because of the following reasons: . Data on wage differences are only available for selected years and a limited number of countries.8 The insufficient country coverage especially may cause seriously biased results when analyzing FDI determinants (Shatz 2003; Matthias Busse, Jens Königer, and Peter Nunnenkamp 2008). Moreover, when available, wage data typically refer to the manufacturing sector only (Kucera 2002; Busse and Spielmann 2006).9 This limitation is problematic, as FDI in developing countries increasingly consists of FDI in the services sector (United Nations Conference on Trade and Development [UNCTAD] 2004; Braunstein 2006). And finally, the problem of reverse causation running from FDI to wages and wage disparity would be all but impossible to resolve.

. Similar arguments apply to labor-force participation rates. Again, problems of reverse causality loom large (Braunstein 2006). The statistically insignificant results Kucera (2002) achieves when adding the proportion of female workers in the industry to his list of FDI determinants may well reflect that causality between FDI and female employment shares goes both ways (see also Elissa Braunstein A R T I C L E S 64 [2002]). Moreover, gender-specific labor-market participation rates do not necessarily reflect discrimination but rather may be based on voluntary decisions of female workers (Busse and Spielmann 2006). Consequently, education-related gender disparity appears to be the first choice when analyzing FDI determinants. While theory indicates that the level of education in a host country should influence FDI inflows (Shatz 2003), the possibility of reverse causation – meaning that higher FDI results in better education – seems to be rather remote in comparison with wages and employment. The empirical studies of Shatz (2003) as well

as Jonathan Eaton and Akiko Tamura (1996), considering education among the determinants of FDI, find that better educated workers in host countries attract higher FDI inflows. However, both studies cover only selected FDI source countries (United States FDI in the case of Shatz, US and Japanese FDI in the case of Eaton and Tamura). Furthermore, Braunstein’s (2006) verdict that most FDI studies are gender blind applies to both Shatz (2003) and Eaton and Tamura (1996). To the best of our knowledge, Kucera (2002) is the only exception in that he considers gender-specific educational variables as determinants of FDI. He does not find evidence suggesting that education-related gender disparity resulted in higher FDI inflows. Yet, his results are far from robust. The positive effect of (relative) female educational attainment on FDI is statistically significant only when high-income host countries are included in the sample, and the coefficient of this variable even changes its sign once the regressions are run with regional dummies. Moreover, Kucera’s study has some shortcomings that we attempt to overcome in the following analysis. First of all, it is purely cross-sectional, while we use a panel analysis to examine changes over time in

the relation between gender gaps in education and FDI. Second, we employ a gravity model on bilateral FDI flows, and we explicitly account for the fact that various host countries have not attracted any FDI flows from particular source countries. Third, we draw on a large, new dataset to cover essentially all (developing) host countries as well as a large number of source countries and thereby avoid, or at least substantially reduce, a sample selection bias. In the regressions reported below, we measure gender gaps in education by comparing females and males with respect to average years of schooling. While we also consider three different levels of education when estimating the Tobit model later in this paper, we confine the subsequent presentation of stylized facts to gender gaps in education at all levels of schooling combined, in order to save space. We compare the situation prevailing in 1980 with that in the most recent years (average of 2000 and 2005). The mean and the range of gender differences at specific levels of schooling are presented in the Appendix. G E N D E R D I S P A R I T Y A N D F D I 65

In Figure 1, ratios far below one reflect larger gender gaps in education working against women. On the other hand, women are overrepresented in some countries with ratios above one (notably in several Latin American countries). Not surprisingly, high-income countries, on average, have a relatively narrow gender gap in education, whereas the gap is widest in low- income countries. This applies to both 1980 and the most recent years. In contrast to what one might expect, however, there is also considerable variation over time.10 Middle-income countries, on average, caught up with high-income countries in terms of narrowing the gender gap; in recent years, middle-income countries resembled the high-income group in that the gender gap in education was less than 10 percent. At the same time, low-income countries, while still lagging behind, made remarkable progress in expanding the schooling of females relative to males. Moreover, the group averages reported in Figure 1 conceal considerably different developments in particular countries. This may be exemplified by three middle-income countries in Latin America. Colombia and Honduras started with a ratio of close to one in 1980 but through

subsequent developments diverged: females spent 24 percent more time in education than males in Colombia in recent years, whereas the ratio of females to males deteriorated to 0.67 in Honduras. Bolivia, starting with a pronounced gender gap (0.68), made substantial progress in closing this gap (to 0.88 in 2000/2005). Similar discrepancies apply to low-income countries in sub- Saharan Africa. Mozambique reported a large gender gap (0.23) at the beginning of the period of observation but a relatively narrow one recently Figure 1 Gender disparity in schooling,a 1980 and 2000/2005 Note: aAverage years of schooling at all levels combined: females divided by males; 2000/2005 represents the average for 2000 and 2005. Sources: Robert J. Barro and Jong-Wha Lee (2001) and UNESCO (2007). A R T I C L E S 66 (0.69). Ghana and Sudan both started at a ratio of females to males in education of about 0.4. While this ratio increased to 0.62 in Sudan, it declined slightly in Ghana. A P P R O A C H A N D D A T A

We follow a standard approach in the large empirical FDI literature11 and estimate a gravity-type model on the determinants of FDI. Gravity models are widely used to analyze the movement of goods, services, and factors of production between different locations within or across countries. The common intuition is to portray spatial transactions analogous to Newton’s Law of Gravity. Consequently, mass and distance are core elements of this class of models. In cross-country contexts as the present one, the focus is on economic size (in terms of income and/or population) and geographical distance between each pair of countries. Economic interaction is supposed to be an increasing function of the economic size of partner countries and a decreasing function of the distance between them; larger countries with higher income are thus expected to be involved in more transactions with nearer-by and larger countries. Various extensions of this basic model structure have been suggested in the literature. The extended version we use is specified below. As noted by Alan V. Deardorff (1998), this class of models first appeared in the empirical economics literature on bilateral trade flows. Deardorff also shows that simple gravity models can be derived from standard

trade theories. More recently, gravity models have also been applied to analyze financial flows. The explanatory power of gravity models on financial flows is comparable with that of models on trade flows (Philippe Martin and Hélène Rey 2004). According to Richard Portes and Hélène Rey, this is hardly surprising as the gravity approach ‘‘emerges naturally’’ from theories of asset trade (2005: 275). Recent examples employing gravity models to analyze bilateral FDI include Shatz (2003) and John H. Mutti and Harry Grubert (2004). Hence, in contrast to Avik Chakrabarti’s earlier verdict of ‘‘measurement without theory’’ (2001: 90), there appears to be widespread agreement by now on the appropriate analytical framework to guide empirical work on the determinants of FDI. Indeed, variables such as market size and openness to trade that the extreme bounds analysis of Chakrabarti (2001) found to be fairly robust determinants of FDI represent important cornerstones of the gravity model. The particular advantage of the extended gravity model in explaining the determinants of bilateral FDI flows is the fact that differences between source and host country characteristics can be used as explanatory variables. A standard FDI analysis using aggregated FDI flows for each country would not be suitable for that task.

While the core variable set of gravity models helps prevent fragile results due to ad-hoc choices on control variables, the estimation results may still G E N D E R D I S P A R I T Y A N D F D I 67 be sensitive to sample selection. Shatz’s (2003) analysis of US FDI clearly reveals that sample selection matters for empirical results.12 Consequently, we cover as many countries as possible in our baseline regressions and, at the same time, perform robustness tests for specific sub- samples.13 Furthermore, when applying gravity models to FDI flows, one must take into account the concentration of FDI in a few host countries. During the period under consideration (1978–2004), about 80 percent of FDI flows to all (150) middle- and low-income countries were concentrated in just twenty countries (World Bank 2006). Bilateral FDI flows are often equal to zero; this applies to roughly three-quarters of all observations in our sample. The censored nature of this variable implies that the results from OLS estimations would very likely be biased. Therefore, we use

a non-linear method of estimation such as Tobit. The Tobit model represents our preferred option among alternative approaches suggested in the literature to avoid biased results when the dependent variable is censored.14 This model estimates FDI flows between a particular pair of countries in one step. The underlying assumption is that the explanatory variables have the same impact on (1) the probability of receiving any bilateral FDI at all (selection decision) and (2) the amount of FDI allocated thereafter (allocation decision). This assumption appears to be reasonable in the context of bilateral FDI flows; it would be difficult to find an exclusion variable that affects selection but does not affect allocation, as two-step models such as the Heckman model would require. In our empirical approach, we principally follow David L. Carr, James R. Markusen, and Keith E. Maskus (2001), who estimate the so- called knowledge-capital model that integrates the previously separate concepts of horizontal (market-seeking) and vertical (efficiency-seeking) FDI into a single model by considering the determinants of both types of FDI within a single estimation equation.15

Our basic specification reads as follows: InðFDI ijtÞ¼ a0 þ a1lnðFDI ijt�1Þþ g 0X jt þ f0Y ijt þ a2 GenderInequalityjt þ lt þ eijt ð1Þ where FDIijt stands for foreign direct investment of country i in country j at period t, FDIijt-1 corresponds to FDI inflows in the previous period t-1, Xjt represents a set of host country control variables, Yijt denotes the difference between source and host country characteristics, lt is a set of year dummies, and GenderInequalityjt corresponds to gender inequality in education between males and females in the host country. The error term of the random effects estimation can be written as: eijt ¼ nijt þ uijt ð2Þ A R T I C L E S 68 where uijt is the random unobserved bilateral effect and vijt represents the remaining error.16 For the dependent variable, we use two measures of FDI: first, FDI flows

from the source to the (developing) host country in percent of host country GDP (the variable is labeled FDI1), and second, the share of FDI attracted by a specific (developing) host country in total FDI flows from the source country under consideration to all developing host countries (FDI2) included in our sample. The second measure captures the attractiveness of a particular country relative to other host countries. We calculate three-year averages to smooth the considerable fluctuation of annual bilateral FDI flows. At the same time, this approach ensures that we have enough variation in the data.17 The limited host country coverage of previous analyses of bilateral FDI flows is overcome by fully exploiting the (largely unpublished) data available upon request from UNCTAD’s Data Extract Service. Yet, some data limitations remain. Most importantly, it is not possible to differentiate between different types of bilateral FDI flows. For instance, vertical FDI should be affected more strongly by gender inequality in education than horizontal FDI; it is mainly the former type of FDI that is supposed to depend on international cost differences as well as the availability of sufficiently qualified labor in the relevant literature. Likewise, the impact of

gender inequality in education may be stronger in the case of greenfield FDI, compared with mergers and acquisitions (M&As), which amount to a change in ownership of existing production facilities and may be driven by asset-seeking motives in the first place.18 Tax-induced distortions in international FDI patterns are minimized by excluding FDI flows to offshore financial centers (see also below); but the problem remains that FDI channeled through offshore centers to the ultimate host country cannot be accounted for appropriately. We include the lagged dependent variable on the right-hand side of the regression equation for two reasons. First, this solves the potential problem of autocorrelation in the pooled time-series regressions.19 Second, this procedure is theoretically plausible as foreign investment in the previous period is highly relevant for FDI in the current period. Above all, countries that already have considerable FDI inflows are much more likely to attract multinational corporations. This has been shown, for example, by Victor M. Gastanaga, Jeffrey B. Nugent, and Bistra Pashamova (1998); the lagged FDI variable is always highly statistically significant in their regressions. By including lagged FDI flows, the econometric specification becomes a

dynamic panel. We employ a fairly standard set of controls, including total host country population and real GDP growth for market-seeking FDI (labeled population and growth, respectively),20 host country inflation (inflation), host country openness to trade (openness), the difference in GDP per capita G E N D E R D I S P A R I T Y A N D F D I 69 between the source and the host country for vertical FDI (DiffGDPpc), and a dummy for the existence of a bilateral or regional trading agreement – that is, a free trade agreement or customs union (RTA). We expect a positive association of population, growth, DiffGDPpc, and RTA with FDI; the opposite applies to inflation, as this variable can be interpreted as a proxy for macroeconomic distortions. Exact definitions and data sources for all variables as well as descriptive statistics can be found in the Appendix. As for time invariant variables, we also closely follow the empirical literature on gravity models and incorporate dummies for a common border, a

common language, and colonial ties as well as the distance between the source and the host country (distance). The first three control variables are expected to be positively associated with FDI flows, whereas the sign of distance is unclear. On the one hand, management and transport costs are likely to increase if two countries are located far away from each other; on the other hand, remote markets might be better served through local production, that is, FDI in the host country. Hence, the net impact on FDI is uncertain. To reduce the skewness in the data, we take the natural logarithm of population, FDI1, FDI2, DiffGDPpc, distance, and inflation. But this would mean that we would lose observations with negative values or zeros. To overcome this problem, we use the following logarithmic transformation that reduces the skewness in the data and, at the same time, retains negative and zero observations: y ¼ ln x þ ffiffiffiffiffiffiffiffiffiffiffiffiffiffiffiffiffi x2 þ 1ð Þ p� � ð3Þ Using this transformation leaves the sign of x unchanged, while

the values of x pass from a linear scale at small absolute values to a logarithmic scale at large values. In addition to these standard control variables, we include the institutional development of host countries, proxied by political constraints on the executive branch (political constraints). Poor institutions may discourage FDI by giving rise to uncertainty (for example, with respect to the protection of property rights [Jeong-Yeon Lee and Edwin Mansfield 1996; Witold J. Henisz 2000]) and additional costs (for example, in the case of corruption [Shang-Jin Wei 2000]). We use the index for political constraints that was developed by Henisz (2000). In contrast to alternative institutional indicators, this variable is available for a large number of countries and years. Political constraints focuses on the political discretion of the executive branch. Less discretion is supposed to render credible commitments to (foreign) investors more likely. The indicator ranges from zero (total political discretion) to one (no political discretion). Thus, we expect a positive link between political constraints and FDI flows. Finally, we include two variables that control for investment liberalization: (1) CapOpen A R T I C L E S

70 for unilateral capital account liberalization of the host country (Menzie D. Chinn and Hiro Ito 2005) and (2) BIT for a bilateral investment treaty ratified between the source and the host country (Busse, Königer, and Nunnenkamp 2008). Both measures are expected to stimulate higher FDI flows. Lastly, we measure the variable of principal interest, gender inequality in education, as the difference between the male and female score for average years of schooling in the population aged 25 and above (education inequality).21 In additional estimations, we use more detailed information of gender inequality in primary, secondary, and tertiary education. This allows us to examine at which level of education gender inequality matters most for the host countries’ attractiveness to FDI. Needless to say, we also control for years of schooling of both sexes combined with respect to either all levels of schooling (education) or specific levels of schooling (primary education, secondary education, and tertiary education). Our analysis covers the period 1978–2004, that is, optimally

nine observations of three-year averages for all indicators. UNCTAD’s Data Extract Service provides FDI data since 1970, but very few countries report FDI flows for the 1970s at a bilateral level. To avoid any biases arising from an extremely small sample of reporting countries, we start with 1978. We exclude financial offshore centers, such as Panama, the Bahamas, or the Cayman Islands.22 Extending the sample to include a large number of poor developing host countries is crucial to avoid a sample selection bias and to assess the chances of these countries becoming more attractive to FDI. Our sample consists of seventy-seven developing host countries, based on the World Bank’s classification of low- and middle-income countries.23 By covering twenty-eight FDI source countries, including various non-OECD source countries, we at least partly capture the recent surge of FDI flows from developing countries to other developing countries. The Appendix includes lists of the source and host countries. M A I N R E S U L T S Following the model specification and the introduction of the variables, we now turn to the empirical results. Table 1 reports the results of the Tobit model for both FDI variables and total years of schooling. The

estimations include all control variables introduced before. Apart from inflation, all control variables have the expected sign, and the significance of the coefficients is not much affected when considering FDI as a share of GDP (FDI1) or FDI shares (FDI2) as the dependent variable. As anticipated, FDI in the past is a strong predictor for current FDI as the coefficient of the lagged dependent variable is positive and highly statistically significant. The strongly positive coefficients of the host countries’ populations, GDP growth rates, and the differences in per G E N D E R D I S P A R I T Y A N D F D I 71 capita income between the host and source countries (DiffGDPpc) reveal that FDI flows to the sample countries are driven by both market-seeking and efficiency-seeking motives (horizontal and vertical FDI). The importance of vertical FDI is also indicated by the significantly positive coefficient of openness; greater openness to trade reflected in this variable improves the host countries’ attractiveness to FDI involving the relocation of particular

segments of the value chain and the offshoring of intermediate production.24 Likewise, less regulated capital transactions are associated with higher bilateral FDI flows, as the coefficient of CapOpen is positive and significant at the 1 percent level. Apart from colonial ties in one specification, all the time- invariant variables traditionally included in gravity models turn out to be statistically significant at the 5 percent level. Bilateral FDI flows between a source and a host country having a common border or speaking the same language are higher than bilateral flows between countries without such common characteristics. The same applies for colonial ties (except column [2]). By contrast, a larger distance between the host and the source country tends to reduce bilateral FDI flows. Distance-related management and transport costs outweigh the source country’s incentive to undertake FDI in remote countries and serve these markets through local production. Table 1 FDI and education, total years of schooling (1) (2) Dependent variable ln (FDIl) ln (FDI2) ln (FDIt-1) 0.299*** (0.012) 0.619*** (0.020) Education 0.108*** (0.021) 0.035*** (0.005) Education inequality -0.128*** (0.043) -0.050*** (0.009)

ln (population) 0.291*** (0.033) 0.101*** (0.008) ln (DiffGDPpc) 0.044*** (0.009) 0.007*** (0.002) Growth 0.027** (0.012) 0.005** (0.002) ln (inflation) 0.004 (0.026) 0.003 (0.005) Openness 0.004*** (0.001) 0.001*** (0.000) Common border 0.674** (0.290) 0.295*** (0.061) Common language 0.476*** (0.110) 0.080*** (0.023) ln (distance) -0.539*** (0.063) -0.143*** (0.014) Colonial ties 0.448** (0.210) 0.037 (0.044) RTA 0.518** (0.200) 0.050 (0.041) Political constraints 0.710*** (0.190) 0.022 (0.038) CapOpen 0.081*** (0.029) 0.020*** (0.006) BIT 0.390*** (0.096) 0.028** (0.012) Observations 8,299 8,299 Country pairs 1,531 1,531 Notes: Marginal effects, computed at the mean, are displayed; standard errors are reported in parentheses; due to space constraints, the coefficients for constant term and the year dummies are not shown; *** significant at 1 percent level, ** significant at 5 percent level, and * significant at 10 percent level. The p-values of the Wald w2 test for the null hypothesis that all explanatory variables equal zero are always statistically significant at the 1 percent level (not reported). A R T I C L E S 72 Results turn out to be weaker for some other control variables. In contrast to our expectations, inflation is positive but never significant.25

RTA has the expected positive coefficient but fails to reach the conventional 10 percent significance level in one specification. We obtain a similar outcome for political constraints – that is, a positive linkage with FDI in both specifications but only one (highly) significant coefficient. Finally, the ratification of bilateral investment treaties (BIT) leads to higher FDI inflows, which is in line with previous findings by Busse, Königer, and Nunnenkamp (2008). Turning to the education-related determinants of FDI, our results corroborate Shatz (2003) as well as Eaton and Tamura (1996) in that average years of schooling of both sexes taken together (education) are associated with higher FDI flows at the 1 percent level. In the present context of gender inequality, it is still more important that education inequality is negatively related to bilateral FDI flows. The coefficient of our variable of principal interest, which captures the difference between male and female years of total (primary, secondary, and tertiary) schooling, turns out to be significant at the 1 percent level for the full sample of (developing) host countries. Hence, our panel analysis produces stronger results than the cross-section analysis of Kucera (2002). While Kucera finds no evidence suggesting that gender disparity in education leads

to higher FDI inflows, our results support the stronger conclusion that gender disparity in education clearly reduces FDI inflows. The quantitative effect of less gender disparity in education on FDI inflows is modest but by no means negligible. Taking the estimated coefficient on education inequality with FDI1 as the dependent variable (70.128) at face value, a decrease in the difference between male and female years of total schooling by 0.25 years (that is, the standard deviation of education inequality) would lead – on average – to an increase in the FDI/GDP ratio by some 2.5 percent.26 The long-run effect would be still more pronounced. The long- run effect can be calculated by dividing the coefficient of education inequality by one minus the coefficient of the lagged dependent variable. Based on the estimate reported in column (1) of Table 1, the long-run FDI effect of a decrease in education inequality by one standard deviation would amount to 3.6 percent of FDI inflows as a share of GDP. Overall, the findings for the effect of gender inequality on FDI underscore the findings for the effect of the education of both sexes combined on FDI: In the first place, the attractiveness of host countries to FDI stems from offering foreign investors the opportunity to draw on

sufficiently qualified labor, be it male or female workers. This does not rule out that foreign investors aim to reduce wage costs for similarly qualified labor.27 But the estimation results suggest that the wage- reducing motive of FDI is dominated by the motive to complement FDI-related production techniques with sufficiently qualified labor in the host country. Gender inequality in education tends to constrain this option as it limits the pool of G E N D E R D I S P A R I T Y A N D F D I 73 locally available labor that meets the standards required by foreign investors. In the next step of our analysis, we differentiate the educational variables (average years of schooling of both sexes combined as well as gender disparity related to years of schooling) by considering three levels of schooling separately. In all other respects, the specification of the Tobit model remains as before.28 The results shown in Table 2 suggest that education at all levels is positively associated with FDI inflows. In contrast to Shatz (2003), we do

not find that primary education had stronger effects on FDI than higher levels of education. The pattern found here for various sources of FDI appears to be plausible given that primary education tends to be a weaker indicator of the availability of skilled labor than higher levels of education. US FDI (analyzed by Shatz) may deviate from this pattern because the motives underlying US FDI differ from those underlying FDI from other sources.29 Off-shoring labor-intensive parts of the production process to lower income host countries appears to figure relatively prominently in US FDI, which may thus depend less on skilled local labor. Moreover, we do not find any evidence that gender inequality results in higher FDI inflows either at the lower level of primary education or at the higher level of secondary and tertiary education. Rather, as before for total schooling, all coefficients on education inequality are statistically significant and have the same negative sign. Apart from inequality in tertiary education, the coefficients reach the 1 percent significance level. However, the results also show that the size of the coefficients at the secondary and tertiary level of education is considerably higher in comparison to the

primary level. In other words, changes in secondary or tertiary education disparities have a much stronger impact on FDI inflows than changes at the primary level. S E N S I T I V I T Y A N A L Y S E S We perform two types of sensitivity analyses in this section. First, we replicate our estimations with average years of schooling at all levels combined for various sub-samples of host and source countries, and second, we perform Tobit fixed-effects estimations to control for country- pair fixed effects. In the estimations reported in Table 3, we return to average years of schooling at all levels combined as a measure of gender inequality in education. To save space, we show only the results for the variable of principal interest in the present context.30 To facilitate comparison, the main results from Table 1 are listed again in the first row of Table 3. The size of the discouraging effect of gender inequality in education on FDI inflows may depend on the stage of development of the host country. A R T I C L E S

74 T a bl e 2 F D I a n d p ri m a ry , se co n d a ry

, a n d te rt ia ry e d u ca ti o n (1 ) (2 ) (3 ) (4 ) (5

) (6 ) E d u ca ti on le v el P ri m a ry ed u ca ti on S ec on

d a ry ed u ca ti on T er ti a ry ed u ca ti on D ep en d en t v

a ri a bl e ln (F D Il ) ln (F D I2 ) ln (F D Il ) ln (F D I2

) ln (F D Il ) ln (F D I2 ) ln (F D I t -1 ) 0 .3 0 0 * * *

(0 .0 1 2 ) 0 .6 2 0 * * * (0 .0 2 0 ) 0 .3 0 2 * * * (0 .0

1 2 ) 0 .6 3 1 * * * (0 .0 2 1 ) 0 .3 0 4 * * * (0 .0 1

2 ) 0 .6 3 9 * * * (0 .0 2 1 ) E d u ca ti o n 0 .1 6 3

* * * (0 .0 3 0 ) 0 .0 5 5 * * * (0 .0 0 7 ) 0 .2 1 3 *

* * (0 .0 6 2 ) 0 .0 5 8 * * * (0 .0 1 3 ) 0 .7 8 6 * * *

(0 .2 0 0 ) 0 .2 3 0 * * * (0 .0 4 2 ) E d u ca ti o n

in e q u a li ty -0 .1 8 1 * * * (0 .0 7 0 ) -0 .0 7 0 * * *

(0 .0 1 5 ) -0 .4 1 3 * * * (0 .1 1 0 ) -0 .1 3 5 * * * (0 .0

2 3 ) -0 .7 4 9 * (0 .4 4 0 ) -0 .3 7 2 * * * (0 .0 9 0

) ln (p o p u la ti o n ) 0 .2 8 4 * * * (0 .0 3 3 ) 0 .0

9 8 * * * (0 .0 0 8 ) 0 .3 1 3 * * * (0 .0 3 3 ) 0 .1 0

5 * * * (0 .0 0 8 ) 0 .2 8 2 * * * (0 .0 3 0 ) 0 .0 9 3 *

* * (0 .0 0 7 ) ln (D if fG D P p c) 0 .0 4 5 * * * (0 .0 0

9 ) 0 .0 0 7 * * * (0 .0 0 2 ) 0 .0 3 7 * * * (0 .0 0 9

) 0 .0 0 4 * * (0 .0 0 2 ) 0 .0 3 5 * * * (0 .0 0 9 ) 0

.0 0 4 * * (0 .0 0 2 ) G ro w th 0 .0 2 8 * * (0 .0 1 2 )

0 .0 0 6 * * (0 .0 0 2 ) 0 .0 2 3 * * (0 .0 1 2 ) 0 .0 0

4 * (0 .0 0 2 ) 0 .0 2 6 * * (0 .0 1 2 ) 0 .0 0 5 * *

(0 .0 0 2 ) ln (i n fl a ti o n ) 0 .0 0 1 (0 .0 2 6 ) 0 .0

0 1 (0 .0 0 5 ) 0 .0 1 5 (0 .0 2 6 ) 0 .0 0 6 (0 .0 0 5

) 0 .0 2 6 (0 .0 2 6 ) 0 .0 0 9 * (0 .0 0 5 ) O p e n

n e ss 0 .0 0 4 * * * (0 .0 0 1 ) 0 .0 0 1 * * * (0 .0 0

0 ) 0 .0 0 5 * * * (0 .0 0 1 ) 0 .0 0 1 * * * (0 .0 0 0

) 0 .0 0 6 * * * (0 .0 0 1 ) 0 .0 0 1 * * * (0 .0 0 0 ) C

o m m o n b o rd e r 0 .6 9 9 * * (0 .2 9 0 ) 0 .3 0

3 * * * (0 .0 6 1 ) 0 .6 5 5 * * (0 .2 9 0 ) 0 .2 8 2 *

* * (0 .0 6 1 ) 0 .6 7 4 * * (0 .2 9 0 ) 0 .2 7 8 * * *

(0 .0 6 1 ) C o m m o n la n g u a g e 0 .4 7 6 * * *

(0 .1 1 0 ) 0 .0 8 0 8 * * * (0 .0 2 3 ) 0 .4 6 9 * * *

(0 .1 1 0 ) 0 .0 7 6 * * * (0 .0 2 3 ) 0 .4 6 6 * * * (0 .1

1 0 ) 0 .0 7 0 4 * * * (0 .0 2 3 ) L n (d is ta n ce )

-0 .5 3 5 * * * (0 .0 6 3 ) -0 .1 4 1 * * * (0 .0 1 4 ) -0

.5 3 3 * * * (0 .0 6 3 ) -0 .1 3 7 * * * (0 .0 1 4 ) -0 .5 1

7 * * * (0 .0 6 3 ) -0 .1 3 3 * * * (0 .0 1 4 ) C o lo n

ia l ti e s 0 .4 5 3 * * (0 .2 1 0 ) 0 .0 3 8 5 (0 .0 4

4 ) 0 .4 2 7 * * (0 .2 1 0 ) 0 .0 2 7 (0 .0 4 4 ) 0 .4

3 5 * * (0 .2 1 0 ) 0 .0 2 9 (0 .0 4 3 ) R T A 0 .4 6

9 * * (0 .2 1 0 ) 0 .0 3 3 (0 .0 4 1 ) 0 .6 5 9 * * * (0

.2 0 0 ) 0 .0 9 3 * * (0 .0 4 1 ) 0 .6 2 0 * * * (0 .2 0

0 ) 0 .0 7 9 * (0 .0 4 1 ) P o li ti ca l co n st ra in ts

0 .7 5 5 * * * (0 .1 9 0 ) 0 .0 3 4 (0 .0 3 8 ) 0 .7 6 7

* * * (0 .1 9 0 ) 0 .0 5 0 (0 .0 3 9 ) 0 .8 9 6 * * *

(0 .1 9 0 ) 0 .0 8 7 1 * * (0 .0 3 7 ) C a p O p e n

0 .0 8 8 * * * (0 .0 2 9 ) 0 .0 2 2 * * * (0 .0 0 6 ) 0

.0 8 5 * * * (0 .0 2 9 ) 0 .0 2 2 * * * (0 .0 0 6 ) 0 .1 0

0 * * * (0 .0 2 9 ) 0 .0 2 5 * * * (0 .0 0 6 ) B IT 0 .3

8 8 * * * (0 .0 9 6 ) 0 .0 2 6 * (0 .0 1 6 ) 0 .4 3 5 *

* * (0 .0 9 6 ) 0 .0 4 2 * * (0 .0 1 9 ) 0 .4 1 9 * * *

(0 .0 9 5 ) 0 .0 3 5 * (0 .0 1 9 ) O b se rv a ti o n s 8

,2 9 9 8 ,2 9 9 8 ,2 9 9 8 ,2 9 9 8 ,2 9 9 8 ,2 9 9 C

o u n tr y p a ir s 1 ,5 3 1 1 ,5 3 1 1 ,5 3 1 1 ,5 3

1 1 ,5 3 1 1 ,5 3 1 N ot es : S e e T a b le 1 ; * * *

si g n ifi ca n t a t 1 p e rc e n t le ve l, * * si g n

ifi ca n t a t 5 p e rc e n t le ve l, a n d * si g n ifi

ca n t a t 1 0 p e rc e n t le ve l. G E N D E R D I S P A R I T Y A N D F D I 75 For this reason, we test whether a differentiation of the fairly heterogeneous group of developing host countries offers additional insights. Indeed, the discouraging effects of gender inequality on FDI are confined to middle-income countries, which (according to the World

Bank’s classification) comprise countries with a per capita income of between US$876 and US$10,725 in 2005 (World Bank 2006). By contrast, gender inequality remains completely statistically insignificant as a determinant of FDI in low-income countries, that is, countries with a per capita income of US$875 or less. Some types of FDI undertaken in low-income countries are rather unlikely to be motivated by the availability of qualified labor. For example, this probably applies to resource-seeking FDI in the primary sector, which accounts for the bulk of total FDI flows to various low-income countries. In any case, qualified labor tends to be in extremely short supply in these host countries, and less gender inequality in education is unlikely to improve this situation substantially. Consequently, foreign investors in low-income host countries may care less about gender inequality than in more advanced host countries. It is important to note, however, that even in low-income countries gender inequality does not induce more FDI. Next, we check whether the impact of gender inequality on FDI has changed over time. One would expect that the discouraging effect had become more pronounced in recent years. The demand of foreign

investors for qualified local labor may have risen with the increasing complexity of production techniques transferred to the host countries. In fact, there is support for this proposition as the size of the coefficients is larger (in both regressions) when the estimations are based on the 1990– 2004 period, instead of between 1978 and 2004. Next, we replicate the estimations for two groups of source countries. As mentioned previously, developing countries have increasingly become sources of FDI. Arguably, the motives underlying FDI from developing countries differ from the motives underlying FDI from more advanced Table 3 Robustness checks and extensions, education inequality (1) (2) Dependent variable ln (FDI1) Ln (FDI2) Full sample (as reported in Table 1) -0.128*** (0.043) - 0.050*** (0.009) Middle-income countries -0.246*** (0.059) -0.066*** (0.013) Low-income countries 0.075 (0.085) 0.007 (0.011) Period 1990–2004 -0.204*** (0.060) -0.107*** (0.017) Developed source countries -0.182*** (0.061) -0.062*** (0.010) Developing source countries 0.021 (0.039) -0.003 (0.019) Notes: To save space, we only report the results for the education inequality variable; *** significant at 1 percent level, ** significant at 5 percent level, and *

significant at 10 percent level. See Table 1 for further notes. A R T I C L E S 76 source countries: On the one hand, wage-related cost savings could be a less important driving force of FDI undertaken by less developed source countries in other developing countries, since wages tend to be similar in the source and the host country. Ceteris paribus, this could have strengthened the discouraging effect of gender inequality on FDI from developing countries. On the other hand, some relatively advanced developing source countries appear to have used FDI as a means to relocate less sophisticated industrial activities to where cost savings could be realized.31 This type of FDI probably draws less on qualified labor in the lower-income host countries.32 It turns out that gender inequality in education enters insignificantly when the estimation is restricted to developing source countries. The results for the full sample of source countries are mainly driven by the discouraging effect of gender inequality on FDI from developed source

countries. For the latter, the significance level closely resembles the general pattern reported in Table 1 and the size of the coefficients is somewhat larger. Finally, the results presented so far are based on a random- effects model, and it may be argued that they are mainly driven by variations across countries rather than over time. To account for this potential weakness of our results, we replicate the analysis from Table 1 using a Tobit fixed-effects model as a robustness check. The results show that the country fixed effects capture a considerable part of the variation in the dependent variables, as a number of independent variables are no longer statistically significant (Table 4). Above all, this applies to education and differences in GDP per Table 4 FDI and total years of schooling, fixed-effects estimation (1) (2) Dependent variable ln (FDIl) ln (FDI2) ln (FDIt-1) 0.023* (0.013) 0.199*** (0.012) Education 0.030 (0.130) 0.000 (0.023) Education inequality -0.051* (0.032) -0.021** (0.011) ln (population) 0.213* (0.174) 0.241* (0.130) ln (DiffGDPpc) -0.028 (0.023) 0.004 (0.004) Growth 0.018 (0.013) 0.004* (0.002) ln (inflation) -0.038 (0.034) -0.001 (0.006)

Openness 0.008*** (0.003) -0.001 (0.001) RTA 0.436 (0.290) 0.096* (0.053) Political constraints 0.352 (0.280) 0.006 (0.051) CapOpen 0.114*** (0.044) 0.008 (0.008) BIT 0.178* (0.101) 0.006* (0.003) Observations 8,299 8,299 Country pairs 1,531 1,531 Notes: See Table 1; *** significant at 1 percent level, ** significant at 5 percent level, and * significant at 10 percent level. G E N D E R D I S P A R I T Y A N D F D I 77 capita. On the other hand, market size (population), economic growth, openness to trade, joining a regional trade agreement, and liberalizing the capital account through unilateral measures or bilateral investment treaties still matter for FDI flows, though significance levels tend to be weaker in comparison to the random-effects model and the coefficients often remain insignificant in one of our two specifications. Importantly, gender inequality in education is always negatively associated with FDI inflows; the coefficient is significant at the 10 percent level or better. Jointly with the previous evidence from additional regressions in this section, this outcome

demonstrates that the link between education inequality and FDI inflows is quite robust. C O N C L U S I O N S With few exceptions, the empirical literature on FDI continues to be gender blind. This paper contributes to filling this gap by assessing the importance of gender inequality in education as a determinant of FDI. We estimate a standard gravity model on bilateral FDI flows, which is augmented by educational variables, including measures of gender inequality in education. Since we lack sufficient data on disparity measures such as wages, labor productivity, and worker qualification by gender, this approach takes an indirect route by testing the opposing propositions that gender disparity in education may either stimulate FDI by reducing unit labor costs or discourage FDI by constraining the local availability of sufficiently qualified labor. We find no evidence whatsoever that multinational companies favor locations where education-related gender disparity exists. Gender disparity in education clearly discourages FDI flows from developed countries to relatively advanced (middle-income) developing countries. However, the

effect is statistically insignificant in low-income host countries. The latter finding can be attributed to the prominence of specific types of FDI that rely considerably less on qualified local labor; resource-seeking FDI in the primary sector of low-income countries is a case in point. Likewise, the motivation underlying FDI from developing countries – including resource- seeking and cost-oriented FDI in fairly poor developing countries – provides an explanation for why this group of foreign investors appears to care less about gender inequality in the host countries. The finding that gender disparity does not attract FDI for any of the sub- groups of host and source countries under consideration has important implications for the fierce international competition for FDI inflows. It would clearly be counter-productive if policy-makers entered into a race to the bottom not only by lowering corporate tax rates or corporate contributions to social security systems but also by being lenient about the still widespread gender gaps in education. It is obviously difficult to A R T I C L E S 78

prove that policy-makers consciously maintained gender gaps in education to contain wage increases for unskilled labor. It cannot be ruled out, however, that policy-makers are tempted not to fight gender gaps in education effectively – in the erroneous belief that having a pool of cheap unskilled labor will attract FDI. Particularly in relatively advanced (middle- income) developing countries, policy-makers would rather be well advised to tackle the persistent gender disparity to improve their countries’ attractiveness to FDI, if not for more general reasons of fairness and equity. But even in low-income developing countries, it would not pay to maintain gender gaps in education, if we recall that the effect on FDI was statistically insignificant for these host countries. This is not to ignore that cost-related dimensions of gender inequality, notably wage discrimination, may offer short-term benefits to investors, help attract FDI that is mainly motivated by the availability of cheap labor, and provide a (temporary) boost to economic growth associated with footloose FDI that is unlikely to stay. In the longer run, however, we argue that policy-makers should be aware of the adverse effects of gender disparity on both FDI inflows and economic growth if persistent inequality

in education adds to the supply of cheap female workers. Our estimation results suggest that the negative effects of gender disparity on FDI are quantitatively modest in the short run but clearly become more important over time. This implies that persistent gender disparity in education would run the risk of developing host countries ending up in a trap of low wages, low labor productivity, and footloose FDI. Multinational companies in the manufacturing and services sectors tend to rely on relatively skilled labor in the host countries. Unskilled labor- intensive FDI – for example, in footloose industries such as clothing and footwear – may have received considerable public attention. But empirical evidence indicates that foreign investors in developing countries typically apply more advanced production techniques than local firms operating in the same industry, and FDI is frequently concentrated in skill- intensive industries (Overseas Development Institute 2002). It also appears that multinational companies are pursuing increasingly complex integration strategies (UNCTAD 1998), in which educated and well-trained labor plays an important role. With labor demand of foreign investors being focused on higher skills, better-educated and high-skilled women would enhance

the attractiveness to FDI by adding to the pool of skilled labor available in a host country. Our findings suggest that less gender disparity in education would promote FDI-related economic growth in the long run. According to the relevant literature, the growth effects of FDI in developing host countries critically depend on the degree to which transfers of technology and know-how are disseminated throughout the host economy.33 Local absorptive capacity plays an important role with regard to FDI- related G E N D E R D I S P A R I T Y A N D F D I 79 spillovers: FDI can only be expected to provide a stimulus to economy- wide productivity gains if local producers and workers are sufficiently qualified to imitate superior technology and acquire advanced skills. According to Eduardo Borensztein, José De Gregorio, and Jong- Wha Lee (1998), local human capital constraints hinder stronger growth effects of FDI. Increasing female education and skills levels can help overcome this constraint and, thereby, help transform larger FDI inflow into

higher economic growth. By analyzing the FDI effects of gender inequality in education, we specify an important transmission mechanism that has received little attention in the literature on gender inequality and economic growth. However, the present paper offers just one more piece of the complex puzzle on gender inequality and economic growth. Further research is clearly required in several respects. First, we do not address dimensions of gender disparity other than education-related disparity. In particular, immediately cost- related dimensions, such as wage discrimination, may have different implications for specific types of FDI. Wage disparity may attract cost- oriented FDI, notably the off-shoring of labor-intensive parts of production. This would resemble the finding of Busse and Spielmann (2006) that wage disparity is positively associated with comparative advantage in labor- intensive export production. Furthermore, the particular dimension of gender inequality matters not only for FDI but also more broadly for the economic development of the host countries (Jean Drèze and Amartya Sen 1989; Lant Pritchett and Lawrence H. Summers 1996; Stephanie Seguino 2000; Stephan Klasen 2002).

Second, disaggregating the different types of FDI may also complement the picture of FDI-related transmissions of gender disparity to growth. Specific types of FDI are likely to respond differently to gender disparity. For example, FDI of the horizontal (or local market-seeking) type may be less affected than FDI of the vertical (or efficiency-seeking) type by gender inequality in terms of both education and labor costs. At the same time, the growth effects may depend on the specific type of FDI. The literature on the growth effects of (aggregate) FDI suggests that attracting more FDI per se is no guarantee to achieve higher growth (for example, Maria Carkovic and Ross Levine 2005). Data constraints render it difficult to assess the transmission mechanisms between gender disparity and growth for specific types of FDI in the context of broad (host and source) country samples. A more promising option could be to conduct case studies for specific host countries attracting different types of FDI. Another option would be to focus on one particular source country, notably the US, which offers detailed data on the operations of foreign affiliates that may allow for a distinction between different types of FDI. Finally, future research may attempt to provide an integrated

account of several transmission mechanisms between gender disparity and growth. An A R T I C L E S 80 important step in this direction would be to simultaneously account for FDI and trade-related effects. These two transmission mechanisms may well work in opposite directions. This might be the case, for instance, if wage inequality strengthened comparative advantage in labor- intensive exports of developing countries (as found in the trade paper of Busse and Spielmann 2006) but no longer attracted FDI. The tendency of multinational companies to contract out unskilled, labor- intensive work to local firms, and to purchase inputs from them, could explain why wage inequality promoted trade, while leaving FDI unaffected or even reducing it. On the other hand, both transmission mechanisms may reinforce each other. This possibility arises from the above finding that less gender disparity in education induces more FDI, in combination with the earlier

finding of Busse and Spielmann (2006) that less gender disparity in education is positively associated with comparative advantage in labor- intensive exports. From a gender perspective, the equity and fairness implications of such a scenario would still remain unresolved. It would have to be assessed whether and to what extent wage discrimination is underlying the positive export and FDI effects of less gender disparity in education. In other words, overcoming gender disparity in education and reaping any ensuing trade and FDI benefits may come at the cost of violating other dimensions of gender equity, such as wage equity. Matthias Busse, Hamburg Institute of International Economics (HWWI) Heimhuder Str. 71, Hamburg, 20148, Germany e-mail: [email protected] Peter Nunnenkamp, Kiel Institute for the World Economy, Duesternbrooker Weg 120, Kiel, 24100, Germany e-mail: [email protected] N O T E S 1 By contrast, Günseli Berik, Yana van der Meulen Rodgers, and Joseph E. Zveglich (2004) consider openness to trade to be a determinant of gender wage gaps, finding that trade openness is inversely related to women’s relative wages in South Korean

and Taiwanese industries. 2 Avik Chakrabarti (2001) subjects the findings of various studies on FDI determinants to extreme bounds analysis and concludes that few determinants are robust to minor changes in sample selection and the specification of the test equation. 3 Restricting the sample to developing host countries is in line with Bruce A. Blonigen and Miao Grace Wang (2005), who argue strongly against pooling rich and poor countries in empirical FDI studies. Later in this paper, we will further differentiate between low- and medium-income countries within the fairly heterogeneous group of developing countries. G E N D E R D I S P A R I T Y A N D F D I 81 4 Overall, the available evidence seems to be in conflict with the hypothesis that exploiting low social standards and repressed worker rights represents an important motivation of FDI. The survey of Drusilla K. Brown (2000) concludes that poor labor practices did not attract FDI; recent studies include Phillipp Harms and Heinrich W. Ursprung (2002), David Kucera (2002), and Matthias Busse (2003, 2004).

5 Footloose industries are not tied to any location but tend to move from country to country following government incentives and/or low wages. 6 The point made by Shatz (2003) and Matthias Busse, Jens Königer, and Peter Nunnenkamp (2008) about sample selection (see below) suggests a further twist to this debate. While multinational companies may shy away from countries that do not pass a basic threshold in terms of social standards and gender equality, companies may exploit cost advantages once this threshold is passed. 7 As noted by Kucera (2002), labor costs tend to decline when some groups of workers are paid less than others for similarly productive work due to discrimination. 8 Moreover, Remco H. Oostendorp (2004) stresses the heterogeneous format of available wage data. 9 Oostendorp (2004) provides a major exception. 10 Note that Shatz (2003) argues against panel analyses on education-related determinants of FDI, as he suspects variation over time to be marginal. 11 See Assaf Razin and Efraim Sadka (2007) for an overview of the relevant literature. 12 As Shatz notes: ‘‘national statistical agencies publish bilateral data about the investment activities of their multinationals only for host

countries that have sizeable inflows of FDI. This means that nearly all research on foreign direct investment focuses on the winners, countries that have achieved at least some success in attracting FDI. This is a major problem since policy advice is most often sought by the countries that are excluded from analysis’’ (2003: 118). 13 By replicating the regressions for specific sub-groups of countries, we assess the sensitivity of results with respect to sample selection, while the extreme bounds analysis of Chakrabarti (2001) is particularly suited to assess the sensitivity of results with respect to variable selection. 14 See Eric Neumayer (2002) for a more detailed discussion of alternative approaches. 15 We divert from the model by Carr, Markusen, and Maskus (2001) in that we use additional control variables. We do not include the interactive terms used by them. 16 A Hausman test showed that there is no clear preference for the random- or fixed- effects model. Depending on the dependent variable or host country sample, we prefer either a random- or a fixed-effects model. 17 Note that bilateral FDI flows take negative values if the source country divests in a particular host country (for example, through selling its equity share to local firms and transferring the proceeds back home). We keep negative

values with respect to FDI1. However, the results for FDI1 hardly change if we exclude negative values. By contrast, negative bilateral flows are set equal to zero when calculating the share variable FDI2. This helps us include as many observations as possible, while avoiding the somewhat odd notion of negative FDI shares. 18 In contrast to M&As, greenfield FDI creates new or additional assets. 19 While a standard Durbin-Watson test showed that we do not necessarily have (first- order serial) correlation in the regressions, we cannot reject the hypothesis of no correlation either. In fact, the evidence is inconclusive. 20 The growth rate of GDP may suffer from endogeneity, as FDI inflows could have an impact on it. In the present context, however, we are not particularly interested in an unbiased estimate of the coefficient on GDP growth. Crucially, any bias in this respect is unlikely to affect the coefficient on our educational indicators – that is, the main interest of the present empirical analysis. A R T I C L E S 82 21 The data have principally been taken from Robert J. Barro and Jong-Wha Lee (2001).

We extended their dataset with more recent figures from United Nations Education, Scientific, and Cultural Organization (UNESCO 2007) to ensure that we can run a panel analysis up to the year 2004. We also performed estimations with average years of schooling for the age group of 15 and above. Unreported results proved very similar to those reported below. The results for the age group of 25 and above may be more reliable, however. This is because average years of schooling for this age group would hardly be affected, even if FDI flows had an impact on the educational attainment of younger cohorts. We owe the point that endogeneity problems may be mitigated in this way to the guest editors of this volume. 22 FDI flows to financial offshore centers can hardly be explained in the context of a gravity model that does not capture tax-related motivations of FDI; including financial offshore centers may thus lead to biased estimation results. We exclude all countries that are on the list of offshore financial centers as reported by Eurostat (2005). For a discussion on tax-induced distortions in international capital flows, see Organisation for Economic Co-operation and Development (OECD 2000). 23 Since we use the 2005 World Bank definition for the distinction between developing und developed countries, economies like Taiwan and the Republic of Korea fall into the latter category. While this has not been the case for the entire 1978–2004 period,

our results do not change much if both countries are treated as developing countries. 24 Obviously, greater openness to trade encourages trade in finished goods, too. In contrast to trade in intermediates, however, the effect of more trade in final goods on FDI flows tends to be ambiguous. This is because the removal of trade barriers for finished goods reduces the incentive to undertake FDI of the ‘‘tariff jumping’’ kind to penetrate protected host-country markets. 25 The results for the remaining variables do not change much if inflation and other insignificant variables are excluded from the analysis. Yet, we include them as they could have an impact on FDI from a theoretical point of view. 26 Note that the mean of 1.06 for FDI1, reported in the Appendix, has to be changed using the reversed transformation equation (3), which results in 1.27. The 2.5 percent increase in the dependent variable then results from the product of -0.128 and -0.25 divided by 1.27. 27 Obviously, it would be desirable to control for the wages of skilled and unskilled labor in our estimations. However, the data situation does not allow us to do so. 28 The results for the control variables are essentially unchanged. Therefore, they are not discussed here in any detail.

29 Another reason for different results is that Shatz (2003) performs a pure cross-country analysis, whereas our findings are based on a panel analysis. 30 Complete results are available from the authors upon request. 31 See Gaute Ellingsen, Winfried Likumahuwa, and Peter Nunnenkamp (2006) for a recent analysis of the case of Singapore in this context. 32 Probably, the same applies to resource-seeking FDI undertaken by developing source countries such as China in low-income regions, notably in Africa. 33 See, for example, the extensive survey of the relevant literature by Robert E. Lipsey (2002). R E F E R E N C E S Barro, Robert J. and Jong-Wha Lee. 2001. ‘‘International Data on Education Attainment: Updates and Implications.’’ Oxford Economic Papers 53(3): 541–63. Berik, Günseli, Yana van der Meulen Rodgers, and Joseph E. Zveglich. 2004. ‘‘International Trade and Gender Wage Discrimination: Evidence from East Asia.’’ Review of Development Economics 8(2): 237–54. G E N D E R D I S P A R I T Y A N D F D I 83

Blonigen, Bruce A. and Miao Grace Wang. 2005. ‘‘Inappropriate Pooling of Wealthy and Poor Countries in Empirical FDI Studies,’’ in Theodore H. Moran, Edward M. Graham, and Magnus Blomström, eds. Does Foreign Direct Investment Promote Development? pp. 221–44. Washington, DC: Institute for International Economics. Borensztein, Eduardo, José De Gregorio, and Jong-Wha Lee. 1998. ‘‘How Does Foreign Direct Investment Affect Economic Growth?’’ Journal of International Economics 45(1): 115–35. Braunstein, Elissa. 2002. ‘‘Gender, FDI and Women’s Autonomy: A Research Note on Empirical Analysis.’’ University of Massachusetts Amherst. http://www.peri.umass. edu/fileadmin/pdf/gls_conf/glw_braunstein.pdf (accessed April 2007). ———. 2006. ‘‘Foreign Direct Investment, Development and Gender Equity: A Review of Research and Policy.’’ Occasional Paper 12, United Nations Research Institute for Social Development, Geneva. Brown, Drusilla K. 2000. ‘‘International Trade and Core Labour Standards: A Survey of the Recent Literature.’’ Occasional Paper 43, Organisation for Economic Co- operation and Development. Busse, Matthias. 2003. ‘‘Do Transnational Corporations Care

About Labour Standards?’’ Journal of Developing Areas 36(2): 49–68. ———. 2004. ‘‘Transnational Corporations and Repression of Political Rights and Civil Liberties: An Empirical Analysis.’’ Kyklos 57(1): 45–66. Busse, Matthias and Christian Spielmann. 2006. ‘‘Gender Inequality and Trade.’’ Review of International Economics 14(3): 362–79. Busse, Matthias, Jens Königer, and Peter Nunnenkamp. 2008. ‘‘FDI Promotion through Bilateral Investment Treaties: More Than a Bit?’’ Kiel Working Paper 1403, Kiel Institute for the World Economy. Carkovic, Maria and Ross Levine. 2005. ‘‘Does Foreign Direct Investment Accelerate Economic Growth?’’ in Theodore H. Moran, Edward M. Graham, and Magnus Blomström, eds. Does Foreign Direct Investment Promote Development? pp. 195–220. Washington, DC: Institute for International Economics. Carr, David L., James R. Markusen, and Keith E. Maskus. 2001. ‘‘Estimating the Knowledge-Capital Model of the Multinational Enterprise.’’ American Economic Review 91(3): 693–708. Chakrabarti, Avik. 2001. ‘‘The Determinants of Foreign Direct Investment: Sensitivity Analyses of Cross-country Regressions.’’ Kyklos 54(1): 89– 113.

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d U N E S C O (2 0 0 7 ) A R T I C L E S 88 Descriptive statistics Source country sample Argentina, Australia, Austria, Belgium-Luxembourg, Brazil, Chile, Colombia, Denmark, Finland, France, Germany, Iceland, Japan, Republic of Korea, Malaysia, Mexico, Netherlands, New Zealand, Portugal, Spain, Sweden, Switzerland, Taiwan, Thailand, Turkey, United Kingdom, United States, Venezuela

Note: Developing source countries in italics. Variable Observations Mean Std. Dev. Minimum Maximum ln (FDI1) 9,743 1.06 2.51 -10.61 11.53 ln (FDI2) 9,743 0.35 0.49 0.00 5.30 ln (population) 9,743 16.44 0.14 13.21 20.98 ln (DiffGDPpc) 9,743 8.94 1.55 -9.92 11.21 Growth 9,743 3.58 2.59 -9.90 16.38 ln (inflation) 9,743 2.99 1.09 -2.45 9.44 Openness 9,743 68.02 13.12 9.31 230.27 Common border 9,743 0.02 0.13 0.00 1.00 Common language 9,743 0.13 0.33 0.00 1.00 ln (distance) 9,743 8.95 0.63 5.21 9.89 Colonial ties 9,743 0.03 0.18 0.00 1.00 RTA 9,743 0.04 0.11 0.00 1.00 Political constraints 9,743 0.27 0.20 0.00 0.68 CapOpen 9,743 -0.20 0.94 -1.75 2.62 BIT 9,743 0.17 0.20 0.00 1.00 Total education 9,743 4.48 0.61 0.30 10.31 Total schooling inequality 9,743 1.06 0.25 -1.22 3.34 Primary education 9,743 3.15 0.35 0.24 7.88 Primary education inequality 9,743 0.65 0.20 -1.04 2.30 Secondary education 9,743 1.12 0.24 0.04 3.27 Secondary education inequality 9,743 0.32 0.09 -0.48 1.27

Tertiary education 9,743 0.22 0.06 0.01 0.84 Tertiary education inequality 9,743 0.08 0.02 -0.17 0.37 G E N D E R D I S P A R I T Y A N D F D I 89 Host country sample Albania, Algeria, Angola, Argentina, Bangladesh, Bolivia, Botswana, Brazil, Bulgaria, Burkina Faso, Cameroon, Chile, China, Colombia, Republic of Congo, Costa Rica, Côte d’Ivoire, Croatia, Dominican Republic, Ecuador, Egypt, El Salvador, Estonia, Ethiopia, Gambia, Ghana, Guatemala, Guyana, Haiti, Honduras, Hungary, India, Indonesia, Jordan, Kazakhstan, Kenya, Latvia, Lithuania, Malaysia, Mali, Mauritius, Mexico, Mongolia, Mozambique, Namibia, Nicaragua, Niger, Nigeria, Oman, Pakistan, Papua New Guinea, Paraguay, Peru, Philippines, Poland, Romania, Russian Federation, Saudi Arabia, Senegal, Slovenia, Sri Lanka, Sudan, Swaziland, Syrian Arab Republic, Tanzania, Thailand, Togo, Trinidad and Tobago, Tunisia, Turkey, Uganda, Ukraine, Uruguay, Venezuela, Vietnam, Zambia, Zimbabwe A R T I C L E S

90 Stronger Database Security Needed, Cyber-Attacks Show http://www.cioinsight.com/print/c/a/Latest-News/CyberAttacks- Highlight-Need-to-Focus-on-Stronger-Database-Security- 342260[2/2/2017 9:31:02 AM] Stronger Database Security Needed, Cyber-Attacks Show By CIOinsight | Posted 06-03-2011 When cyber-attackers breach an organization's network, the database is usually their target. However, many organizations are so focused on protecting the perimeter that they don't think abo protecting the database itself, according to several security experts. Many organizations still think that protecting the perimeter is sufficient to protect the data, but as recent data breaches at Epsilon and Sony have shown, traditional perimeter security can't relied on to protect the data, Josh Shaul, CTO of Application Security, told eWEEK. It's a "losing battle" to try to protect every single endpoint within the organization, Shaul said. That's not to suggest that organizations shouldn't be investing in firewalls and other security products. Shaul recommended the layered model, where attackers have to get past multip gatekeepers before they even get to the database. Organizations should be thinking, "When the perimeter fails, what's next?" and combining all the layers to pinpoint when something is wron according to Shaul.

It's ironic that "the closer we get to the data, we see fewer preventive controls and more detection measures," Shaul said. IT departments are more likely to have deployed products that send o alerts that a breach has occurred, than ones that actively block the threat from getting in to the database. Most blocking technologies are still deployed on the perimeter, according Shaul. Organizations still assume that all activity hitting the database is "untrusted," Shaul said. Instead, they should monitor all requests to figure out whether the activity is normal or malicious. Continuous, real-time monitoring is crucial to detect suspicious or unauthorized activity within the database, Phil Neray, vice president of data security strategy and information management IBM, told eWEEK. Database activity monitoring allows security managers to catch anyone who is trying to get access to information they shouldn't be able to obtain. To read the original eWeek article, click here: Cyber-Attacks Highlight Need to Focus on Stronger Database Security. http://www.cioinsight.com/c/a/Security/Epsilon-Data-Breach- Hits-Banks-Retail-Giants-154971/ http://www.cioinsight.com/c/a/Security/Sony-Networks-Lacked- Firewall-Ran-Obsolete-Software-Testimony-103450/ http://www.eweek.com/c/a/Security/CyberAttacks-Highlight- Need-to-Focus-on-Stronger-Database-Security- 342260/cioinsight.comStronger Database Security Needed, Cyber-Attacks Show

· WEEK 1 Databases and SecurityLesson· Databases and Security.docx

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