The Effect of Unemployment on Life Satisfaction: A Cross-National Comparison Between Canada, Germany, the United Kingdom and the United States

The Effect of Unemployment on Life Satisfaction: A Cross-National Comparison Between Canada, Germany, the United Kingdom and the United States. Wen-Hao Chen, Feng Hou. Applied Research in Quality of Life, September 2019, Volume 14, Issue 4, pp 1035–1058. https://link.springer.com/article/10.1007/s11482-018-9638-8

Abstract: This paper investigates the effect of unemployment on life satisfaction from a comparative perspective. It also tests whether the link between unemployment and life satisfaction is moderated or reinforced by contextual unemployment across regions within a country—either through a negative spillover or a positive social-norm effect, or both. The results suggest that noticeable non-pecuniary costs are associated with unemployment in the four countries studied. Cross-national differences also emerged in the impact of the moderating factors. Regional unemployment is a strong moderating factor of own unemployment in Canada and to a lesser extent in the United States; the effect is ambiguous in the United Kingdom and exacerbating in Germany. The results also support a negative spillover effect of regional unemployment on the employed in the United States and Germany, no spillover effect in the United Kingdom and, surprisingly, a positive overall spillover effect in Canada. Sensitivity testing further revealed that this Canadian anomaly was a phenomenon mainly in Atlantic Canada, not across the whole country.

Keywords: Unemployment Subjective well-being Employment insecurity Social norm


Cross-National Differences and Possible Explanations
The results reveal some interesting cross-country similarities and differences in the well-being of the employed and the unemployed among the four countries studied. In all cases, noticeable non-pecuniary costs are associated with unemployment, net of observable characteristics. Unemployed Canadians, like the unemployed in the other countries, tend to have lower levels of life satisfaction than the employed, witheverything else held equal. The estimated employed–unemployed life satisfaction gap looks remarkably similar in Canada, the United States and the United Kingdom. It is significantly wider in Germany.

This overall pattern is consistent withour prior expectation that the negativeeffect of own unemployment on life satisfaction is exacerbated in countries where employment protection legislation (EPL) is strict. Empirical studies often find that strict EPL negatively affects unemployment inflow and outflow,because stricter EPL entails higher firing costs and reduces the propensity tohire by employers (for a review, see OECD2013). The differences in EPL among the four countries (over their respective sample periods) reported inTable5 support this claim. Germany had relatively more stringent EPL againstdismissals for regular and temporary workers than, in order, the United Kingdom, Canada and the United States.

Job Security and Prospects
The other possible explanation for the wide employed–unemployed life satisfaction gapin Germany is labour market security and prospects. Knabe and Ratzel (2011) showed that individuals’ perceptions of labour market risk have a greater impact on their well-being than their employment status. Clark et al. (2010) divided sample in the labourforce into four mutually exclusive groups according to their job security and prospects. They found that the employed with high job security had the highest life satisfaction, while the unemployed with poor job prospects had the lowest life satisfaction. Interestingly, the average life satisfaction of the insecurely employed and of the unemployed with good job prospects is similar. The wide employed–unemployed life satisfactiongap in Germany may indicate a polarized labour market, where the employed are secure and the unemployed have poor job prospects (as a result of, for instance, significantentry or re-entry barriers).

The summary labour market and institution indicators for the four countries reported in Table5 were used to determine whether the labour market in Germanyis polarized.23The results support the idea of a polarization between job security and job prospects in the German labour market. Germany stands out as having the lowest incidence of short job tenures, as well as the highest incidence of long job tenures, among the four countries. Over the study period, nearly 60% of German employees had five years or more of tenure with their current employer. Only 46%to 48% of employees in the three other countries had five years of more of tenure with their current employers over their corresponding periods. The employee turnover rate was lower in Germany: only about 15% of employees experienced a short job tenure of less than a year, versus 20% or more of employees in thethree other countries. These results suggest that German employees are relatively more secure in their jobs. Table 5 also confirms that prospects for job seekers are relatively gloomy in Germany: over 50% of the unemployed were out of work longer than one year. The long-term unemployment rates are much lower in thethree other countries, at 12%for Canada, 15% for the United States and 27% United Kingdom. This also explains why unemployed Germans had significantly lower predicted life satisfaction scores (see Section 4.3) than their otherwise-equal counterparts in other countries.

Happier Indonesians in 2007 earned more money, were more likely to be married, less likely to be divorced or unemployed, and in better health when the survey was conducted again seven years later

Does Happiness Pays? A Longitudinal Family Life Survey. Sujarwoto Sujarwoto. Applied Research in Quality of Life, November 23 2019. https://link.springer.com/article/10.1007/s11482-019-09798-x

Abstract: Most of the research on happiness has documented that income, marriage, employment and health affect happiness. Very few studies examine whether happiness itself affect income, marriage, employment and health. This study does so, benefiting from data drawn from the panel longitudinal Indonesian Family Life Survey (IFLS) 2007 and 2014. The survey includes 23,776 individuals from 15,067 households living in about 262 neighborhoods between 2007 and 2014. The findings show that happier Indonesians in 2007 earned more money, were more likely to be married, were less likely to be divorced or unemployed, and were in better health when the survey was conducted again seven years later. Policy makers may consider that increasing citizen happiness is vital to achieve citizen success on labor markets, to improve their job performance and to maintain their health.

Keywords: Happiness Positive cognitive bias Income Indonesia

Check also Individual and Contextual Factors of Happiness and Life Satisfaction in a Low Middle Income Country. Sujarwoto Sujarwoto, Gindo Tampubolon, Adi Cilik Pierewan. Applied Research in Quality of Life 13(4), October 2017. DOI: 10.1007/s11482-017-9567-y:

Abstract: Understanding individual and contextual factors of happiness and life satisfaction in a low- and middle-income country setting are important in the study of subjective well-being. This study aims to examine individual and contextual factors of happiness and life satisfaction in one of the happiest countries in the world: Indonesia. Data comes from the Indonesian Family Life Survey 2014 (N individual = 31,403; N household = 15,160; N district = 297). Results from a three-level ordered logit model show that factors of happiness and life satisfaction are beyond individual factors. Happiness and life satisfaction are also strongly associated with factors within an individual’s household and at the district government level. Individuals living in households with better economic welfare are happier and more satisfy. Poor health and unemployment have a detrimental effect on happiness and life satisfaction. Individuals living in districts whose governments’ better deliver public services are happier and more satisfy. In contrast, those living in areas with conflict and violence is less happy and satisfy. Individual religiosity and community social capital in the form of indigenous tradition benefit individual happiness and life satisfaction.

People higher in dark-personality constructs may perceive others’ dark characteristics as less undesirable (“darkness tolerance”); it may result from similarity-liking mechanisms

Desirability of others' dark characteristics: The role of perceivers’ dark personality. William Hart, Kyle Richardson. Personality and Individual Differences, November 22 2019, 109722. https://doi.org/10.1016/j.paid.2019.109722

Abstract: Previous research suggests people higher in dark-personality constructs may perceive others’ dark characteristics as less undesirable (“darkness tolerance”). To advance understanding, we examined darkness tolerance in the context of various dark-personality constructs and addressed different explanations for its occurrence. Participants (N = 567) indicated the desirability of dark characteristics in others under conditions where dark characteristics would be either highly or less undesirable; also, participants self-reported possession of these characteristics and completed measures of dark-personality constructs (narcissism constructs, Machiavellianism, and psychopathy). Dark-personality constructs related to enhanced desirability ratings for dark characteristics in ways that generally matched how the constructs related to enhanced self-ratings on the same characteristics; also, dark characteristics were rated less favorably under highly-undesirable vs. less-undesirable conditions, and this effect was not modified by any dark-personality construct. In sum, darkness tolerance occurs across dark-personality constructs; it may result from similarity-liking mechanisms but not from insensitivity to contextual features that make dark characteristics particularly undesirable.

Keywords: Dark personalityPerson perceptionPersonality


3. Discussion

Prior work had generally focused on darkness tolerance in the context of narcissism, but the present findings suggest that darkness tolerance is a common thread running through dark-personality constructs; darkness tolerance was present across six dark-personality constructs, albeit it appeared smaller for arguably the least “dark” constructs (GN, VN, and NARQ-A). Indeed, Lamkin et al. (2018) showed darkness tolerance as a function of trait antagonism, and trait antagonism might unite dark-personality constructs. This study provided novel insights on the underpinnings of darkness tolerance. Theorists suggested that darkness tolerance could spring from similarity-liking mechanisms and/or an apparent tone-deafness (Adams et al., 2015; Wallace et al., 2015). The present study provided novel tests of these accounts. Inconsistent with the tone-deafness account, perceivers high and low in dark-personality constructs agreed on the extent to which dark (and bright) characteristics were undesirable as a function of the darkness-desirability manipulation. These null interaction findings are unlikely due to limited power or the un-modifiable nature of the darkness-desirability effect on desirability ratings; indeed, the darkness-desirability effect was moderated by scenario. Hence, darkness tolerance might not arise from tone-deafness. The present data supported the similarity-liking account using a novel method for testing this account (for other support, Burton et al., 2017; Hart & Adams, 2014); specifically, dark-personality constructs generally related to desirability ratings of the dark/bright characteristics in ways that matched how they related to self-ratings on these characteristics.5 But, this matching was less impressive for vulnerable narcissism. Perhaps people high in VN showed weak similarity-liking effects because they have low self-esteem (Klohnen & Mendelsohn, 1998); in this study, VN inversely related to self-esteem (r = −0.34). To check on this idea, we first removed self-esteem variance from VN; we predicted VN from self-esteem and saved the residual. Second, we related this residualized version of VN to (a) the 13 desirability ratings and (b) the 13 self-ratings (for these rs, see supplemental Table S4). Finally, we computed the relation between these two “sets” of rs (i.e., set “a” and set “b” rs). This computation revealed weak profile mismatch (r = −0.18). So, it seems that low self-esteem, if anything, contributed similarity-liking effects in VN. Perhaps people high in VN showed weak similarity-liking effects because their self-views are not held with much confidence (Hart, Adams, Burton & Tortoriello, 2017) and are therefore less likely to factor into evaluating others’ characteristics. The present study is not without limitations. First, because we used superficial methods to investigate darkness tolerance and examined the phenomenon in a college sample with a high proportion of females, it remains unclear whether the results are applicable to different settings or populations. Second, because the study was limited to unidimensional indicators of Machiavellianism and psychopathy, it remains unclear whether the present results are applicable to all features of these constructs. Third, we used single items to assess desirability and selfratings of the 13 dark/bright characteristics, and single item indicators are prone to low reliability. This suggests findings pertaining to analyses of the single items might not be highly precise. Finally, although evidence was consistent with a similarity-liking mechanism, this evidence was correlational and does not show causation.

Moral foundations (Care/Injury, Fairness/Deception, Loyalty/Betrayal, Authority/Subversion, Sacred/Fallen) are based on emotions and intuitions, but are substantially associated with cognitive ability

Moral foundations and cognitive ability: Results from a Japanese sample. Tetsuya Kawamoto, Takahiro Mieda, Atsushi Oshio. Personality and Individual Differences, Volume 149, October 15 2019, Pages 31-36. https://doi.org/10.1016/j.paid.2019.05.050

Abstract: Research has indicated that human morality is associated with cognitive ability. However, morality is not a unified concept but rather is a multi-faceted concept. Moral Foundation Theory suggests that human beings have at least five innate moral foundations: Care/Injury, Fairness/Deception, Loyalty/Betrayal, Authority/Subversion, and Sacred/Fallen. The present study aimed to investigate the associations between these moral values and cognitive ability. A large-scale cross-sectional survey was conducted online, which was completed by a total of 4863 Japanese adults (2922 females, 1935 males, and 6 “other;” Mage = 48.78, SDage = 10.93, range 20–70). Correlation and multiple regression analyses revealed that cognitive ability was positively linked to Care/Injury, Fairness/Deception, and Sacred/Fallen. Notably, the positive associations of cognitive ability with Loyalty/Betrayal and Authority/Subversion were statistically significant only for people below the age of approximately 50. These findings indicate that although moral foundations are based on emotions and intuitions, they are substantially associated with cognitive ability. In addition, the positive direction of associations of cognitive ability with Loyalty/Betrayal and Authority/Subversion was inconsistent with previous findings in Western countries, which suggests that these associations are culture-dependent.

Keywords: Cognitive abilitySyllogism-solvingMoral foundationMoral decision-makingAgeJapanese


4. Discussion

The present study examined the associations between moral foundations and cognitive ability and the moderating effect of age on these
associations using a large Japanese adult sample. Cognitive ability was
positively associated with individualizing foundations (Care/Injury and
Fairness/Deception) in this study. This result is basically consistent
with previous findings (Carl, 2014, 2015; Deary et al., 2008; Schoon
et al., 2010; Van Leeuwen et al., 2014). In addition, cognitive ability
was also positively correlated with binding foundations (Loyalty/Betrayal, Authority/Subversion, and Sacred/Fallen). These associations
were inconsistent with previous Western findings (Pennycook et al.,
2014; Sidanius et al., 1996; Stankov, 2009). However, as Kemmelmeier
(2008) noted, the associations between cognitive ability and
conservatism-related values might be moderated by environmental
political interest. Since Japanese people, especially young ones, have
among the lowest political interest and engagement rates (Cabinet
Office, 2014), the positive associations of cognitive ability with binding
foundations were expected. Hence, the present hypotheses on the
moral-cognitive ability associations were supported by this study. As for
age as a moderator, interaction terms were significant for all moral
foundations. A subsequent simple slope test showed that the links to
each moral foundation were stronger for younger people. Moreover, the
Johnson–Neyman test revealed that the moral-cognitive ability links
became significant among people below the age of approximately 50 for
binding foundations other than Sacred/Fallen. Therefore, the present
age-as-moderator hypothesis was partially supported in this study.
Five moral foundations are considered as innate psychological
systems that give rise to moral intuitions throughout the world (Graham
et al., 2009, 2011, 2013). According to the MFT, the five intuitive
systems are respectively linked to distinct moral values: Care/Injury,
Fairness/Deception, Loyalty/Betrayal, Authority/Subversion, and
Sacred/Fallen (Graham et al., 2009, 2011, 2013). The MFT focuses on
the role of emotional intutions in determining moral judgment and
insists that moral judgements derive almost entirely from intuitive
processes (Graham et al., 2013). That is, although they do not deny the
role of cognitive reasoning processes (Graham et al., 2011), deliberate
cognitive reasoning is not overly assumed in MFT. The present results
showed that higher cognitive ability is linked to higher levels of every
moral foundation, especially Care/Injury, Fairness/Deception, and
Sacred/Fallen. While the present results basically conformed to previous findings (Carl, 2014, 2015; Deary et al., 2008; Schoon et al.,
2010; Van Leeuwen et al., 2014), the underlying reason why these
foundations are positively related to cognitive ability remains unknown. Moreover, the present study showed that age exerted a moderating effect on the associations between cognitive ability and moral
foundations. In particular, with respect to Loyalty/Betrayal and Authority/Subversion, the associations with cognitive ability were observed only in younger people. Considering that Japanese young people
show low political interest and engagement (Cabinet Office, 2014), this
age-by-cognitive ability interaction is consistent with previous research
findings (Kemmelmeier, 2008; Kemmelmeier et al., 1999; Krosnick,
1988, 1990).
The associations between cognitive ability and moral foundations
observed in this study can be interpreted from the viewpoint of social
learning or enculturation. According to MFT, moral foundations are
basically innate (Graham et al., 2011, 2013). However, they also note
that these foundations are malleable, and there is room for cultural
learning (Graham et al., 2013). Cognitive ability is a fundamental tool
for human learning, and is also a useful tool for social learning
(Christie, 2017). Although there are some differences in the moral values regarded as important among various cultures or societies, care
and justice are seen as basic morality in at least developed countries.
Children socially learn their importance as moral principles and become
socialized (Smetana, 1999). Cognitive ability plays an important role in
this social learning process, which promotes positive associations with
Care/Injury and Fairness/Deception, that is, individualizing foundations.
Sacred/Fallen is considered as one of the binding foundations
(Graham et al., 2009). However, the present results on Sacred/Fallen
were not consistent with Loyalty/Betrayal and Authority/Subversion,
but rather similar with Care/Injury and Fairness/Deception. Considering the raw inter-scale correlations among moral foundations,
Sacred/Fallen can be construed as more closely associated with individualizing foundations in Japanese people. Previous research also
showed that openness to experience and honesty-humility are positively
associated with Care/Injury, Fairness/Deception and Sacred/Fallen in
Japanese (Kawamoto et al., 2017). Based on these findings, the location
of Sacred/Fallen in moral foundation factor structure for Japanese
people may be different from that for Westerners.
The present results pointing to age as a moderator of the associations between cognitive ability and moral foundations are novel. In
particular, the associations with Loyalty/Betrayal and Authority/
Subversion were observed only for people below the age of 50. Loyalty/
Betrayal and Authority/Subversion are linked to social conservatism
(Graham et al., 2009, 2011). People with little political interest tend to
have less opportunities to think about and discuss political issues; this
tendency is conducive to maintaining high consistency between values
and beliefs (Kemmelmeier et al., 1999; Krosnick, 1988, 1990). Besides,
Japanese young people have little political interest and engagement
(Cabinet Office, 2014). Considering these previous findings, Japanese
young people may passively accept conservative values that are widely
endorsed in the Japanese society without questioning them, since their
surroundings are characterized by low political interest. This holds
especially true for young individuals with relatively high cognitive
ability, since the processes involved in learning social values or norms
necessitate high cognitive ability (Buckholtz & Marois, 2012; Christie,
2017). Therefore, it is plausible to argue that younger and smarter
people are more likely to passively accept conservative values, which
are widely endorsed in Japan, without analyzing them.
The present study had some strengths including a sufficient sample
size and the use of a Japanese sample. As research on moral foundations
has been mainly conducted in Western countries, the present results
extended previous findings. Despite these strengths, this study also had
some limitations. First, we relied on self-reported moral foundation
scores; mono-rater measures of psychological traits are vulnerable to
random and systematic errors (Campbell & Fiske, 1959). Future research ought to use other self-rating measures or other methods to assess people's morality. Second, the present sample was composed only
of Japanese individuals. Accordingly, future research should recruit
culturally and ethnically diverse samples. Third, cognitive ability was
measured with the BAROCO Short, which has been well validated but is
a relatively simple test. In order to further examine the associations
between moral foundations and detailed facets of cognitive ability,
future research should use tests that can capture the multiple facets of
cognitive ability.
This study offered the positive associations between binding foundations and cognitive ability. Notably, the associations were significantly moderated by the participants' age. These findings were
novel, and additionally, suggested that the associations between moral
values and cognitive ability could be modified by external factors. This
study focused only on the participants' age as a moderator, however,
future research should explore possible interactions between cognitive
ability and other factors. In addition, the present findings implied that
lower political interest predispose people to accept moral values that
widely prevailed in their society without thinking. We can say that it is
important to encourage younger people to get an interest in their society and politics in education.

Spearman’s g Found in 31 Non-Western Nations, 52000 people: Strong Evidence That g Is a Universal Phenomenon

Spearman’s g Found in 31 Non-Western Nations: Strong Evidence That g Is a Universal Phenomenon. Russell T. Warne. Psychological Bulletin, 145(3), 237-272. Nov 2010. http://dx.doi.org/10.1037/bul0000184

Abstract: Spearman’s g is the name for the shared variance across a set of intercorrelating cognitive tasks. For some—but not all—theorists, g is defined as general intelligence. While g is robustly observed in Western populations, it is questionable whether g is manifested in cognitive data from other cultural groups. To test whether g is a cross-cultural phenomenon, we searched for correlation matrices or data files containing cognitive variables collected from individuals in non-Western, nonindustrialized nations. We subjected these data to exploratory factor analysis (EFA) using promax rotation and 2 modern methods of selecting the number of factors. Samples that produced more than 1 factor were then subjected to a second-order EFA using the same procedures and a Schmid-Leiman solution. Across 97 samples from 31 countries totaling 52,340 individuals, we found that a single factor emerged unambiguously from 71 samples (73.2%) and that 23 of the remaining 26 samples (88.5%) produced a single second-order factor. The first factor in the initial EFA explained an average of 45.9% of observed variable variance (SD = 12.9%), which is similar to what is seen in Western samples. One sample that produced multiple second-order factors only did so with 1 method of selecting the number of factors in the initial EFA; the alternate method of selecting the number of factors produced a single higher-order factor. Factor extraction in a higher-order EFA was not possible in 2 samples. These results show that g appears in many cultures and is likely a universal phenomenon in humans.

Public Significance Statement: This study shows that one conceptualization of intelligence—called Spearman’s g—is present in over 90 samples from 31 non-Western, nonindustrialized nations. This means that intelligence is likely a universal trait in humans. Therefore, it is theoretically possible to conduct cross-cultural research on intelligence, though culturally appropriate tests are necessary for any such research.

KEYWORDS: Spearman’s g, cross-cultural psychology, general cognitive ability, human intelligence, factor analysis

General Discussion of Results

We conducted this study to create a strong test of the theory that general cognitive ability is a cross-cultural trait by searching for g in human populations where g would be the least likely to be present or would be weakest. The results of this study are remarkably similar to results of EFA studies of Western samples, which show that g accounts for approximately half of the variance among a set of cognitive variables (e.g., Canivez & Watkins, 2010). In our study, the first extracted factor in 97 EFAs of data sets from non-Western, nonindustrialized countries was 45.9%.

Moreover, 73.2% of the data sets unambiguously produced a single factor, regardless of the method used to select the number of factors in the EFA. Of the remaining data sets, almost every one in which a second-order EFA was possible produced a single general factor. The only exceptions were from Grigorenko, Ngorosho, Jukes, and Bundy (2006) and Gurven et al. (2017). The Grigorenko et al. (2006) dataset produced two general factors only if one sees the modified Guttman method of selecting the number of first-order factors as being a more realistic solution than MAP. Given the modified Guttman rule’s penchant for overfactoring and the generally accurate results from MAP in simulation studies (Warne & Larsen, 2014), it seems more likely that even the Grigorenko et al. (2006) dataset has two first-order factors and one general factor that accounts for 49.9% of extracted variance. The Gurven et al. (2017) samples both produced two factors in an initial EFA, but the factor extraction process failed for the second-order EFAs for both samples. The inability to test whether the two initial factors could form a general factor makes the Gurven et al. (2017) data ambiguous in regards to the evidence of the presence of g in its Bolivian samples.

Although we did not preregister any exact predictions for our study, we are astonished at the uniformity of these results. We expected before this study began that many samples would produce g, but that there would have been enough samples for us to conduct a post hoc exploratory analysis to investigate why some samples were more likely to produce g than others. With only three samples that did not produce g, we were unable to undertake our plans for exploratory results because g appeared too consistently in the data.

Thus, Spearman’s g appeared in at least 94 of the 97 data sets (97.0%) from 31 countries that we investigated, and the remaining three samples produced ambiguous results. Because these data sets originated in cultures and countries where g would be least likely to appear if it were a cultural artifact, we conclude that general cognitive ability is likely a universal human trait. The characteristics of the original studies that reported these data support this conclusion. For example, some of these data sets were collected by individuals who are skeptical of the existence or primacy of g in general or in non-Western cultures (e.g., Hashmi et al., 2010; Hashmi, Tirmizi, Shah, & Khan, 2011; O’Donnell et al., 2012; Pitchford & Outhwaite, 2016; Stemler et al., 2009; Sternberg et al., 2001, 2002). One would think that these investigators would be most likely to include variables in their data sets that would form an additional factor. Yet, with only three ambiguous exceptions (Grigorenko et al., 2006; Gurven et al., 2017), these researchers’ data still produced g. Additionally, many of these data sets were collected with no intention of searching for g (e.g., Bangirana et al., 2015; Berry et al., 1986; Engle, Klein, Kagan, & Yarbrough, 1977; Kagan et al., 1979; McCoy, Zuilkowski, Yoshikawa, & Fink, 2017; Mourgues et al., 2016; Ord, 1970; Rehna & Hanif, 2017; Reyes et al., 2010; Tan, Reich, Hart, Thuma, & Grigorenko, 2014). And yet a general factor still developed anyway. It is important to recognize, though, that the g factor explained more observed variable variance in some samples than in others.

For those who wish to equate g with a Western view of “intelligence,” this study presents several problems for the argument that Western views of intelligence are too narrow. First, in our search, we discovered many examples of non-Western psychologists using Western intelligence tests with little adaptation and without expressing concern about the tests’ overly narrow measurement techniques. Theorists who argue that the Western perspective of intelligence is too culturally narrow must explain why these authors use Western (or Western-style) intelligence tests and why these tests have found widespread acceptance in the countries we investigated (Oakland, Douglas, & Kane, 2016). Another difficulty for the argument that Western views of intelligence are too narrow is the fact that tests developed in these nonindustrialized, non-Western cultures positively correlate with Western intelligence tests (Mahmood, 2013; van den Briel et al., 2000). This implies that these indigenous instruments are also g-loaded to some extent, which would support Spearman’s (1927) belief in the indifference of the indicator.

One final issue bears mention. Two peer reviewers raised the possibility that developmental differences across age groups could be a confounding variable because a g factor may be weaker in children than adults. To investigate this possibility, we conducted two post hoc nonpreregistered analyses. First, we found the correlation between the age of the sample (either its mean or the midpoint of the sample’s age range) and the variance explained by the first factor in the dataset was r = .127 (r2 = .016, n = 84, p = .256). Because a more discrete developmental change in the presence of strength of a g factor was plausible, we also divided the data sets five age groups: <7 years (10 samples), 7–12.99 years (34 samples), 13–17.99 years (12 samples), 18–40.99 years (21 samples), and ≥41 years (five samples). All of these age groups had a mean first factor that had a similar strength (between 41.79% and 49.63%), and the null hypothesis that all age groups had statistically equal means could not be rejected (p = .654, η2 = .031) These analyses indicate that there was no statistical relationship between sample age and the strength of the g factor in a dataset.

Methodological Discussion

A skeptic of g could postulate that our results are a statistical artifact of the decisions we used to conduct a factor analysis. Some data sets in our study had been subjected to EFA in the past, and the results often differed from ours (Attallah et al., 2014; Bulatao & Reyes-Juan, 1968; Church, Katigbak, & Almario-Velazco, 1985; Conant et al., 1999; Dasen, 1984; Dawson, 1967b; Elwan, 1996; Guthrie, 1963; Humble & Dixon, 2017; Irvine, 1964; Kearney, 1966; Lean & Clements, 1981; McFie, 1961; Miezah, 2015; Orbell, 1981; Rasheed et al., 2017; Ruffieux et al., 2010; Sen, Jensen, Sen, & Arora, 1983; Sukhatunga et al., 2002; van den Briel et al., 2000; Warburton, 1951). In response, we wish to emphasize that we chose procedures a priori that are modern methods accepted among experts in factor analysis (e.g., Fabrigar et al., 1999; Larsen & Warne, 2010; Thompson, 2004; Warne & Larsen, 2014). The use of promax rotation, for example, might be seen as an attempt to favor correlated first-order factors—which are mathematically much more likely to produce a second-order g than orthogonal factors. However, promax rotation does not force factors to be correlated, and indeed uncorrelated factors are possible after a promax rotation. Therefore, the use of promax rotation permitted a variety of potential factor solutions—including uncorrelated factors—and permitted the strong test of g theory that we desired.

Another potential source of criticism would be our methods of retaining the number of factors in a dataset. The original Guttman (1954) rule of retaining all factors with an eigenvalue of 1.0 or greater is the most common method used in the social sciences, probably because it is the default method on many popular statistical analysis packages (Fabrigar et al., 1999). However, the method can greatly overfactor, especially when a dataset has a large number of variables, the sample size is large, and when factor loadings are weak (Warne & Larsen, 2014). These circumstances are commonly found in cognitive data sets, which are frequently plagued by overfactoring (Frazier & Youngstrom, 2007). This is why we chose to use more conservative and accurate methods of retaining the number of factors (Warne & Larsen, 2014). The use of MAP is especially justified by its strong performance in simulation studies and its tendency to rarely overfactor. MAP is insensitive to sample size, the correlation among observed variables, factor loading strength, and the number of observed variables (Warne & Larsen, 2014), all of which varied greatly among the 97 analyzable data sets.

Indeed, it is because of our use of modern methods of factor selection and rotation that we believe that prior researchers have never noticed g as a ubiquitous property of cognitive data in non-Western groups. Many prior researchers used varimax rotation and the original Guttman rule, likely because these methods mathematically and computationally were easier in the days before inexpensive personal computers or because both are the default method in popular statistics packages today. (Additionally, the older data sets predate the invention of promax rotation and/or MAP). But both of these methods obscure the presence of g. As an extreme example, Guthrie’s (1963) data consist of 50 observed variables (the most of any dataset in our study) that produced 22 factors when he subjected them to these procedures. Some of Guthrie’s (1963) factors were weak, uninterpretable, or defined by just one or two variables. In our analyses we found five (using MAP) or 10 (using the modified Guttman rule) first-order factors; when subjected to the second-order EFA, the data clearly produced a single factor with an obvious interpretation: g.

The results of this study are highly unlikely to be a measurement artifact because the original researchers used a wide variety of instruments to measure cognitive skills in examinees. While some of these instruments were adaptations of Western intelligence tests (e.g., Abdelhamid, Gómez-Benito, Abdeltawwab, Bakr, & Kazem, 2017), some samples included variables that were based on Piagetian tasks (e.g., Dasen, 1984; Kagan et al., 1979; Orbell, 1981). Other samples included variables that were created specifically for the examinees’ culture (e.g., Mahmood, 2013; Stemler et al., 2009; Sternberg et al., 2001; van den Briel et al., 2000) or tasks that did not resemble Western intelligence test subtests (Bangirana et al., 2015; Berry et al., 1986; Gauvain & Munroe, 2009). There were also several samples that included measures of academic achievement in their data sets (e.g., Bulatao & Reyes-Juan, 1968; Guthrie, 1963; Irvine, 1964). The fact that g emerged from such a diverse array of measurements supports Spearman’s (1927) belief in the “indifference of the indicator” and shows that any cognitive task will correlate with g to some degree.

Other readers may object to our use of EFA at all, arguing that a truly strong test of g theory would be to create a confirmatory factor analysis (CFA) model in which all scores load onto a general factor. However, we considered and rejected this approach because CFA only tests the model(s) at hand and cannot generate new models from a dataset (Thompson, 2004). In this study, EFA procedures did not “know” that we were adherents to g theory when producing the results. Rather, “EFA methods . . . are designed to ‘let the data speak for themselves,’ that is, to let the structure of the data suggest the most probable factor-analytic model” (Carroll, 1993, p. 82). Thus, if a multifactor model of cognitive abilities were more probable in a dataset than a single g factor, then EFA would be more likely to identify it than a CFA would. The fact that these EFAs so consistently produced g in their data is actually a stronger test of g than a set of CFAs would have been because EFA was more likely to produce a model that disproved g than a CFA would. CFA is also problematic in requiring the analyst to generate a plausible statistical model—a fact that Carroll (1993, p. 82) recognized when he wrote:

It might be argued that I should have used CFA. . . . But in view of wide variability in the quality of the analyses applied in published studies, I could not be certain about what kind of hypotheses ought to be tested on this basis. (Carroll, 1993, p. 82)

We agree with Carroll on this point. CFA also requires exactly specifying the appropriate model(s) to be tested. While this is a positive aspect of CFA in most situations, it was a distinct disadvantage when we were merely trying to establish whether g was present in a dataset that may not have been collected for that purpose. This is because most authors usually did not report a plausible theoretical model for the structure of their observed variables, and there was often insufficient information for us to create our own plausible non-g models that could be compared with a theory of the existence of Spearman’s g in the data.3 Indeed, some researchers did not collect their data with any model of intelligence in mind at all (e.g., McCoy et al., 2017). By having EFA to generate a model for us, we allowed plausible competing models to emerge from each dataset and examined them afterward to see if they supported our theory of the existence of Spearman’s g in non-Western cultures. Another problem with CFA’s requirement of prespecified models is that some theories of cognitive abilities include g as part of a larger theoretical structure of human cognition (e.g., Canivez, 2016; Carroll, 1993). How the non-g parts of a model might relate to g and to the observed variables is rarely clear.

Another advantage to EFA over CFA is that the former uses data to generate a new model atheoretically, and the subjective decisions (e.g., factor rotation method, second-order procedures, standards used to judge the number of factors) in an EFA are easily preregistered, whereas the subjective decisions in a CFA (e.g., when to use modification indices, how to arrange variables into factors, the number of non-g factors to include in a model) often cannot be realistically preregistered—or even anticipated before knowing which variables were collected—in secondary data analysis if the data were not collected in a theoretically coherent fashion (as was often the case for our data sets). By preregistering the subjective decisions in an EFA, we could ensure that subjective decisions could not bias our results into supporting our preferred view of cognitive abilities.

Finally, we want to remind readers that our dataset search and analysis procedures were preregistered and time stamped at the very beginning of the study before we engaged in any search procedures or analyses. This greatly reduces the chance for us to reverse engineer our methods to ensure that they would produce the results we wanted to obtain. Still, deviations from our preregistration occurred. When we deviated from the preregistration protocol, we stated so explicitly in this article, along with our justification for the deviation. Additionally, some unforeseen circumstances presented themselves as we conducted this study. When these circumstances required subjective decisions after we had found the data, we erred on the side of decisions that would maximize the chances that the study would be a strong test of g theory. Again, we have been transparent about all of these unforeseen circumstances and the decisions we made in response to them.

Full paper, maps, references, etc., at the link at the beginning.

There is a generalized bias towards negativity, but with individual-level differences, which appear to be partly pre-dispositional (durable, with correlations with demographic, partisan & personality measures)

Individual-level differences in negativity biases in news selection. Sarah Bachleda et al. Personality and Individual Differences, November 23 2019, 109675. https://doi.org/10.1016/j.paid.2019.109675

Abstract: Literatures across the social sciences highlight the tendency for humans to be more attentive to negative information than to positive information. We focus here on negativity biases in news selection (NBNS) and suggest that this bias varies across individuals and contexts. We introduce a survey-based measure of NBNS which is used to explore the correlates of negative news bias in surveys in the U.S., Canada, and Sweden. We find that some respondents are more prone to NBNS than others. There is evidence of contextual effects, but panel data suggests that some of the individual-level differences persist over time. NBNS likely reflects some combination of long-term personality differences and short-term situational factors, and is systematically related to a number of economic and political attitudes.

Keywords: Political communicationPersonality differencesNews consumptionNegativity bias


1. Durable versus context-driven individual-level variation in negativity biases

"including Lilienfeld and Latzman's (2014) finding that although conservatives are more responsive to negative information on average, both conservatives and liberals respond to negative information when it poses a threat to their partisan identity; or Federico, Johnston and Lavine's (2014) finding that evidence of negativity biases will be conditional on political engagement."


6. Discussion

There is reason to expect that individual-level variation in negativity biases has an important and durable impact on individuals’ news media use, as well as on a range of economic and political attitudes. This paper has taken a first step toward measuring a negativity bias in news selection. We find that while on balance there is a bias towards negativity, there are individual-level differences. These differences appear to be partly pre-dispositional; that is, they appear to be durable, demonstrated both by correlations with demographic, partisan and personality measures, and by within-respondent correlations across time. We also find that these individual-level differences are correlated with a variety of economic and political attitudes. We take these results as evidence of the potential importance of negativity biases in news selection (NBNS) in understanding attitudes about governments, the economy, and other politically and economically-relevant attitudes. We also suspect that NBNS moderates the impact of news content – those who are high in NBNS may select into a rather different information stream than those who are low in NBNS, which could subsequently shape their political perspectives. Although this application of the measure is not tested here, we thus see disentangling the relationship between political news selection and political preferences as an important avenue for future research. There is also potential for work that explores the degree to which more nuanced variation in tone – i.e., not just positive or negative, but gradations across that range – matters for story selection and measures of negativity biases. Our headlines do not vary in tone much within the negative and positive categories (see Appendix Fig. 2); this was done by design. But past work has suggested nonlinearities in negativity biases (e.g., Ito and Cacioppo, 2005), and these could be more fully explored using headlines that vary systematically in degrees of positivity or negativity. Finally, an exploration of the relationship between NBNS and other measures of negativity biases will be critical for future work. Given that other more standard measures of negativity biases are primarily labbased, we have not examined them in the survey data used here. However, understanding the extent to which NBNS is a domain-specific negativity bias, versus the consequence of a more domain-general bias, requires further research. Our results provide only a first step in this direction. In doing so, however, we regard the preceding analyses as a first signal that individual-level differences in news preferences may be one way in which personality differences are relevant to political attitudes and behavior.



Samples

U.S. Sample
Data for the U.S. study were collected as part of an online panel survey from a sample provided by Qualtrics, which recruited subjects using ClearVoice research. ClearVoice maintains a standing panel of survey respondents who were recruited to the platform through a combination of targeted emails, advertisements, and website intercepts. These individuals then opt-in to taking surveys and are recruited to participate in individual studies either by email or by clicking on a dashboard link. ClearVoice sent emails to 61,865 panelists with the goal of recruiting a broad national sample of at least 3,667 Americans to participate in the study.

Swedish Sample
Data for the first Swedish sample come from the Citizen Panel (original Swedish name: Medborgarpanelen – MP), which is a panel survey fielded online by the Laboratory of Opinion Research (LORE). Specifically, the data come from Citizen Panel 16 (MP16), which was fielded between June 9 and June 30, 2015. The panel used a mixed sampling design whereby 84 percent of the gross sample were opt-in and the remaining 16 percent were probability based. The panel wave included five separate modules and our data come from module 3 (Negativity Biases). This module yielded 12,867 complete responses for an AAPOR participation rate (RR5) of 92%.

Data for the second Swedish sample also come from the Citizen Panel. Specifically, the data come from Citizen Panel 29 (MP29), which was fielded between March 22 and April 16, 2018. The panel used a mixed sampling design whereby 76 percent of the gross sample were opt-in and the remaining 24 percent were probability based. The panel wave included five separate modules and our data come from module 2 (Negativity Biases in News Selection). Additional information about the Citizen Panels can be found at http://lore.gu.se/surveys/citizen.

Canadian Sample
The Canadian data come from the 2015 Canadian Election Study. Full documentation for the study can be found at: http://ces-eec.arts.ubc.ca/ english-section/surveys/. The study was funded by the Social Sciences and Humanities Research Council of Canada.