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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
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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
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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
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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
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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
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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
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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
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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
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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
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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,
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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.
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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
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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
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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.
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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.
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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
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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