1.

The Effect of Individualism on Opportunism Propensity

in International Strategic Alliances

Olivier Furrer, Radboud University Nijmegen

Brian Tjemkes, VU University Nijmegen

Abstract

The objective of this study is to examine the effect of cultural values on opportunistic

propensity in strategic alliances. Alliance relationships constitute mixed-motive ventures

which are often plagued with opportunism. However, opportunistic propensity may not be as

universal as currently described in the literature. More specifically, though previous alliance

studies investigated opportunistic propensity and suggested that cultural values may affect a

manager’s likelihood to act opportunistically three issues permeate current literature:

empirical tests are virtually absent, studies have primarily used country-level data, and studies

tend to neglect the impact of situational factors. To address these issues, we collected survey

data in the Netherlands and Turkey from alliance managers and empirically examine the

moderating effect of individualism on the relationships between four exchange variables and

opportunistic propensity. The results demonstrate that (1) the effects of economic

dissatisfaction and alliance specific investments are stronger for managers with individualist

values, whereas (2) the effects of social dissatisfaction and alternative attractiveness are

stronger for mangers with collectivistic values. Thus, we advance the international strategic

alliance literature by showing that opportunism possesses a similar meaning across the two

countries, but in addition that opportunistic propensity is affected interactively by cultural

values and exchange variables.

Keywords: Opportunism, strategic alliances, cultural values

2.

INTRODUCTION

International strategic alliances are voluntary, long-term, contractual, cross-border

relationships between two firms, designed to achieve specific objectives through collaboration

(Brouthers & Bamossy, 2006). However, international strategic alliances are also mixed-

motive ventures in which partners cooperate and compete simultaneously (Kumar & Nti,

2004). This simultaneity opens the door to opportunism, which is then likely to influence

alliances’ evolution and performance (Das & Rahman, 2010), resulting in the high failure rate

of international strategic alliances often (Park & Ungson, 2001).

The marriage of firms from different cultures creates an additional potential for

opportunism, conflict, and mistrust (Johnson, Cullen, Sakano, & Takenouchi, 1996). To

insure the success of an international strategic alliance, trust between partners is crucial (Das

& Teng, 2001). As demonstrated by Johnson and colleagues (1996), a lack of cultural

sensitivity affects trust building between partners. Thus, one of the key drivers of

opportunism in international strategic alliances is a lack of sensitivity to cultural differences

while managing the alliance (Kumar & Nti, 2004). Alliance partners often manage the

alliance based on their own frame of reference and cultural values with the implicit

assumption that opportunism is the same across cultures. This issue has been identified by a

few scholars who recognized a need for the identification of the conditions under which

opportunism is most likely to occur (Chen, Peng, & Saparito, 2002; Maitland, Bryson, & Van

de Ven, 1985). However, implicit universalism still pervades much opportunism research

(Boyacigiller & Adler, 1991). If opportunism, as an economic factor, is likely to be universal,

the opportunistic propensity of alliance partners, which is a human factor, is likely to be

influenced by cultural values.

There are some indications that cultural values, and in particular individualism and

collectivism, influence people’s opportunistic propensity in general (Chen et al., 2002; Furrer

3.

et al., 2011; Sakalaki, Kazi, & Karamanoli, 2007), as well as in strategic alliances (Johnson et

al., 1996; Tjemkes et al., 2011). For example, Chen and colleagues (2002) argued that

individualists have a higher opportunistic propensity in intra-group transactions and

collectivists in inter-group transactions. Furrer and colleagues (2011) also suggested that

opportunism could be perceived as more morally wrong in some cultures than in others.

However, as Chen and colleagues (2002), they did not empirically test their propositions.

Surprisingly then, there is still limited understanding of how cultural values influences

international strategic alliance partners’ opportunistic propensity.

Beside these conceptual studies, only a small number of empirical studies started to

investigate differences in opportunism across cultures (Furrer et al. 2011; Johnson, Cullen and

Sakano, 1996; Sakalaki, Kazi, & Karamanoli, 2007; Tjemkes et al., 2011). They all found

cross-cultural differences in alliance partners’ opportunistic propensity. However, these

studies only assessed cultural differences at the country level, neglecting within-country

differences. Advances in cross-cultural research (e.g., Au, 2000; Au & Cheung, 2004)

demonstrated the importance of taking into account within-country differences, as individual-

level cultural differences are often better able to explain behaviors and behavioral intentions

than societal-level differences (e.g., Ralston et al., 2009). This is, cultural values often have a

more significant effect than national culture.

Another limitation of previous studies investigating opportunism across cultures is that

they only investigate the direct effect of culture, neglecting the mechanisms through which

culture influences opportunistic propensity. However, the results of the study by Tjemkes et

al. (2011) suggest that culture not only directly influence opportunistic propensity but also

moderate the effect of antecedents, such as satisfaction and exit barriers. Building on strategic

issue categorization theory (Dutton & Jackson, 1987) and its cross-cultural developments

(Kumar & Nti, 2004; Sallivan & Nonaka, 1988; Schneider & Meyer, 1991), it is likely that

4.

cultural values influence how managers interpret signals from their external environment and

thus moderate the effect of these environmental factors on their likelihood to behave

opportunistically.

In this study, we empirically investigate the moderating effect of individual-level

individualism on the relationship between situational factors and opportunistic propensity

with a survey of a sample of Dutch and Turkish alliance managers. Specifically, we

hypothesize that individualism moderates the effects of four situational factors indentified by

economic and social exchange theories: economic dissatisfaction, social dissatisfaction,

alliance-specific investments, and the alternative attractiveness. We focus on individualism as

it dominates cross-cultural research and is perhaps the most commonly used value dimension

to explain cultural differences (Hofstede, 2001; Triandis, 1995). Furthermore, Chen and

colleagues(2002) argued that opportunistic propensity is affected by individualism and

Sakalaki, Kazi and Karamanoli (2007) empirically demonstrated that individualism is

significantly related to opportunistic propensity. In addition, we measure individualism at the

individual level and control for country effects. By doing so, we are able to disentangle the

effects of individualism values from those of other national-level factors, such as economic

development and institutions. By doing so, we empirically demonstrate that if opportunism is

a universal construct, opportunistic propensity is affected by cultural values.

Addressing questions of the cross-cultural generalizability of opportunism is

fundamental to combat the implicit universalism that pervades much organizational and

strategic research (Boyacigiller & Adler, 1991; Thomas & Au, 2002). Thus, we contribute to

the debate between universalist and relativist approaches in cross-cultural management

research by demonstrating that if opportunism is a universal construct, people’s opportunistic

propensity is culturally influenced. Furthermore, our study also advances the literature on the

management of international strategic alliances. As opportunism and even the assumption of

5.

opportunism have the potential of negatively affecting the performance of international

strategic alliances (Luo, 2007). Indeed, assuming the opportunism of one’s partner is likely to

become a self-fulfilling prophecy (Ghoshal & Moran, 1996). As demonstrated by Davis,

Schoorman and Donaldson (1997), the assumption of opportunism is likely to trigger the

development of control mechanisms, which implementation is likely to frustrate one’s partner

who is likely to feel betrayed and to start behaving opportunistically in retaliation. John

(1984) also found empirical evidence that bureaucratic control can damage trust and

exacerbate opportunism in interpartner relationships. Thus, in international strategic alliance,

managers’ awareness of cultural differences in opportunistic propensity becomes critical to

establish fair control mechanisms and instill trust and benevolence, which are important for

the success of international strategic alliances (Das & Teng, 2001).

We organize the remainder of this article as follows: In the next section, we define and

distinguish between opportunism as a behavior and opportunistic propensity in strategic

alliances and their antecedent. Then, we review the embryonic literature cross-national

variations in opportunistic propensity. We then discuss the effect of individualism at the

individual-level on opportunistic propensity and develop hypotheses about its moderating

effect on the relationship between exchange variables and opportunistic propensity. In the

method section, we describe the sample and the design of the survey we use to test the

hypotheses. Finally, we present the results and conclude with a discussion of the theoretical

and managerial implications of our study, along with limitations and directions for further

research.

THEORETICAL BACKGROUND AND HYPOTHESES

Opportunism and opportunistic propensity in strategic alliances

In light of its specific strategic objectives for entering in an international strategic alliance and

resources commitments, each firm seeks to optimize its individual position. Because of the

6.

high stakes typically involved, alliance partners may be motivated to behave opportunistically

or take an opportunistic view in managing the alliance (Johnson et al., 1996). International

strategic alliances are mixed-motive ventures in which partners cooperate and compete

simultaneously (Kumar & Nti, 2004). Therefore, opportunism is likely to occur when the

expected gains from behaving opportunistically exceed potential payoffs from forestalling

malfeasance (Axelrod, 1986). To inhibit opportunism and safeguard firms’ individual

investment (Beamish & Banks, 1987), partners in international strategic alliances often

establish mutual hostage situations (Kogut, 1988) and contractual safeguards (Deeds & Hill,

1999). However, alliance partners still represent separate self-interested constituencies with

their own individual objectives. Thus, even though alliance partners succeed in internalizing

transactions, and therefore reduce opportunism, to some extent, opportunism remains present

and problematic in international strategic alliances (Johnson et al., 1996).

The most often used definition of opportunism was put forth by Williamson (1985:

47–48):

By opportunism I mean self-interest seeking with guile. This includes but is scarcely

limited to more blatant forms, such as lying, stealing, and cheating. Opportunism often

involves subtle forms of deceit […]. More generally, opportunism refers to the

incomplete or distorted disclosure of information, especially to calculated efforts to

mislead, distort, disguise, obfuscate, or otherwise confuse. It is responsible for real or

contrived conditions of information asymmetry, which complicate problems of

economic organization.

Despite this behavioral definition, Williamson used the term opportunism both in the

sense if an attitude (i.e., proclivity, inclination, and propensity) and in the sense of a behavior,

which should be distinguished (Ghoshal & Moran, 1996). For example, he refers to

opportunistic attitudes, which he presents as one of the rudimentary attributes of human

7.

nature. At the same time, he sees opportunism as behavior such as lying, stealing, and

cheating. In strategic alliance, opportunistic behavior involves several elements: (i) distortion

of information, including overt behaviors such as lying, cheating and stealing, as well as more

subtle behaviors such as misrepresenting information by not fully disclosing. (ii) reneging on

explicit or implicit commitments such as shirking, or failing to fulfill promises, and

obligations (Wathne & Heide, 2000). Opportunistic propensity, as a behavioral tendency,

represents the attitude (i.e., proclivity, inclination) to act opportunistically (Ghoshal & Moran,

1996). Although, opportunistic behavior is assumed to be universal and triggered by

economic situational and structural factors and exchange variables (Williamson, 1985),

opportunistic propensity is ultimately caused by the nexus of a given human nature of self-

interest with certain cultural norms and values (Chen et al., 2002).

Extent strategic alliance studies examined the conditions leading to increased and

decreased opportunism (e.g., Das & Rahman, 2001; Deeds & Hill, 1999; Judge & Dooley,

2006; Luo, 2007). Building on transaction cost economics (Williamson, 1985) and social

exchange theory (Blau, 1964), empirical studies identified four exchange variables affecting

opportunism: economic and social satisfaction, alliance specific investments and alternative

availability. For example, Das & Rahman (2001) explain that a partner may perceive its share

of reward from the alliance to be inequitable, and feel economically dissatisfied, which will

motivate it to restore a sense of equity by any mean possible, including opportunism. Judge

and Dooley (2006) empirically found that alliance performance and partner trustworthiness

were negatively related to opportunistic behavior. Luo (2007) also demonstrated that

environmental turbulence putting at risk the performance of the alliance increases partners’

opportunism. Deeds & Hill (1999) also found that a good, socially satisfying working

relationship with an alliance partner reduces the likelihood of opportunism. Investigating

deterrence mechanisms, Das and Rahman (2001) found that the presence of alliance-specific

8.

investments decrease partners’ opportunism. Similarly, the availability of attractive alternative

partners outside the alliance increases opportunism (Luo & Shenkar, 2002).

Cross-national variations in opportunistic propensity

People from different cultures have different preferences for dealing with similar set of

problems. These different preferences are described by Kluckhohn and Strodtbeck (1961) as

variations in value orientations, which derive from assumptions regarding relationships with

the environment as well as relationships among people (Schneider & Meyer, 1991). These

value orientations or cultural values influence the way people perceive, think, feel, and

evaluate, and thus affect the process by which the environment is “known” and responded to

(Hofstede, 2001). As such, culture plays an important role in strategic decision making,

including in international strategic alliances.

A few studies seem, indeed, to indicate that, at the country level, culture influences

opportunistic propensity. For example, in a conceptual study, Chen and colleagues (2002)

suggest that opportunistic propensity is affected by cultural prior conditioning of

individualism. Specifically, they argue that individualists have a higher opportunistic

propensity in intra-group transactions, and collectivists in inter-group transactions. In

addition, they also propose that when there is a conflict of interest between an in-group and an

out-group (like in the case of a international strategic alliance), collectivists will have a

greater opportunistic propensity on behalf of the in-group than will individualists. In their

study of the cross-cultural validity of a circumplex model of response strategies, Furrer and

colleagues (2011) found that whereas six of the response strategies they studied were

universally organized along the same circular structure, opportunistic propensity deviate from

this structure in some cultures but not is others. They suggested that opportunistic propensity

could be more morally wrong in some cultures compared to others. Building on Hofstede’s

(2001) work they explain that in countries with low uncertainty avoidance, people may more

9.

tolerant of transgressions of moral norms, such as opportunism, whereas in countries with

higher uncertainty avoidance scores, such transgressions are considered morally wrong.

In an early study of opportunistic propensity in international strategic alliances

between Western or Asian and Japanese firms, at the country-level, Johnson, Cullen and

Sakano (1996) found significant differences between partners from Western cultures and the

Japanese, but not between other Asians and the Japanese. Western partners reported a smaller

propensity for opportunistic behavior than did their Japanese counterparts. In addition, they

found the opportunistic propensity did not diminish as the alliance relationship aged, which

suggests that opportunistic propensity is a stable cultural trait. In an experimental study with

business students in an international strategic alliance context, Tjemkes and colleagues (2011)

also found significant country differences in opportunistic propensity. They found that

Turkish participants reported significantly higher level of opportunistic propensity than

British, Dutch, and Swiss participants, which in turn, reported higher level of opportunistic

propensity than Japanese participants. But they also discovered that culture also interacted

with social satisfaction in influencing opportunistic propensity; whereas opportunistic

propensity increases as social dissatisfaction increases in the U.K., the Netherlands, and

Japan, it decreases in Turkey and Switzerland. These results suggest that the adversity of a

poor working relationship with one’s partner is perceived differently across cultures.

However, all these studies assessed culture at the country level neglecting important

within-country differences. To the best of our knowledge, the only study that investigates the

effect of cultural values on opportunistic propensity at the individual level is the study by

Sakalaki, Kazi and Karamanoli (2007). However, contrary to the previous studies, they

looked at relationships with in-groups rather than out-groups and found in a Greek sample

that opportunistic propensity is positively correlated with individualism and negatively with

collectivism, which is consistent with the predictions made by Chen and colleagues (2002) for

10.

intra-group relationships. Thus, these findings might not be valid in the international strategic

alliance context, which mostly involve inter-group relationships (Johnson et al., 1996;

Tjemkes et al., 2011). Indeed, Chen and colleagues (2002) argued that given that

individualists and collectivists differ in self–other relationships both within and between

groups, the effect of cultural values on opportunistic propensity is also likely to vary between

in-group and out-group transactions.

The effect of individual-level individualism on opportunistic propensity

Individualism and collectivism contrast values that focus on the individual as the most

meaningful social unit (e.g., autonomy) with those that emphasize social groups (e.g., group

norms) (Markus & Kitayama, 1991). An empirical examination of cultural values measures

developed in different parts of the world suggested that individualism and collectivism might

be basic dimensions of human values (Smith, Dugan, & Trompenaars, 1996). Thus,

individualism is likely to have an impact on interpreting and responding to adverse situations

in international strategic alliances.

Individualism has shown to exist and have different effects at the national societal-

level and at the individual-level (Ralston et al., 2009). Because, we are interested in both

between and within country differences, we develop our next hypotheses in reference to

individual-level individualism and collectivism. To minimize confusion over levels of

analysis, some researchers have encouraged the use of the term “idiocentrism” as an

individual level parallel to individualism at the country level and “allocentrism” as an

individual level parallel to collectivism (Triandis et al., 1985). However, as Smith and Bond

(1999, p. 62) note, “level-appropriate terms have not yet been adopted by other researchers.”

Therefore, following Kirkman and Shapiro (2001), we retain the terms individualism and

collectivism but caution the reader that we are referring to an individual-level construct.

At the individual-level, individualism refers to values that refers to a self-orientation,

11.

an emphasis on self-sufficiency and control, the pursuit of individual goals that may or may

not be consistent with in-group goals, and a willingness to confront members of the in-group

to which a person belongs (Markus & Kitayama, 1991). People high in individualism tend to

put forth and promote their own welfare over the interests of their group or organization

(Hofstede, 2001; Triandis, 1995). Individualists are motivated by self-interest and

achievement of personal goals. They are hesitant to contribute to collective action unless their

own efforts are recognized, preferring instead to benefit from the efforts of others.

Collectivism involves the subordination of personal interests to the goals of the larger work

group, an emphasis on sharing, cooperation, and group harmony, a concern with group

welfare, and hostility toward out-group members (Hofstede, 2001). Collectivists believe that

they are an indispensable part of the group, and will readily contribute without concern for

advantage being taken of them or for whether others are doing their part (Markus &

Kitayama, 1991). They feel personally responsible for the group product and are oriented

towards sharing group rewards (Kluckhohn & Strodtbeck, 1961; Triandis, 1995) and are

likely to place great emphasis on social acceptance, group identity, smooth interpersonal

relations, and close emotional ties (Grimm, Church, Katigbak, & Reyes, 1999).

Wong, Tjosvold and Yu (2005) argue that opportunism in strategic alliances can be

understood in terms of how partners conclude that their self-interests are related to each other.

When partners believe that their goals are competitively but not cooperatively related, they

are tempted to pursue their self-interests opportunistically. Individualists make decision based

motives pertaining to the protection of individual profits, as justified by utilitarian principles

(Thomas, Au, & Ravlin, 2003). They assess strategic alliances based on cost-benefits

calculations (Triandis, 1995). Therefore, opportunism is likely to occur when individualists’

expected gains from behaving opportunistically exceed potential payoffs from forestalling

malfeasance (Axelrod, 1986). Compared with individualists, collectivists are more inclined to

12.

consider their partners as out-groups (Triandis, 1995), especially if they are foreigners

(Johnson et al., 1996). Collectivistic managers in an out-group situation with foreign partners

might have more competitive goals than individualistic managers, and therefore are more

likely to behave opportunistically. Furthermore, out-group transactions present a test for

collectivists to demonstrate their willingness to self-sacrifice for the preservation of their in-

group’s interest (Chen et al., 2002). Therefore, collectivists, as faithful agents of their in-

groups, will be more willing than individualists to fight on behalf of their in-group against the

out-group, employing all possible means including guileful ones (Chen et al., 2002).

Moderating effects of individual-level individualism

As argued by Ghoshal and Moran (1996), opportunistic propensity is likely to be affected by

individual dispositions as well as by the situation that shapes the individual perceptions and

instrumentalities. In other words, individual dispositions, such as individual values, and

situational factors, such as exchange variables, are likely to interact in influencing people’s

opportunistic propensity. Strategic issue categorization theory (Dutton & Jackson, 1987), and

its cross-cultural developments (Kumar & Nti, 2004; Sallivan & Nonaka, 1988; Schneider &

Meyer, 1991), provides theoretical foundations to understand this interaction effect.

Strategic issue categorization theory (Dutton & Jackson, 1987) proposes that

managers’ cognitions and motivations systematically affect the processing of issues and the

types organizational actions taken in response to them. Specifically, Dutton and Jackson

(1987) argue that the labeling of strategic issues as either threats or opportunities by

managers’ influence their information processing and responses. This is, in a strategic alliance

context, exchange variables might either be perceived as threats or opportunities. For

example, a manager might perceive alliance-specific investments as a threat, as they increase

the dependence of their firm on its partner, or as opportunity to signal to its partner the firm’s

commitment to the alliance.

13.

In international situations, Sallivan and Nonaka (1988) and Schneider and Meyer

(1991) empirically found that the interpretation and categorization of strategic issues are

influenced by culture leading to different responses to these issues. For example, Sallivan and

Nonaka (1988), found Japanese managers more likely than American managers to interpret

issues as threats and to restrict information scanning and sharing as a function of that

interpretation. Similarly, Schneider and Meyer (1991) found that culture influence the

interpretation and response to strategic issues. Consistent with these findings, Kumar and Nti

(2004) argue that culture affect strategic alliance evolution by influencing partner’s sensitivity

to discrepancy detection, shaping the nature of attributions they make, and by affecting the

partners’ reactions to discrepancies. In addition, in a strategic alliance experimental context,

Tjemkes and colleagues (2011) found that the effect of economic satisfaction, social

satisfaction, alliance-specific investments, and alternative propensity on response strategies

was moderated by national culture.

In a different study context and at the individual-level, Thomas and colleagues (2002,

2003) also found that cultural values moderates the effect exchange variables and behavioral

responses to dissatisfaction. Empirical studies focusing on alliances identified four exchange

variables that influence opportunistic propensity: economic and social satisfaction, alliance

specific investments and alternative availability (e.g., Ping, 1993; Tjemkes & Furrer, 2010).

Thus, in an international strategic alliance, we propose that alliance managers’ cultural

individualism influences the way they perceived exchange variables moderating the effects of

these variables on their opportunistic propensity.

Moderating the effect of economic dissatisfaction

Economic dissatisfaction pertains to managers’ evaluation of the financial outcomes of an

alliance (Geyskens and Steenkamp, 2000). According to Geyskens and Steenkamp (2000), an

economically dissatisfied manager considers the alliance a failure with respect to goal

14.

attainment, effectiveness, productivity, and the resulting financial outcomes. Kumar and Nti

(2004) argue that in strategic alliances outcome discrepancy generates economic

dissatisfaction. Prior empirical studies have produced inconclusive results about the

relationship between economic dissatisfaction and opportunistic propensity. For example,

Ping (1993) and Tjemkes and Furrer (2010) hypothesized a negative relationship between

economic satisfaction and opportunistic propensity but the results of their empirical studies

were not statistically significant, which might be due to specific cultural contexts.

Achieving economic satisfaction is a more important goal for individualist managers

than collectivist managers, because strategic alliances in the former are governed by more

rational cost–benefit calculations (Triandis, 1995). Therefore, when economic dissatisfaction

increases, managers with more individualistic values are more likely than their counterparts

with collectivist values to be opportunistic. In contrast, managers with collectivist values are

less sensitive to changes in the economic outcomes of the alliance, as the quality of the

relationship with their partner is more important than its short-term financial outcomes

(Hofstede, 2001). Therefore, we hypothesize:

Hypothesis 1: In strategic alliances, the positive effect of economic dissatisfaction

on managers’ opportunistic propensity is stronger the more his/her

values are individualistic.

Moderating the effect of social dissatisfaction

Social dissatisfaction pertains to managers’ negative evaluations of the psychosocial aspects

of an alliance (Tjemkes and Furrer, 2010); it implies that interactions with counterparts are

problematic (Anderson & Narus, 1990) and lacking in transparent communication (Ariño, De

la Torre, & Ring, 2001). Socially dissatisfying relationships are also characterized by negative

criticisms and dishonesty (Geyskens & Steenkamp, 2000). Managers’ perceptions of

relational quality affect their social satisfaction; if relational quality is poor, the alliance

15.

suffers dysfunctional conflicts, distrust, and low commitment (Ariño et al., 2001). Similarly,

Kumar and Nti (2004) argue that, process discrepancy increases social dissatisfaction. High

social dissatisfaction creates greater suspicion about a counterpart’s intentions and reduces

expectations about the potential future benefits of the relationship (Geyskens & Steenkamp,

2000). Partners dissatisfied with the relationship become less worried about endangering the

relationship and may act opportunistically to extract additional benefits (Deeds and Hill,

1999). In a strategic alliance context, experimental results suggest that social satisfaction

reduces managers’ opportunistic propensity (Tjemkes and Furrer, 2010).

Compared to individualists, people with collectivist values feel personally responsible

for the group product and are oriented towards sharing group rewards (Kluckhohn &

Strodtbeck, 1961; Triandis, 1995). They are also more likely to place great emphasis on social

acceptance, group identity, smooth interpersonal relations, and close emotional ties (Grimm et

al., 1999). Therefore, for managers with more collectivist values, social dissatisfaction should

have a stronger effect on opportunistic propensity than it does more managers with more

individualistic values. As social dissatisfaction increases, managers with collectivist values,

who value consensus and close relationships (Hofstede, 2001), respond more

opportunistically, compared to managers with more individualistic values who are less

sensitive to personal relationships and social dissatisfaction. Thus:

Hypothesis 2: In strategic alliances, the positive effect of social dissatisfaction on

managers’ opportunistic propensity is stronger the more his/her

values are collectivist.

Moderating the effect of alliance-specific investments

Alliance-specific investments represent sunk costs that cannot be redeployed easily to another

alliance without some sacrifice in the productivity of the assets or cost to adapt them (Ping,

1993). These investments would be lost if the alliance were dissolved, so they act as exit

16.

barriers. Their presence constitutes a source of dependence for the firm that makes them,

which implies an adverse situation for managers who need to reduce the negative

consequences of their firms’ vulnerable position. High levels of alliance-specific investments

encourage managers to work cooperatively with their partner to resolve any problems to

maintain the relationship (Gulati, Khanna, & Nohria, 1994). Conversely, by increasing the

costs of terminating the alliance, alliance-specific investments reduce the likelihood of any

action that could prompt the partner to exit, such as opportunism (Deeds and Hill, 1999).

Compared to collectivist managers, individualist managers are more likely to rely on

rational cost–benefit calculations in managing their strategic alliances (Triandis, 1995).

Therefore, individualist managers are likely to be more sensitive to alliance-specific

investments than alliance managers with collectivist values. This is because if their

opportunism is detected, they are more likely to fear the retaliation of their partner (John,

1984), which could lead to the loss of the investments. Thus, as the amount of alliance-

specific investments increases, individualists are less likely than collectivists to be

opportunistic to safeguard these investments (Beamish & Banks, 1987). Therefore, for

managers with more individualist values, alliance-specific investments should have a stronger

negative effect on opportunistic propensity than it does for more managers with more

collectivistic values. Thus:

Hypothesis 3: In strategic alliances, the negative effect of alliance-specific

investments on managers’ opportunistic propensity is stronger the

more his/her values are individualistic.

Moderating the effect of alternative availability

Alternative availability refers to the extent to which the firm possesses attractive alternatives

outside the alliance that could enable it to attain its objectives (Ping, 1993). The presence of

attractive alternatives provides firms with a source of power, whereas a dearth of alternatives

17.

increases dependence on counterparts. In a situation without alternatives, alliance managers

have strong incentives to make the current alliance work and are less likely to endanger the

relationship by acting opportunistically (Buchanan, 1992). On the other hand, Peng (1993)

and Tjemkes and Furrer (2010) demonstrated that when managers perceive that they have

other alternatives for achieving their objectives, and they depend less on the current

relationship, they are more likely to act opportunistically.

People with collectivistic values better tolerate dependence than people with

individualistic values, which push then to be independent and autonomous. Thus, in strategic

alliances, individualistic managers without attractive alternatives might feel threatened by the

risk that their counterpart will exit the alliance; to reduce their dependence (Thomas and Au,

2002). Collectivist managers instead are used to depending on their group and therefore might

feel less threatened by a dependence situation created by a lack of alternatives. Therefore,

they are likely to be less influenced by the existence or absence of alternatives, because they

do not consider the situation especially threatening. Thus, for collectivists, the absence of

alternatives deters them less than individualists from opportunism. As the availability of

attractive alternatives increases, managers with collectivistic values will respond more

opportunistically, compared to managers with more individualistic values whose opportunistic

propensity is less likely to be influenced by the increasing presence of alternatives. Thus, we

hypothesize:

Hypothesis 4: In strategic alliances, the positive effect of alternative availability on

managers’ opportunistic propensity is stronger the more his/her

values are collectivist.

METHOD

To test our hypotheses, we developed a survey and collected data from Dutch and Turkish

alliance managers.

18.

Country selection and samples

We chose for Turkey and the Netherlands because they face opposite positions on the indexes

of Hofstede’s individual-collectivist cultural dimension. For individualism are the scores 80

for the Netherlands 80 and 37 for Turkey. Though, the Netherlands and Turkey differ on the

other cultural dimensions the relative distance is smaller. Power distance has scores of 66 for

Turkey and 38 for the Netherlands., the Netherlands has a score of 14 and Turkey of 45 on

masculinity, and scores of 53 for the Netherlands and 85 for Turkey on uncertainty avoidance

(Hofstede, 2001).

From the alliance managers we contacted in the Netherlands we obtained 248 valid

questionnaires whereas alliance managers in Turkey returned 171 questionnaires. We ensured

that any data from non-native respondents were dropped from the study (n = 14) and after

accounting for incomplete questionnaires, the response for Turkey is 157 valid questionnaires.

No significant differences in organizational characteristics emerged between early and later

respondents (Armstrong & Overton, 1977).

In the Netherlands, respondents work for firms in three main industries:

production/manufacturing, business services, and construction. On average, these firms had

4,400 employees (standard deviation [SD] = 18,203) and had managed 17 alliance

relationships (SD = 49,4) in the past five years. The average duration of an alliance was 7.1

years (SD = 8.9). The respondents, mostly male managers (229, or 93.5%), were 46.6 years

(SD = 7.7) and had 10.2 years of experience with alliance relationships (SD = 7.2) on average.

In Turkey, respondents work for firms primarily in the business service industry. On average,

these firms had 44 employees (standard deviation [SD] = 62.6) and had managed 21.7

alliances (SD = 66.7) in the past five years. The average duration of an alliance was 7.4 years

(SD = 4.6). The respondents, mostly male managers (125, or 79.6%), were 40.7 years (SD =

8.9) and had 13.0 years of experience with alliance relationships (SD = 7.1) on average.

19.

Study setting and procedure

We asked these respondents to read two screening questions before participating in the project

to exclude any who could not provide the necessary information. The first screening question

asked respondents to select a strategic alliance that involved no equity; thus, we decreased the

likelihood that governance forms other than contractual alliances, such as joint ventures,

appeared in the final data set. Then we asked them to select a long-term alliance formalized

with a contract, to steer them away from short-term, transactional relationships (Dwyer,

Schurr, & Oh, 1987). Thus, only strategic alliances consistent with the scope of our research

are included in the final sample.

Consistent with previous alliance research (e.g., Johnson et al., 1996), we used key

informants and collected data from only one side of the dyadic relationship. This choice

offered several advantages, including increased response rates, fewer resources, and relatively

faster and easier data collection. However, one-sided key informants also produce noisy data

as a result of selection and perceptual biases (Kumar, Stern, & Anderson, 1993). To reduce

concerns about these response biases, we asked two questions to ensure the respondent (1)

was knowledgeable about the strategic alliance under investigation and (2) possessed

decision-making authority. On a seven-point Likert scale, with a cutoff value of 3, a low score

indicated the respondent had less knowledge about the alliance; only three informants in the

Netherlands and zero in Turkey did not meet these criteria and were eliminated from further

analysis resulting in a final sample of 245 for the Netherlands. The average scores across the

sample were 5.4 (SD = 1.6) for knowledge possession and 5.8 (SD = 1.2) for decision-making

authority, comparable to the levels in prior research (Lambe, Spekman, & Hunt, 2002), which

suggests that we used appropriate respondents for our data analysis.

Measures

The questionnaire was developed in English that we translated into Dutch and Turkish, using

20.

standard translation–back translation procedures (Brislin, 1986). To measure managers’

opportunistic propensity, we adapted existing scales from John (1984) and Ping (1993) that

refers to withholding information, exaggerating the averse nature of the situation, and seeking

to escape contractual obligations. The seven-point Likert scales ranged from (1) “I would

definitely not react in this way” to (7) “I would definitely react in this way.”

Because Hofstede’s cultural value measures were designed to be used only for

country-level analyses, we chose Yoo, Donthu and Lenartowicz’s (1998) cultural-value

measures that were constructed specifically for use at the individual level of analysis. We use

the Individual Cultural Values Scale (CVSCALE) (Yoo et al., 2011) to assess the cultural

values of individualism/collectivism. The CVSCALE has shown to possess good reliability

and validity and to be cross-cultural invariant (Patterson, Cowley, & Prasongsukarn, 2006;

Yoo et al., 2011). All four items were measured on seven-point Likert scales, ranging from (1)

strongly disagree to (7) strongly agree.

To measure economic satisfaction, social dissatisfaction, alliance specific investments

and availability of alternatives we built on prior work (Geyskens & Steenkamp, 2000; Ping,

1993; Tjemkes & Furrer, 2010). Economic dissatisfaction is measured by indicating the extent

to which a manager is financially dissatisfied with the alliance, with four items measuring

managers’ level of satisfaction with the alliance with regard to profit, performance,

achievement of goals, and efficiency (Geyskens & Steenkamp, 2000). The four social

dissatisfaction items measure the extent to which the interaction between the partners is

perceived as complicated, unfulfilling, and disappointing (Geyskens & Steenkamp, 2000).

The presence of relationship-specific investments increases costs of terminating it by creating

a potential hold-up situation (e.g., Ping, 1999; Williamson, 1985), whereas the presence of

attractive alternatives decreases the risks of terminating the relationship. Alternative

availability and relationship-specific investments are each operationalized by four items

21.

adapted from Ping (1999) and are measured on seven-point Likert scales. All items were

measured on seven-point Likert scales, ranging from (1) strongly disagree to (7) strongly

agree.

Control variables.

Responses could be influenced by factors other than the four exchange variables and

individualism, so we controlled for country-, firm-, alliance, and individual-level variables.

The significant differences between the two countries on many of the constructs confirmed

that sampling from multiple countries increased variance on the constructs and also confirmed

the need to control for country in all of our analyses (Gibson, 1999). To ensure that the

cultural values explained unique variance above and beyond country, we included country as

a dummy variable in all analyses (per Kirkman & Shapiro, 2001).

At the firm level, we control for firm size (natural logarithm of the number of

employees), because larger firms with more resources may respond differently to

dissatisfaction (Lambe, Spekman, & Hunt, 2002). In addition, we created three dummy

variables to capture the firm’s industry: production/manufacturing, business service, and

construction. However, the firm size and industry dummies were not significant, so we

removed them from further analyses for parsimony.

We also controlled for alliance duration (natural logarithm of years in operation) as

over time partners’ identification with the alliance might take precedence over their

identification with their parent firms, opportunistic propensity is likely to diminish (Johnson

et al., 1996; Liu et al., 2010). However, as the effect was not significant we removed it from

further analysis. We also control for an alliance manager’s intention to exit the alliance, as an

exit propensity may affect a manager’s intention to act opportunistically.

At the individual level, personal characteristics might influence preferences for

opportunistic behavior (Pansiri, 2005). We captured a manager’s risk propensity with three

22.

seven-point Likert scale question questions (Cronbach’s alpha = .72). Managers who engage

in risk-taking behavior are more likely to act opportunistically. In addition, older managers

might be more experienced and respond differently than younger managers (Tjemkes &

Furrer, 2010; thus we controlled for personal experience with alliances (seven-point Likert

scale). In addition, we controlled for managers’ social desirability tendency by including the

M-C2 version of the Marlowe-Crowne social desirability scale (Strahan & Gerbasi, 1972);

opportunism in particular may be influenced by social desirability bias (Hawkins et al., 2009).

Controlling for social desirability also helps us reduce common method variance (Podsakoff,

MacKenzie, Lee, & Podsakoff, 2003).

RESULTS

Psychometric characteristic and cross-cultural invariance

We employed maximum likelihood (ML) estimation procedures, because the data did not

strongly violate multivariate normality assumptions (McDonald & Ho, 2002). Following

common practice (Hu & Bentler, 1999), we used multiple indicators to assess model fit,

namely, normed chi-square (χ2/d.f.), root mean square error of approximation (RMSEA),

standardized root mean square residual (SRMR), non-normed fit index (NNFI), and

comparative fit index (CFI). We first subjected 20 items pertaining to the independent

variables to an EFA in each country and computed the Cronbach’s alpha for each variable.

Consistent with our expectations, five factors emerged with acceptable construct reliability;

except one item for alternatives availability was removed from further analysis. We then

subjected items with factor loadings greater than .50 in each culture and no cross-loadings to

separate CFAs, as well as a pooled sample. The error variances were all positive and did not

significantly differ from 0; no correlations were greater than 1, and standard errors were not

too large (Cheung & Rensvold, 2002).

The country models possessed good fit; the normed chi-square values were 1.20 and

23.

1.78 for the Netherlands and Turkey, respectively. In addition, other goodness-of-fit indices

suggested acceptable fit: the RMSEA values are .029 [90% confidence interval (CI): .002,

.044] for the Netherlands and .070 [.049, .090] for Turkey, the latter score slightly below the

cut-off value. For the Netherlands, the other indices also suggested a good fit with the

statistics, including .067 (SRMR), .96 (NNFI), and .97 (CFI) and for Turkey, they were .059

(SRMR), .90 (NNFI), and .90 (CFI). The Turkish CFI thus was slightly below the expected

value; attributable to a relative smaller sample size. The model with the pooled sample (n =

402) also produced good fit indices, with a normed chi-square value of 1.84 and fit index

values of .046 (RMSEA) [.037, .054], .051 (SRMR), .94 (NNFI), and .93 (CFI).

To assess convergent validity, we examined the factor loadings, which exceeded the .50

threshold, ranging from .50 to .75 in the Dutch sample and .50 to .84 in the Turkish sample.

The Cronbach’s alphas and composite reliability values were greater than .70, with a few

exceptions that still remained above .60. We conducted the Fornell and Larcker’s (1981) test

to assess discriminant validity. The results indicate that in both countries discriminant validity

was satisfactory.

To evaluate measurement and construct invariance, we used multigroup structural

equation models (AMOS 16.0), performed mean and covariance structure (MACS) analyses,

and considered group comparisons across the two countries. The MACS analysis involved

two nested models that corresponded to the different levels of invariance across groups (e.g.,

Cheung & Rensvold, 2002). In addition to the overall fit indices, we used two comparative fit

indices to evaluate the difference between nested models. First, we used the chi-square

difference test (Δχ2). Second, as recommended by Cheung and Rensvold (2002), we examined

changes in CFI (ΔCFI), which is less affected by sample size. An absolute value of ΔCFI less

than or equal to |.01| would indicate that the invariance hypothesis cannot be rejected.

Regarding configural invariance, all five scales were invariant and unidimensional across

24.

samples. The fit indices of an unconstrained Model were good, with only the CFI slightly

below .95. Regarding metric invariance, the fit indexes of a constrained Model were below

the fit indexes of Model 1 (Δχ2[14] = 43.5, p < .000, ΔCFI = .018). Therefore, we estimated a

second Model, in which we released two factor loadings. The fit indexes of this Model were

as good as those of the unconstrained model (Δχ2[11] = 13.8, p = .28, ΔCFI = .001), in support

of partial metric invariance. Each item loaded on its relevant measure at approximately equal

strength across the two countries. We repeated the same procedure for the dependent variable,

opportunistic propensity, and the results also suggests measurement equivalence (i.e. metric

invariance). A comparison of a unconstrained Model comprising of four valid items with a

constrained Model indicates fit indexes as good as the unconstrained model (Δχ2[2] = 1.69, p =

.43, ΔCFI = .001). We also computed a partial scalar invariance model (both independent and

dependent measurement models), which however revealed significant decrease in model fit,

which might be attributed to cross-cultural differences in scale response styles. Therefore, we

used within-subject standardized scores in the analysis (Hanges, 2004).

Although we reduced some concerns about common method bias by designing a

questionnaire with different scale endpoints and creating psychological separation between

the independent and dependent variables (Podsakoff et al., 2003), we conducted Harman’s

one-factor test. Specifically, we loaded all items of the independent variables in an

exploratory factor analysis and examined the unrotated factor solution to determine the

number of factors needed to account for variance (Podsakoff et al., 2003). The results

indicated limited concerns for common method bias, because the six factors explained,

respectively, 14.8%, 12.9%, 9.2%, 8.6%, 6.9%, and 5.1%.

Hypothesis testing

To test our hypotheses, we conducted hierarchical regression analyses. After averaging the

items related to each variable to compute a construct score we centered the independent

25.

variables to minimize distortion due to possible multicollinearity between the independent

variables and the interaction terms (Aiken & West, 1991). Furthermore, in order to avoid

multi-collinearity between the four exchange variables we conducted four separate sets of

regressions. For each regression, we assessed the possibility of multicollinearity by examining

the variance inflation factors, which were all smaller than the cutoff value of 3; thus,

multicollinearity was not a problem for our data (Hair et al., 2010). To find how much

additional variance is explained by the independent variables after accounting for the controls,

we used regression analysis, such that we entered the control variables in step 1, independent

variables in step 2 and the interaction in step 3, tracking the changes in the multiple squared

correlation coefficient (R2) in each step. Table 1 contains the correlations and Table 2 contains

the regression results. To clarify the interactions we plot them in Figure 2.

[Insert Table 1 and 2 about here]

In step 1, we regressed opportunistic propensity on the controls. The variance explained

by the control variables was between 20.7%. Exit propensity (β = .34, p < .01) and risk

propensity (β = .18, p < .01) were positive and significantly related to opportunism

propensity, such that managers with a tendency to dissolve the alliance or an inclination to

engage in risk-prone behavior are more likely to act opportunistically. Social desirability bias

(β = -.18, p < .01) was negative and significantly related to opportunism propensity, such that

managers who tended to respond in socially desirable ways were less likely to indicate that

they engaged in opportunism. Country (β = -.09, p > .10) and personal experience (β = -.01, p

> .10) were not significantly related to opportunistic propensity.

Economic satisfaction. In step 2, we added economic dissatisfaction and individualism

to the model; the model is significant, with a F-value of 12.56 (p < .001), but with no

significant additional explained variance (ΔR2= .01, p > .10) compared to the base-line model.

The direct effects of economic dissatisfaction (β = .03, p > .10) and individualism (β = -.06, p

26.

> .10) were not significant. In step 3, we included the interaction between economic

satisfaction and individualism; the interaction effects explains 1.4% (p < .05) more variance,

and the F-value in the significant model is 11.87 (p < .001). We proposed in Hypothesis 1 that

individualism moderates the relationship between economic dissatisfaction and opportunism

propensity. The results show that this interaction is significant (β = .13, p < .05). The positive

sign means that the positive effect of economic dissatisfaction on opportunistic propensity is

stronger for managers with individualistic values than for managers with collectivistic values,

in support of Hypothesis 1.

Social dissatisfaction. In step 2, we added social dissatisfaction and individualism to the

model; the model is significant with a F-value of 19.37 (p < .001) and with significant

additional explained variance (ΔR2 = .07, p < .01) compared to the base-line model. The direct

effect of social dissatisfaction (β = .24, p > .10) is positive and significant such that managers

experiencing a poor working relationship are more likely to act opportunistic. The direct

effect of individualism is however not significant (β = -.02, p > .10). In step 3, we included

the interaction between social satisfaction and individualism; the interaction effect explains

1.0% (p < .05) more variance, and the F-value in the significant model is 17.86 (p < .001). We

proposed in Hypothesis 2 that individualism moderates the relationship between social

dissatisfaction and opportunism propensity. The results show that this interaction is

significant (β = -.11, p < .05). The negative sign means that the positive effect of social

dissatisfaction on opportunistic propensity is stronger for managers with collectivistic values

than for managers with individualistic values, in support of Hypothesis 2.

Availability of alternatives. In step 2, we added alternative attractiveness and

individualism to the model; the model is significant, with a F-value of 15.03 (p < .001) and

with significant additional explained variance (ΔR2 = .02, p < .01) compared to the base-line

model. The direct effect of alternative attractiveness (β = -.10, p < .10) is negative and

27.

significant at 10% indicating that managers perceiving more alternatives are less likely to act

opportunistic. The direct effect of individualism is also significant and negative (β = -.13, p >

.10). In step 3, we included the interaction between alternative attractiveness and

individualism; the interaction effect explains 1.0% (p < .05) more variance, and the F-value in

the significant model is 13.81 (p < .001). We proposed in Hypothesis 3 that individualism

moderates the relationship between social dissatisfaction and opportunism propensity. The

results show that this interaction (β = .10, p < .05) is significant and positive, which in part

contrasts our expectations. That is, the results support the expectation that the effect of

alternatives attractiveness on opportunistic opportunism is stronger for managers with

collectivistic values than for managers with individualistic values. However, whereas we

expected a positive effect of alternative attractiveness of opportunism, the result indicate a

negative effect. This means that the results provide partial support for Hypothesis 3.

Alliance specific investments. In step 2, we added alliance specific investments and

individualism to the model; the model is significant with a F-value of 14.64 (p < .001) and

with significant (at 10%) additional explained variance compared to the base-line model (ΔR2

= .01, p < .10). The direct effect of alliance specific investments (β = -.07, p > .10) is not

significant, whereas the direct effect of individualism is significant and negative (β = -.11, p <

.05). In step 3, we included the interaction between alliance specific investments and

individualism; the interaction effect explains 1.0% (p < .05) more variance, and the F-value in

the significant model is 17.86 (p < .001). We proposed in Hypothesis 4 that individualism

moderates the relationship between social dissatisfaction and opportunism propensity. The

results show that this interaction is significant (β = -.10, p < .05). The negative sign means

that the negative effect of alliance specific investments on opportunistic propensity is stronger

for managers with individualistic values than for managers with collectivistic values, in

support of Hypothesis 4.

28.

[Insert Figure 2 about here]

DISCUSSION AND CONCLUSION

The objective of the study was to examine the effect of cultural values on opportunistic

propensity in strategic alliances. More specifically, we examined the interactive effect of

individualism on the relationship between four exchange variables and opportunistic

propensity. The results show that economic and social dissatisfaction increase, whereas

alliance specific investments decrease opportunistic propensity, which is consistent with

previous findings (Deeds & Hill, 1999; Judge & Dooley, 2006; Luo, 2007). Contrary to our

expectations, the availability of attractive alternatives negatively influences opportunistic

propensity. Moreover, we also found that these effects are moderated by individualism, such

that (1) the effects of economic dissatisfaction and alliance specific investments are stronger

for managers with individualist values, whereas (2) the effects of social dissatisfaction and

alternative attractiveness are stronger for managers with collectivistic values.

Theoretical and managerial implications

We advance the literature on international strategic alliances in two ways. First, we contribute

to the debate between universalist and relativist approaches in cross-cultural management

research by demonstrating that if opportunism is a universal construct, managers’

opportunistic propensity is culturally influenced. As opportunism and even the assumption of

opportunism have the potential of negatively affecting the performance of international

strategic alliances (Luo, 2007). Indeed, assuming the opportunism of one’s partner is likely to

become a self-fulfilling prophecy (Ghoshal & Moran, 1996). As demonstrated by Davis,

Schoorman and Donaldson (1997), the assumption of opportunism is likely to trigger the

development of control mechanisms, which implementation is likely to frustrate one’s partner

who is likely to feel betrayed and to start behaving opportunistically in retaliation. John

(1984) also found empirical evidence that bureaucratic control can damage trust and

29.

exacerbate opportunism in interpartner relationships. However, the introduction of different

cultural values in international strategic alliances exacerbates the opportunism issue

considerably. In general, cross-cultural interaction, often replete with misunderstandings and

miscommunication, can foster opportunistic propensity. The fundamental reason is that

managers with different cultural values are likely to have different frames of reference, and it

is the differences in frames of reference that may give rise to opportunism. Cultural values

can differ quite drastically regarding expected patterns of interactions. Thus, an important

managerial implication is that managers’ awareness of cultural differences in opportunistic

propensity becomes critical to establish fair control mechanisms and instill trust and

benevolence, which are important for the success of international strategic alliances (Das &

Teng, 2001).

Limitations and conclusion

The study also has some limitations. First, we measure behavioral intentions rather than actual

behaviors. Although intentions are not always flawless predictors of behavior, our approach

attempts to assess opportunistic propensity, an objective achieved more readily by measuring

behavioral intentions. However, a field study recording alliance managers’ actual behavior

would complement and corroborate our findings. Second, as Johnson and colleagues (1996)

we ask alliance managers to rate their own opportunistic propensity rather than assess the

opportunistic propensity of the partner. Even if we control for social desirability bias it would

be interesting to also investigate the perceptions of counterpart’s alliance managers as their

judgment may be more objective. In addition, assessing one’s partner’s opportunistic

propensity might be interesting as it is likely to influence managers’ own behavior. Third, it

would also be valuable to investigate if managers’ opportunistic propensity differs across

different phases of the alliance relationship. The required longitudinal study would also

enable future research to investigate the consequences of opportunistic propensity to better

30.

understand under what conditions it is beneficial or detrimental to the alliance.

Addressing questions of the cross-cultural generalizability of opportunism is

fundamental to combat the implicit universalism that pervades much organizational and

strategic research (Boyacigiller & Adler, 1991). This study advances international alliance

research by providing a better understanding of how managers may act opportunistically in

alliance relationships in different countries. It demonstrates that opportunism possesses a

similar meaning across two countries, but in addition that opportunistic propensity is affected

interactively by cultural values and exchange variables.

31.

REFERENCES

Aiken, L. S., & West, S. G. 1991. Multiple regression: Testing and interpreting interactions.

Newbury Park, CA: Sage.

Anderson, J. C., & Narus, J. A. 1990. A model of distributor firm and manufacturer firm

working partnerships. Journal of Marketing, 54(1): 42–58.

Ariño, A., De la Torre, J., & Ring, P. S. 2001. Relational quality: Managing trust in corporate

alliances. California Management Review, 44(1): 109–131.

Armstrong, J. S., & Overton, T. S. 1977. Estimating nonresponse bias in mail surveys.

Journal of Marketing Research, 14(3): 396–402.

Au, K. Y. 2000. Intra-cultural variation as another construct of international management: A

study based on secondary data of 42 countries. Journal of International Management, 6:

217–238.

Au, K. Y., & Cheung, M. W. L. 2004. Intra-cultural variation and job autonomy in 42

countries. Organization Studies, 25(8): 1339–1362.

Axelrod, R. 1986. An evolutionary approach to norms. American Political Science Review,

80(4): 1095–1111.

Beamish, P. W., & Banks, C. W. 1987. Equity joint ventures and the theory of the

multinational enterprise. Journal of International Business Studies, 18(2): 1–16.

Blau, P. M. 1964. Exchange and power in social life. Wiley: New York.

Boyacigiller, N. A., & Adler, N. J. 1991. The parochial dinosaur: Organizational science in a

global context. Academy of Management Review, 16(2): 262–290.

Brislin, R. W. 1986. The wording and translation of research instruments. In Field methods in

cross-cultural research, Lonner, W. J., & Berry, J. W. (eds). Sage: Beverly Hills, CA:

137–164.

Brouthers, K. D., & Bamossy G. J. 2006. Post-formation processes in Eastern and Western

European joint ventures. Journal of Management Studies, 43: 203–229.

Buchanan, L. 1992. Vertical trade relationships: The role of dependence and symmetry in

attaining organizational goals. Journal of Marketing Research, 29: 65–75.

Chen, C. C., Peng, M. W., & Saparito, P. A. 2002. Individualism, collectivism, and

opportunism: A cultural perspective on transaction cost economics. Journal of

Management, 28(4): 567–583.

Cheung, G. W., & Rensvold, R. B. 2002. Evaluating goodness-of-fit indexes for testing

measurement invariance. Structural Equation Modeling, 9: 233–255.

32.

Das, T. K., & Rahman, N. 2010. Determinant of partner opportunism in strategic alliances: A

conceptual framework. Journal of Business Psychology, 25: 55–74.

Das, T. K., & Teng, B. S. 2001. Trust, control, and risk in strategic alliances: An integrated

framework. Organization Studies, 22(2): 251–283.

Davis, J. H., Schoorman, F. D., & Donaldson, L. 1997. Toward a stewardship theory of

management. Academy of Management Review, 22(1): 20–47.

Deeds, D. L., & Hill, C. W. L. 1999. An examination of opportunistic action within research

alliances: Evidence from the biotechnology industry. Journal of Business Venturing,

14(2): 141–163.

Dutton, J. E., & Jackson, E. E. 1987. Categorizing strategic issues: Links to organizational

action. Academy of Management Review, 12(1): 76–90.

Dwyer, F. R., Schurr, P. H., & Oh, S. 1987. Developing buyer-seller relationships. Journal of

Marketing, 51(2): 11–27.

Fornell, C., & Larcker, D. F. 1981. Evaluating structural equation models with unobservable

variables and measurement error. Journal of Marketing Research, 18(1): 39–50.

Furrer, O., Tjemkes, B. V., Ülgen Aydinlik, A., & Adolfs, K. in press. Responding to adverse

situations within exchange relationships: The cross-cultural validity of a circumplex

model. Journal of Cross-Cultural Psychology, forthcoming.

Geyskens, I., & Steenkamp, J-B. E. M. 2000. Economic and social satisfaction: Measurement

and relevance to marketing channel relationships. Journal of Retailing, 76(1): 11–32.

Ghoshal, S., & Moran, P. 1996. Bad for practice: A critique of the transaction cost theory.

Academy of Management Review, 21(1): 13–47.

Gibson, C. B. 1999. Do they do what the believe the can? Group effectiveness across tasks

and cultures. Academy of Management Journal, 42(2): 138–152.

Grimm, S. D., Church, A. T., Katigbak, M. S., & Reyes, J. A. S. 1999. Self-described traits,

values, and moods associated with individualism and collectivism: Testing I-C theory in an

individualistic (US) and a collectivistic (Philippine) culture. Journal of Cross-Cultural

Psychology, 30(4): 466–500.

Gulati, R., Khanna, T., & Nohria, N. 1994. Unilateral commitments and the importance of

process in alliances. Sloan Management Review, 35(3): 61–69.

Hair, J. F., Black, W. C., Babin, B. J., & Anderson, R. E. 2010. Multivariate data analysis: A

global perspective. New York: Macmillan.

Hanges P. L. 2004. Response bias correction procedure used in GLOBE. In Culture,

Leadership, and Organizations: The GLOBE Study of 62 Societies, House R.J., Hanges

33.

P.J., Javidan M., Dorfman P.W., Gupta V. (eds). Sage: Thousand Oaks, CA; 737-751.

Hawkins, T. G., Knipper, M. G., & Strutten, D. 2009. Opportunism in buyer-supplier

relations: New insights from quantitative synthesis. Journal of Marketing Channels,

16(1): 43–75.

Hofstede, G. 2001. Culture’s consequences: Comparing values, behaviors, institutions, and

organizations across nations. Second edition. Thousand Oaks: London.

Hu, L. T., & Bentler, P. M. 1999. Cutoff criteria for fit indexes in covariance structure

analysis: conventional criteria versus new alternatives. Structural Equation Modeling, 6:

1–55.

Hui, C. H., & Triandis, H. C. 1985. Measurement in cross-cultural psychology: A review and

comparison of strategies. Journal of Cross-Cultural Psychology, 16: 131–152.

John, G. 1984. An empirical-investigation of some antecedents of opportunism in a marketing

channel. Journal of Marketing Research, 21(3): 278–289.

Johnson, J. L., Cullen, J. B., & Sakano, T. 1996. Opportunistic tendencies in IJVs with the

Japanese: The effects of culture, shared decision making, and relationship age. The

International Executive, 38(1): 79–94.

Johnson, J. L., Cullen, J. B., Sakano, T., & Takenouchi, H. 1996. Setting the stage for trust

and strategic integration in Japanese-US cooperative alliances. Journal of International

Business Studies, 27: 981–1004.

Judge, W. Q., & Dooley, R. 2006. Strategic alliance outcomes: A transaction-cost economics

perspective. British Journal of Management, 17(1): 23–37.

Kirkman, B. L., & Shapiro, D. L. 2001. The impact of team members’ cultural values on

productivity, cooperation, and empowerment in self-managing work teams. Journal of

Cross-Cultural Psychology, 32(5): 597–617.

Kluckhohn, F., & Strodtbeck, F. L. 1961. Variations in value orientations. Evanston, IL:

Row.

Kogut, B. 1988. Joint ventures: Theoretical and empirical perspectives. Strategic

Management Journal, 9: 319–332.

Kumar, N., Stern, L. W., & Anderson, J. C. 1993. Conducting interorganizational research

using key informants. Academy of Management Journal, 36(6): 1633–1651.

Kumar, R., & Nti, K. O. 2004. National cultural values and the evolution of process and

outcome discrepancies in international strategic alliances. Journal of Applied Behavioral

Science, 40(3): 344–361.

Lambe, C. J., Spekman, R. E., & Hunt, S. D. 2002. Alliance competence, resources, and

34.

alliance success: Conceptualization, measurement, and initial test. Journal of the Academy

of Marketing Science, 30(2): 141–158.

Liu, Y., Su, C. Li, Y., & Liu, T. 2010. Managing opportunism in a developing interfirm

relationship: The interrelationship of calculative and loyalty commitment. Industrial

Marketing Management, 39(5): 844–852.

Luo, Y. 2007. Are joint venture partners more opportunistic in a more volatile environment?

Strategic Management Journal, 28: 39–60.

Luo, Y., & Shenkar, O. 2002. An empirical inquiry of negotiation effects in cross-cultural

joint ventures. Journal of International Management, 8: 141–162.

Maitland, I., Bryson, J., & Van de Ven, A. 1985. Sociologists, Economists, and Opportunism.

Academy of Management Review, 10(1): 59–65.

Markus, H.R., & Kitayama, S. 1991. Culture and the self: Implications for cognition, emotion,

and motivation. Psychological Review, 98(2): 224–253.

McDonald, R. P., & Ho, M.-H. R. 2002. Principles and practice in reporting structural

equation analyses. Psychology Methods, 7: 64–82.

Pansiri, J. 2005. The influence of managers’ characteristics and perceptions in strategic

alliance practice. Management Decision, 43(9): 1097–1113.

Park, S. H., & Ungson, G. R. 2001. Interfirm rivalry and managerial complexity: A

conceptual framework of alliance failure. Organization Science, 12: 37–53.

Patterson, P. G., Cowley, E., & Prasongsukarn, K. 2006. Service failure recovery: The

moderating impact of individual-level cultural value orientation on perceptions of justice.

International Journal of Research in Marketing, 23(3): 263–277.

Ping, R. A. 1993. The effects of satisfaction and structural constraints on retailer exiting,

voice, loyalty, opportunism, and neglect. Journal of Retailing, 69(3): 320–352.

Podsakoff, P. M., MacKenzie, S. B., Lee, J. Y., & Podsakoff, N. P. 2003. Common method

biases in behavioral research: A critical review of the literature and recommended

remedies. Journal of Applied Psychology, 88(5): 879–903.

Ralston, D. A., Egri, C. P., de la Garza Carranza, M. T., Ramburuth, P. et al. 2009. Ethical

preferences for influencing superiors: A 41-society study. Journal of International

Business Studies, 40: 1022–1045.

Sakalaki, M., Kazi, S., & Karamanoli, V. 2007. Do individualists have a higher opportunistic

propensity than collectivists? Individualism and economic cooperation, Revue

Internationale de Psychologie Sociale, 20: 59–76.

Sallivan, J., & Nonaka, I. 1988. Culture and strategic issue categorization theory.

35.

Management International Review, 28(3): 6–10.

Schneider, S. C., & Meyer, A. de. 1991. Interpreting and responding to strategic issues: The

impact of national culture. Strategic Management Journal, 12(4): 307–320.

Smith, P. B., & Bond, M. H. 1999. Social psychology across cultures. Boston: Allyn &

Bacon.

Smith, P. B., Dugan, S., & Trompenaars, F. (1996). National culture and the values in

organizational employees: A dimensional analysis across 43 nations. Journal of Cross

Cultural Psychology, 27: 231–264.

Steenkamp, J-B. E. M., & Baumgartner, H. 1998. Assessing measurement invariance in cross-

national consumer research. Journal of Consumer Research, 25: 78–90

Strahan, R., & Gerbasi, K. C. 1972. Short, homogeneous versions of Marlow-Crowne social

desirability scale. Journal of Clinical Psychology, 28(2): 191–193.

Thomas, D. C., & Au, K. 2002. The effect of cultural differences on behavioral responses to

low job satisfaction. Journal of International Business Studies, 33(2): 309–326.

Thomas, D. C., Au, K., & Ravlin, E. E. 2003. Cultural variation and the psychological

contract. Journal of Organizational Behavior, 24(5): 451–471.

Tjemkes, B., & Furrer, O. 2010. The antecedents of response strategies in strategic alliances.

Management Decision, 48: 1103–1133.

Tjemkes, B. V., Furrer, O., Adolfs, K., & Ülgen Aydinlik, A. in press. Response strategies in

an international strategic alliance experimental context: Cross-country differences. Journal

of International Management, forthcoming.

Triandis, H. C. 1995. Culture and social behavior. New York: McGraw-Hill.

Triandis, H. C., Leung, K., Villareal, M., & Clack, F. L. 1985. Allocentric versus idiocentric

tendencies: Convergent and discriminant validation. Journal of Research in Personality,

19(4): 395–415.

Wathne, K. H., & Heide, J. B. 2000. Opportunism in interfirm relationships: Forms,

outcomes, and solutions. Journal of Marketing, 64(4): 36–51.

Williamson, O. E. 1985. The economic institutions of capitalism. New York: Free Press.

Wong, A., Tjosvold, D., & Yu, Z.-Y. 2005. Organizational partnerships in China: Self-

interest, goal interdependence, and opportunism. Journal of Applied Psychology, 90(4):

782–791.

Yoo, B., Donthu, N., & Lenartowicz, T. 2011. Measuring Hofstede’s five dimensions of

cultural values at the individual level: Development and validation of CVSCALE. Journal

of International Consumer Marketing, 23(3/4): 193–210.

36.

Table 1: Descriptive Statistics and Correlations

Variables Mean S.D. 1. 2. 3. 4. 5. 6. 7. 8. 9. 10. 11.

1. Opportunistic propensity 2.39 1.10

2. Economic dissatisfaction 4.01 1.58 .05

3. Social dissatisfaction 2.24 1.20 .42*** .20***

4. Alliance spec. investments 4.42 1.25 -.03 -.17*** .07

5. Alternative attractiveness 5.28 1.17 -.04 .03 -.15** -.08

6. Individualism/collectivism 2.69 1 .05 -.10* .08 .07 -.19***

7. Nationality n.a. .49 .00 .60*** .09 -.20*** .14** -.16**

8. Exit propensity 2.42 1.21 .38*** .09 .41*** -.04 .01 .05 .11*

9. Personal experience 3.86 1.97 .02 .52*** .07 -.18** .07 -.19*** .71*** .15**

10. Risk propensity 4.17 1.17 .23*** .10 .18** .22*** .03 .05 .14* .12** .08

11. Social desirability bias 7.34 1.60 -.22*** -.15** -.12* .13* -.07 -.03 -.28*** -.16** -.21*** -.12*

n = 402 (pooled sample); *** p < .001 ** p < .01 * p < .05

16430

37.

Table 2: Hierarchical Regression Results

Variables 1a 1b 1c 2a 2b 2c 3a 3b 3c 4a 4b 4c

Nationality

Exit propensity

Experience

Risk propensity

Social desirability bias

-.12

.37***

-.00

.20***

-.17**

-.14

.36***

-.01

.19***

-.17**

-.11

.34***

-.01

.18***

-.16**

-.12

.37***

-.00

.20***

-.17**

-.10

.26***

.02

.18***

-.15**

-.09

.24***

.02

.17***

-.15**

-.12

.37***

-.00

.20***

-.17**

-.17**

.36***

-.02

.20***

-.16**

-.17*

.36***

-.01

.19***

-.16**

-.12

.37***

-.00

.20***

-.17**

-.15*

.35***

-.03

.18***

-.17**

-.15*

.33***

-.01

.18***

-.17**

Individualism (Ind.)

Economic dissatisfaction

Social dissatisfaction

Alliance specific. investments

Alternative attractiveness

-.12*

-.05

-.11

.06

-.02

.28***

-.05

.24***

-.11*

-.07

-.10*

-.04

-.13*

-.10

-.13*

-.12*

Ind. Eco. dissat.

Ind. Soc. dissat.

Ind. All. spec. invest.

Ind. Alternative attract.

.11*

-.11*

-.10*

.10*

F-value

R²

∆R²

19,17***

.23

14,45***

.24

.01

13.40***

.25

.01*

19,17***

.23

19.73***

.29

.06***

17.68***

.30

.01*

19.17***

.23

14.65***

.24

.01

13.42***

.25

.01*

19.17***

.23

15.03***

.24

.01*

13.81***

.25

.01*

n = 402 (pooled sample) *** p < .001 ** p < .01 * p < .05

16430

38.

Figure 1. Conceptual Model

16430

39.

Figure 2. Interaction Effects

----------------- = Individualism —————— = Collectivism