Although the importance of stream condition for leaf litter decomposition has been extensively studied, little is known about how processing rates change in response to altered riparian vegetation community composition. We investigated patterns of plant litter input and decomposition across 20 boreal headwater streams that varied in proportions of riparian deciduous and coniferous trees. We measured a suite of in‐stream physical and chemical characteristics, as well as the amount and type of litter inputs from riparian vegetation, and related these to decomposition rates of native  and introduced (lodgepole pine) litter species incubated in coarse‐ and fine‐mesh bags. Total litter inputs ranged more than fivefold among sites and increased with the proportion of deciduous vegetation in the riparian zone. In line with differences in initial litter quality, mean decomposition rate was highest for alder, followed by birch, spruce, and lodgepole pine . Further, these rates were greater in coarse‐mesh bags that allow colonization by macroinvertebrates. Variance in decomposition rate among sites for different species was best explained by different sets of environmental conditions, but litter‐input composition  was overall highly important. On average, native litter decomposed faster in sites with higher‐quality litter input and (with the exception of spruce) higher concentrations of dissolved nutrients and open canopies. By contrast, lodgepole pine decomposed more rapidly in sites receiving lower‐quality litter inputs. Birch litter decomposition rate in coarse‐mesh bags was best predicted by the same environmental variables as in fine‐mesh bags, with additional positive influences of macroinvertebrate species richness. Hence, to facilitate energy turnover in boreal headwaters, forest management with focus on conifer production should aim at increasing the presence of native deciduous trees along streams, as they promote conditions that favor higher decomposition rates of terrestrial plant litter. For studies that have tested how litter mixing regulates decomposition rates, results are equivocal: mixing high‐quality with low‐quality litter species may promote, retard, or show no effect on, overall decomposition rates  now cover close to 600,000 ha, or approximately 3%, of the productive forest area in Sweden  and a corresponding shift from dominance by deciduous to coniferous species in the riparian zone , Scots pine (Pinus sylvestris), birch (Betula sp.), willow , and alder (Alnus sp.). For each stream, a 100‐m reach dominated by riffles was selected as the study site. Along each of these reaches, we surveyed all riparian trees within 5 m of the stream bank to generate estimates of relative abundance for different broad‐leaved and coniferous species , light conditions  were estimated in August before leaf abscission, using a spherical densitometer. We measured stream width, depth (at both edges and in the middle), and velocity at three locations along each study reach. Due to seasonally fluctuating water levels, depth and stream velocity were measured three times during the study period , to provide mean depth and velocity values for the entire period. Water samples were taken four times during the period  for analysis of conductivity and pH, and to measure ambient concentrations of dissolved organic carbon (DOC), soluble reactive phosphorus (SRP), nitrate (NO3−), ammonium (NH4+), and total nitrogen (TN). Water samples were filtered in the field with a 0.45‐μm nylon membrane filter  and stored on ice until transferred to the laboratory refrigerator (4°C) or freezer  within 6 hr.The study was carried out from late May to early November, 2014. We monitored stream temperature continuously from mid‐August to early November, using HOBO2 in size) on the bank immediately adjacent to each reach in late May  yielded C and N quantities. All chemical analyses were performed by a certified laboratory at the Swedish University of Agricultural Science, Umeå, Sweden.To investigate leaf litter decomposition rates with respect to variation in riparian community composition, we selected litter from four different tree species. Three of these species  are native to northern Sweden, while lodgepole pine is introduced  bags, which excludes macroinvertebrate access. Another set of 12 × 17‐cm coarse‐mesh (5 mm) bags was used to measure leaf litter decomposition in the presence of both macroinvertebrates and microbes. Litter from all four species was used for the fine‐mesh bags, while only birch was used for the coarse‐mesh bags. Stalks on birch and alder leaves were removed before weighing and insertion into the bags, to increase accuracy in measures of consumable leaf litter mass  were affixed at randomly selected locations to an anchored chain. Bags containing coniferous litter  were introduced in mid‐August, while bags containing deciduous litter were introduced in the end of September, to allow each species to reach similar stages of decomposition when collected. All litterbags were collected in mid‐November, rendering a total of 87–91 field days for coniferous litter and 53–56 field days for deciduous litter. The litterbags were brought to the laboratory and were frozen at −18°C, for later processing and analyses.k) was calculated, using the negative exponential decay model (k = ln[Mt/M0]/t), where Mt is the AFDM at time t and M0 is the initial AFDM.In the laboratory, litter removed from the bags was rinsed before being oven‐dried at 60°C for 48 hr. The dried litter was then combusted to obtain AFDM, following the procedure described above. Litter AFDM, used later in calculations of decomposition rates, was obtained by subtracting ash weight from dry mass. Following the same procedure, initial AFDM was obtained by combusting three 1.0‐g samples of each of the four litter species. To calculate decomposition rates for the four litter species, and for both fine‐ and coarse‐mesh bags, the decomposition constant , NH4+ (Method G‐171‐96 Rev. 12), and SRP (Method G‐297‐03 Rev. 1) were analyzed using a SEAL Analytical AutoAnalyzer 3 . Dissolved organic nitrogen (DON) was estimated as the difference between TN and total inorganic N . From this, we calculated the DOC:DON ratio (mass) as a measure of “dissolved organic matter (DOM) character” that has been shown to be positively related to “humic‐like” compounds and inversely related to rates of microbial DOM degradation . 2.6We used mixed‐effects models with site as a random factor to compare decomposition rates between fine‐ and coarse‐mesh litterbags (for birch only), among litter species , and to assess whether there were significant differences in input among different litter types and between seasons. This test was followed by pairwise comparisons using Tukey's HSD test. We used principal component analysis (PCA) to explore associations among seasonal (summer and autumn) patterns of litter inputs and physical and chemicals conditions. Moreover, to obtain litter‐input compositional variables  that could be used in subsequent statistical analyses, we performed PCAs using only the summer or autumn senesced mass of different litter types. These compositional measures, which are distinct from absolute abundances of single litter species, were used as predictor variables for both microbial‐  and macroinvertebrate‐mediated (birch) litter decomposition rates. Hence, instead of only using the input of each litter type as a predictor variable, these component 1 scores allow us to test effects of the entire litter‐input community  on litter decomposition rates.R2), loadings of the independent variables, and the variable influence on projection (VIP). The independent variable loading describes the relative strength and direction of the relationship between independent and response variable. The VIP value summarizes the importance of each variable. In the models, the limit for a variable to be included in the final model was a VIP value at 0.7  regressions to explore the predictors of litter decomposition from the suite of environmental variables. PLS regression relates two data matrices  to each other by a linear multivariate model and produces latent variables (PLS components) extracted from predictor variables that maximize the explained variance in the dependent variables. PLS regression is especially useful when predictor variables are correlated, and when there are more predictor variables than observations . DOC concentrations varied sixfold among sites (6.85–35.73 mg/L), the mass ratio of DOC:DON ranged from 48 to 80, pH varied from acidic (4.9) to circumneutral (6.4), SRP concentrations from 1.1 to 11.3 μg/L, and DIN concentrations were below 50 μg/L, except for at BCC (Table S2). Among sites, mean species richness, abundance, and biomass of the macroinvertebrate detritivores, based on individuals found in coarse‐mesh litterbags, ranged from 0.6 to 7.6 species, 6.0 to 113.2 individuals, and 1.2 to 16.3 mg DW, respectively (Table S3).Among sites, mean stream width ranged from 39.7 to 185.0 cm, depth from 5.7 to 26.6 cm, velocity from 0.03 to 0.31 m/s, and water temperature from 3.8 to 4.9°C (see Table S1). Canopy openness was mostly ≤15%, with the exception of the recently harvested site . In terms of composition, spruce and birch were the most dominant riparian trees, and the dominant forms of conifer and broad‐leaved species, respectively (Table S1), and the relative abundance of these species was inversely correlated among sites (2 (mean ± 1 SD). Total input increased across sites with the relative abundance of broad‐leaved trees in the riparian zone  and decreased with the abundance of spruce . Similarly, the proportion of inputs composed of broad‐leaved material declined strongly among sites with spruce cover . Seasonally, mean riparian plant litter input was five times lower in summer than in autumn, largely driven by significantly greater inputs of birch litter during autumn . During autumn, birch contributed the greatest amount of litter on average, followed by spruce, “other,” and alder, while pine and SWD contributed very little to the total input  to 265.8 (±79.6) g AFDM /m2 mean ±  SD. Tota2 mean ±  SD. TotaNO3− and DIN, while input of “other” litter was positively related to these environmental conditions. Further, in both summer and autumn, input of low‐quality litter  was positively associated with the DOC:DON ratio. Total litter input was most strongly associated with spruce, pine, and “other” litter input in summer  in the fine‐mesh litterbags differed significantly among all litter species , with alder showing the highest k , followed by birch , spruce , and lodgepole pine  than in coarse‐mesh  litterbags , which on average lost 52% of the initial mass over 53–56 days (range of mass loss for coarse mesh: 39–70%).Mass loss rates  and autumn input of SWD , and with nitrogen concentrations. Also k of spruce litter was positively associated with high‐quality litter input, that is, birch litter input  and a mixture of birch, “other,” and alder , and negatively related to input of low‐quality litter . However, in contrast to alder and birch, spruce litter showed higher k in streams with high pH and low DOC concentrations, and canopy openness did not show a significant association , and negatively associated with high‐quality litter . Between these extremes, the gradual replacement of deciduous trees by coniferous species resulted in predictable changes in the amount, quality, and timing of litter inputs. In this way, both the short‐term effects of forest management , the set of environmental factors that best predicted rates of litter decomposition differed among species. For both alder and birch litter, decomposition rates tended to increase with N concentrations, a result consistent with patterns of N limitation widely observed in other studies of litter processing  in adjacent and downstream aquatic food webs in this region. Ultimately, by enhancing both litter supply and biological processing rates, management decisions that lead to greater cover by deciduous riparian trees have the potential to strongly influence the transfer of energy through stream food webs in the boreal region.http://www.slu.se/en/krycklan).Data used in the study are housed in the Krycklan Experimental Catchment database (None declared. Click here for additional data file.


P = 0.04), while TI-RADS had higher sensitivity (P = 0.02). In nodules d > 20 mm, the specificity of ATA patterns was higher than TIRADS (P = 0.003). Our results indicated that nodule sizes may influence the diagnostic performance of TIRADS and ATA ultrasound patterns. The ATA patterns may yield higher specificity than TIRADS, especially in nodules larger than 20 mm.To evaluate the impact of thyroid nodule sizes on the diagnostic performance of thyroid imaging reporting and data system (TIRADS) and ultrasound patterns of 2015 American Thyroid Association (ATA) guidelines. Total 734 patients with 962 thyroid nodules were recruited in this retrospective study. All nodules were divided into three groups according to the maximal diameter . The ultrasound images were categorized based on TIRADS and ATA ultrasound patterns, respectively. A total of 931 (96.8%) and 906 (94.2%) patterns met the criteria for TIRADS and ATA ultrasound patterns. The AUC (0.849) and sensitivity (85.3%) of TIRADS were highest in d = 10–20 mm group. However, ATA had highest AUC (0.839) and specificity (89.8%) in d > 20 mm group. ATA ultrasound patterns had higher specificity ( However, only 10–15% of these nodules are malignantet al. have compared the diagnostic efficiency between the new ATA ultrasound patterns and the Korean TIRADS proposed in differentiating malignancy from benign lesions, indicating that ATA classifications may yield higher specificity, while TIRADS may offer a relatively higher sensitivityRecently, the 2015 American Thyroid Association (ATA) guidelines constructed a new ultrasound risk stratification model from very low suspicion to high suspicion for malignancy according to sonographic featuresP > 0.05). The malignancy rates of nodules d > 20 mm, d = 10–20 mm and d < 10 mm were 22.2%, 45.7% and 48.5%, respectively.A total of 962 thyroid nodules in 734 patients were included in our study with 578 women and 156 men. The average age was 46.8 ± 13.1 years old and the mean diameter of the nodules was 17.7 ± 12.8 mm. All 375 malignant lesions and 328 benign nodules were confirmed by histopathology. The remaining 259 nodules were regarded as benign lesions due to the repeated benign cytology or follow-up ultrasound after the first benign cytology . The epiA total of 931 patterns (96.8%) were able to be categorized based on TIRADS classification. The malignancy rates of TIRADS 2, 3, 4A, 4B and 5 were 0, 14.1% (62 of 439 nodules), 50.0% (118 of 236 nodules), 80.4% (156 of 194 nodules) and 100.0% (27 of 27 nodules), respectively, with significant differences between categories (P < 0.001). The correlations between the TIRADS classification and final diagnosis according to nodule size were shown in The remaining 31 nodules couldn’t be categorized, of which 12 (38.7%) nodules were validated as PTCs by surgery. Among them, there was 11, 12, 8 cases in d < 10 mm, 10–20 mm and > 20 mm group, respectively, indicating that nodule size had no influence on this aspect. Moreover, in these nodules beyond the range of TIRADS classifications, nodules with hypoechogenicity and tall-than-wide shape had 100% malignancy risk, hypoechogenicity accompanied with irregular shape and ill-defined margin were likely to have 75.0% malignancy rate.P < .001). The correlations between the ATA ultrasound patterns and final diagnosis according to size were shown in A total of 906 patterns (94.2%) were able to be categorized based on ATA ultrasound patterns. According to histopathology or follow-up results, the malignancy rates of the nodules with very low, low, intermediate, and high suspicion for malignancy were 5.3%, 10.0%, 21.8% and 71.8%, respectively, with significant differences between patterns  nodules were validated as PTCs by surgery. There was 17, 20, 19 cases in d ≤ 10 mm, 10–20 mm and > 20 mm group, respectively, indicating that nodule size had little relations with the nodules that couldn’t be classified by ATA ultrasound patterns. Furthermore, we compared ultrasound features of these benign and malignant lesions and found that hyper-/isoechogenecity accompanied with irregular shape had much tendency to be malignant (42.9%).P = 0.04), while TIRADS had a higher sensitivity . The AUC was higher in TIRADS than 2015 ATA classification, though, not significantly .Compared to TIRADS, the 2015 ATA guidelines yielded a significant higher specificity . In d = 10–20 mm group, there were no significant difference between the two models . While in d > 20 mm group, the specificity of ATA ultrasound patterns was significantly higher compared with TIRADS classification , while the sensitivity and AUC showed no significant difference between the two models   .In this study, we evaluated the impact of thyroid nodule sizes on the diagnostic performance of newly published ultrasound patterns of 2015 ATA guidelines and original TIRADS classifications. We found that TIRADS performed best for differentiating nodules between 10–20 mm, while ATA ultrasound patterns had best value in lesions larger than 20 mm. The ATA ultrasound patterns may yield higher specificity, especially in nodules larger than 20 mm.TIRADS established by Horvath had been widely applied in clinical setting for the evaluation of thyroid nodules. Based on 10 US patterns, TIRADS related the rate of malignancy according to the patterns3et al. considered that solid iso/hyperechoic nodule with any calcification beard a malignancy risk of 24.7%The 2015 ATA guidelines for patients with thyroid nodules established a 5-tier risk classification of ultrasound patterns by combining several individual sonographic characteristics11et al.P = 0.024), whereas specificity, PPV, and accuracy were higher with the ATA guidelines . Similar to Yoon’s study, our study found that original TIRADS model had a higher sensitivity (P = 0.02), while specificity were higher with the ATA ultrasound patterns (P = 0.04). In addition, Yoon et al. also revealed that 44 (3.4%) patterns did not meet the criteria for any ATA pattern including hyper- to isoechoic solid or partially cystic nodules with microlobulated or irregular margins, microcalcifiations or mixed calcifiations, or nonparallel shape and the malignancy risk was 18.2%.Recently, Yoon et al.et al.The novel finding in our study was that nodular size markedly influenced the diagnostic performance of the TIRADS and 2015 ATA US patterns in differentiating benign and malignant lesions. The value of US between large lesions and small ones was controversial. Andrej The limitations of our study should also be addressed. Firstly, all classifications were performed based on the static images of US, which might cause misinterpretation of ultrasound classification. Secondly, description of features was reported by different radiologists, which may cause inter-observer variability. Thirdly, this was a retrospective study, the selection bias such as patients who underwent thyroid surgery and gender bias  may cause the high percentage of carcinomas (39.0%), resulting in the overestimation of PPV and underestimation of NPV, both in TIRADS and ATA ultrasound patterns16In conclusion, both TIRADS and the 2015 ATA ultrasound patterns provide effective malignancy risk stratification for thyroid nodules. Nodule sizes may influence the diagnostic performance of the two models. The TIRADS showed best value in nodule between 10–20 mm, while ATA patterns had highest value in lesions larger than 20 mm. Both models are less reliable in lesions smaller than 10 mm. The ATA patterns may yield higher specificity than TIRADS, especially in nodules larger than 20 mm. Those nodules beyond the range of TIRADS categories and ATA patterns had little relation with nodule size and may still have a relatively high risk of malignancy (38.7% and 28.6%). However, due to the limitations of this study, our findings still need to be further validated in the clinical practice.a) patients who underwent thyroid surgery regardless of cytologic results, (b) patients who underwent fine-needle aspiration cytology at least two times within a 1-year interval for benign thyroid lesions, (c) patients who had benign results on cytology and showed no change or decreased size at follow-up US for at least a year (7). The increase in size was defined as more than a 50% change in volume or a 20% increase in at least two nodule dimensions with a minimal increase of 2 mm in solid nodules or in the solid portion of mixed cystic-solid nodules (8). A total of 734 patients with 962 nodules  were included preliminarily. There were 156 men  and 578 women . All nodules were divided into three groups according to the maximal diameter . Informed consent was obtained from all patients and the study was performed in accordance with the ethical guidelines of the Helsinki Declaration and approved by the First Affiliated Hospital with Nanjing Medical University ethics review committee (2012-SR-058).This retrospective study was based on patient data collected from eight tertiary hospitals around Jiangsu province in China from January 6, 2014 to December 20, 2014. A total of consecutive patients underwent US-guided FNAB or thyroidectomy for thyroid nodules. Patients who met the following criteria by reviewing US patterns and clinical data were included in this study:  test or Fisher exact test. Compared to the final diagnosis (according to pathology or follow-up results), the sensitivity, specificity, positive predictive value (PPV) and negative predictive value (NPV) were calculated for each method. Receiver Operating Characteristic (ROC) curve analysis with MedCalc 11.4.2.0 software  was used to compare the two models and to determine the optimal cut-off value between benign and malignant nodules. Area under the curves (AUCs) and P value were calculated. P < 0.05 was considered significant in all tests.Statistical analysis was performed using SPSS 20.0 software . All quantitative values were expressed as means ± SD. Differences in the values of continuous variables between three groups were evaluated by the one-way ANOVA test or non-parametric test. Differences in the distribution of categorical variables between groups were evaluated by the 2-tailed Chi-square .Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Although it seems reasonable to assume that activating patriotism might motivate citizens to cooperate with the state in reaching societal goals, the empirical evidence supporting this contention is based mostly on correlational rather than experimental studies. In addition, little is known on whether patriotism can be manipulated without simultaneously triggering nationalism and on the psychological processes which determine the patriotism‐cooperation relation. This current article reports results of one survey and three experiments that manipulate patriotism by displaying either a national flag or national landscapes or by priming national achievements. The outcomes indicate that reported and manipulated patriotism indirectly increase tax compliance, although the national flag also increases nationalism. National achievements, on the other hand, seemingly increases trust in national public institutions and the voluntary motivation to cooperate, whereas national landscapes only increase the voluntary motivation to cooperate. Hence, it is possible to increase social capital in the form of trust and cooperation through patriotism without fostering nationalism as well. The state and its public institutions depend on such citizen cooperation as tax compliance, which might potentially be increased by using patriotism as a promotional tool . Most participants had a university degree (33.3%), were employed in the private sector (45.2%), and reported a monthly income between 1,501 and 3,000 Euro (40.5%). When asked to indicate their experience with tax authorities on a 7‐point scale, participants rated themselves overall as rather experienced .Our sample was made up of 84 Austrians, the majority of whom were male (66.7%), with an average age of 44.13  which impacts voluntary cooperation , and in turn impacts tax compliance  or pride and voluntary cooperation alone , however, is not significant. The results for identification with the local community show that identification with the local community triggers voluntary cooperation , which in turn impacts tax compliance . Other indirect paths are not significant  or marginally not significant . Nationalism seems also to be indirectly connected to tax compliance: it triggers trust in tax authorities , which fosters voluntary cooperation , and in turn tax compliance . The other indirect effects are not significant . In sum, we identify only indirect relations between identification with the community and tax‐compliance intention: national pride and nationalism are related to trust in tax authorities, which fuels voluntary motivation to cooperate and in turn tax compliance. Identification with the local community, in contrast, seems to be related to voluntary cooperation (and less to reason‐based trust), which fuels tax compliance. Hence, whereas national pride and nationalism tend to trigger deliberate processes in which evaluations about whether or not to trust tax authorities play a role, identification with the local community seems to be a rather emotional process that fuels patriotic motivations based on civic duty and not deliberate evaluation whether or not tax authorities can be trusted.The results indicate that, as expected, national pride leads to trust . Most participants had a university degree (34.5%), were employed in the private sector (36.4%), and had a monthly income of between 1,501 and 3,000 Euro. Again, when asked on a 7‐point scale about their experience with the tax authorities, participants rated themselves overall as rather experienced .The first experimental sample consisted of 110 Austrians, the majority of them male (70.9%), with an average age of 41.44 . Next, participants were presented with the same tax‐compliance scenario as in the survey and filled in the same items on national pride, identification with the local community, nationalism (Cronbachs’ α = .80), trust in tax authorities (Cronbachs’ α = .92), voluntary cooperation (Cronbachs’ α = .92), and sociodemographic characteristics.t‐tests, with the national versus fictitious flag as the independent variable and national pride, identification with the local community, nationalism, trust, voluntary cooperation, and tax‐compliance intention as dependent variables. The national flag led to greater national pride than the fictitious flag (t(101) = 2.31, p = .02, Cohen's d = .47) and also to marginally more nationalism (t(98) = 1.76, p = .08, Cohen's d = .35). No differences emerged for identification with the local community (t(102) = 1.15, p = .25). The results also indicate, however, that the national flag induced no higher trust in tax authorities (t(90) = 0.65, p = .52), tax‐compliance intentions (t(108) = 1.14, p = .26), or voluntary cooperation (t(84) = 1.56, p = .12) than the fictitious flag. Respective means and standard deviations of the experimental groups are displayed in Table The differences between the experimental groups were identified using p = .02), which in turn impacts tax compliance , 95% CI . The results also show that national pride leads to more reason‐based trust in tax authorities  and in turn fosters voluntary cooperation  and tax compliance . The fact that no path without national pride is significant  implies that the national flag increases intended tax compliance indirectly through an explicit and deliberative process based on trust in tax authorities. Given the t‐test result of no impact for the national flag on identification with the local community, no indirect effect via identification with the local community is possible. Also only tendencies but no significant indirect effects emerge for elicited nationalism . These results indicate that an indirect effect similar to that found in the survey study occurs only if the national flag impacts national pride thereby indirectly fostering trust and voluntary cooperation and in turn tax‐compliance intention. No such indirect effect on tax compliance is observed if the flag impacts nationalism.The corresponding regression analyses, again on a bootstrap sample of 1,000 participants, indicate an indirect effect. The national flag elicits national pride . Most participants had a university degree (40.4%), were employed in the private sector (38.4%), and reported a monthly income between 1,501 and 3,000 Euro (40.4%). When asked to indicate their experience with tax authorities on a 7‐point scale, participants rated themselves overall as rather experienced .Our sample comprised 99 Austrians, the majority of whom were male (63.6%), with an average age of 43.09  = 1.84, p = .07, Cohen's d = .39), but no differences are observable for identification with the local community (t(92) = 1.06, p = .29) or nationalism (t(91) = −1.00, p = .32). Positive priming also generated more trust in tax authorities (t(97) = 1.89, p = .01, Cohen's d = .57) than did negative priming and also lead to marginally more tax‐compliance intention  = 3.58, p = .06, Cohen's d = .38). No differences emerge, however, for voluntary cooperation (t(81) = 0.90, p = .37). Respective means and standard deviations of the experimental groups are illustrated in Table The differences between experimental groups were again identified using p = .07) which influences trust  and in turn impacts voluntary motivation to cooperate  and thus, tax compliance . The other paths with national pride are not significant . However, the analysis also suggests paths without national pride: Compared to negative priming, positive priming leads to more trust in tax authorities  which in turn impacts tax compliance  or impacts first voluntary cooperation  and then tax‐compliance intention . Likewise, the t‐test results indicate that priming has no impact on identification with the local community or nationalism, meaning that no indirect effect is possible via identification or nationalism. Taken together, these findings indicate that priming increases national pride and in turn, similar to the survey study, trust, voluntary motivation to cooperate, and tax compliance. However, the present results also suggest that reason‐based trust in tax authorities seems to be particularly important for the indirect impact of the priming on tax compliance. Thus, the priming of national achievements seems to increase tax‐compliance intention through an explicit deliberative process.To identify the underlying psychological mechanisms, we again tested for indirect effects using a bootstrap sample of 1,000 participants. The results indicate that the priming impacts national pride . As before, most participants had a university degree (40.5%), were employed in the private economy (37.8%), and earned between 1,500 and 3,000 Euro (55.4%) a month. When asked on a 7‐point scale about their experience with the tax authorities, participants again rated themselves overall as rather experienced .The third experimental sample consisted of 74 Austrians, the majority male (91.9%), with an average age of 43.14  did not recognize or (2) recognized the landscape shown.As in the national flag experiment and priming task, the dependent variables were tax‐compliance intention, national pride, identification with the local community, nationalism (Cronbachs’ α = .83), trust in tax authorities (Cronbachs’ α = .83), voluntary cooperation (Cronbachs’ α = .89), and sociodemographic characteristics. Participants were also asked to indicate whether they recognized the landscapes presented and to write down what they thought each picture showed. Based on this information, we were able to calculate a F = 3.93, p = .05, η2 = .05), the national (Austrian) landscape pictures did not generate more national pride  = 0.14, p = .71) or nationalism  = 1.02, p = .32) than the foreign  landscape controls. The results also reveal no differences between the conditions for trust in tax authorities  = 0.29, p = .60) or voluntary cooperation  = 1.68, p = .20). They do, however, indicate that national landscapes led to higher tax‐compliance intention than foreign landscapes  = 5.56, p = .02, η2= .07). Respective means and standard deviations of the experimental groups are shown in Table The differences between experimental groups were identified using ANCOVA, with national versus foreign landscape as the independent variable; national pride, identification with the local community, nationalism, trust, voluntary cooperation, and tax compliance as dependent variables; and recognition as the control variable. Although the national landscape pictures led to more identification with the local community  = 8.58, p = .01, η2= .11) also indicates that nonrecognition increases tax‐compliance intention  in a similar manner in both experimental conditions . The covariate is insignificant, however, for pride, identification with the local community, nationalism, trust, and voluntary cooperation .The significant recognition covariate for tax‐compliance intention (p = .05), which impacts voluntary tax cooperation  and in turn tax‐compliance intention . No other mediation models produced significant results: identification with the local community , identification with the local community and trust in tax authorities , trust in tax authorities , trust and voluntary cooperation , or via voluntary cooperation . According to the ANCOVA results, the landscapes have no impact on national pride or nationalism and can thus have no indirect effect via pride or nationalism. These findings imply that the landscapes’ impact on tax‐compliance intention is indirect and rather implicit. This assumption is also supported by the finding that the landscapes particularly increased tax‐compliance intentions of those participants who did not deliberately recognize them. As with the survey results, identification with the local community triggers the voluntary motivation to cooperate and in turn tax‐compliance intention without rational consideration of whether national tax authorities can be trusted.An additional analysis with recognition as a control variable was aimed at identifying the mechanisms determining the landscapes’ impact on tax compliance. This analysis, again conducted on a bootstrap sample of 1,000, points to an indirect impact via identification with the local community  compared to identification on the national level  run a smaller risk of triggering nationalism next to patriotism , it is only significant in the foreign landscape condition. Concerning the foreign landscapes, it is perhaps not surprising that explicit recognition of a foreign country is negatively related to national tax compliance. Nonrecognition of a foreign landscape might be perceived as a neutral landscape which does not influence tax‐compliance intentions. In contrast, the explicit recognition of a foreign country might lead to the impression that tax compliance does not support one's own community but a foreign community, a perception which might reduce tax‐compliance intentions. However, nonrecognition of national landscapes also tends to be related with more tax‐compliance intention than recognition of national landscapes. One could speculate that recognized national landscapes might be linked to concrete and sometimes negative attitudes which in turn reduce tax compliance. In contrast, nonrecognized national landscapes might unconsciously only be perceived as familiar and therefore might implicitly elicit more positive emotions and more tax‐compliance intentions. However, these are only speculations, and the underlying sample sizes necessitate approaching such arguments with caution. Future studies are needed to replicate the current finding and to shed light on the reasons why nonrecognition of landscapes might be positively related to cooperation with the state.The present research is subject to certain limitations. First, although public institutions like tax authorities need the cooperation of foreign citizens, all participants in our research had Austrian citizenship. This choice was made based on the assumption that the Austrian national flag, national achievements, and/or national landscapes might not be as familiar to foreigners and might not, therefore, have equal power to increase their identification, trust, and cooperation. Nonetheless, future studies should investigate how identification with the country or region of residence can be triggered to increase trust and cooperation in foreigners.Second, unlike the findings in previous research Torgler, , the surFinally, our research design does not take into account the fact that Austria has one of the highest tax morale and tax‐compliance rates across the globe  induce nationalism and should be avoided, others, such as national achievements or national landscapes, can increase citizens’ identification and cooperation with their community. Nevertheless, if a long‐term impact on patriotism, trust, and cooperation is pursued, campaigns should use salient and explicitly positive evaluations of state achievements Bar‐Tal, . Such anAdditional supporting information may be found in the online version of this article at the publisher's website:Photos of fictional country and Austrian flagsPhotos of Australian and Austrian national landscapesClick here for additional data file.


The effects of video games on children’s psychosocial development remain the focus of debate. At two timepoints, 1 year apart, 194 children  reported their gaming frequency, and their tendencies to play violent video games, and to game (a) cooperatively and (b) competitively; likewise, parents reported their children’s psychosocial health. Gaming at time one was associated with increases in emotion problems. Violent gaming was not associated with psychosocial changes. Cooperative gaming was not associated with changes in prosocial behavior. Finally, competitive gaming was associated with decreases in prosocial behavior, but only among children who played video games with high frequency. Thus, gaming frequency was related to increases in internalizing but not externalizing, attention, or peer problems, violent gaming was not associated with increases in externalizing problems, and for children playing approximately 8 h or more per week, frequent competitive gaming may be a risk factor for decreasing prosocial behavior. We argue that replication is needed and that future research should better distinguish between different forms of gaming for more nuanced and generalizable insight. Video games have rapidly become a universal aspect of child development  externalizing problems, (H2) internalizing problems, (H3) peer problems, and (H4) overall psychosocial problems by the second time point. Given the lack of consensus in the research, no predictions were made for the influence of gaming on hyperactivity and inattention, or on prosocial behavior, although exploratory analyses were conducted. We also explored the potential relationships between violent video game content and both externalizing problems and prosocial behavior. Finally, we also hypothesized that (H5a) cooperative gaming at the first time point would be associated with increases in prosocial behavior, whereas (H5b) competitive gaming at the first time point would be associated with decreases in prosocial behavior.T1 and T2; days between visits: range 265–510, M = 392.22, SD = 59.05). Participants were recruited from a pool of 298 participants already participating in research which tracked children’s psycho-social health ; 86.6% of parent reporters were female (n = 168); with the exception of three adopted mothers and one adopted father, all parents were the child’s biological parent; finally, 96.9% of parents were Dutch (n = 188), with the others coming from Suriname (n = 1) or nearby European countries (n = 5). The study’s procedures were approved by the Behavioural Science Institute’s Ethical Review Board under the Radboud University, and informed consent forms were obtained from parents at both timepoints. Descriptive statistics for the sample at T1 and T2 are reported in Table T1 declined to participate at T2. Additionally, data from ten parent reports were missing at T2 because their data was not properly saved by the recording software. With the exception of five parents, all parent reports were provided by the same parent at T1 and T2. Among parents, education level was low for 6.7%, medium for 30.4%, and high for 60.3%Data were collected during home visits 1 year apart (T1 and T2 respectively.Children provided self-reports during a face-to-face interview with an experimenter. Parents provided their survey responses via an online questionnaire. Families were rewarded a 20 and 30 Euro voucher check (per child) for their participation at Not true to Very true) and is comprised of five sub-scales: (a) internalizing problems, (b) externalizing problems, (c) hyperactivity/inattention, (d) peer relationship problems, and (e) prosocial behavior. Consistent with Stone and colleagues  internalizing problems ; (b) externalizing problems ; (c) hyperactivity/inattention ; (d) peer problems ; and (e) prosocial behavior . All sub-scales consist of five items with sum scores being calculated for each sub-scale. The SDQ also includes a total difficulties score, calculated as the sum scores of all scales except for prosocial behavior ; this reflects children’s general psychosocial health. Descriptive statistics for the SDQ measures are presented in Table Psychosocial health was measured by parent’s reports on the Dutch version of the Strengths and Difficulties Questionnaire  Parental reports for the number of hours their child plays on average per week; (2) child reports for the number of hours they had played video games during the past week; (3) children’s ability to recall their gaming hours across a whole week was scaffolded by an additional measure of gaming frequency: in interviews, children looked over a calendar with the experimenter and indicated for each day over the past full week whether or not they had played a video game in the morning, afternoon, and evening. “Video games” were explicitly described to parents and children as any game that can be played on an electronic device, and several example games were listed.T1: parent reports: M = 5.76, SD = 3.87, outliers n = 4; child reports: M = 4.86, SD = 4.25, outliers n = 6. T2: parent reports M = 6.83, SD = 5, outliers n = 2; child reports: M = 5.92, SD = 5.9, outliers n = 2). Children reported an average of 7.88 discrete play sessions per week (SD = 4.15) at T1 and 8.11 (SD = 4.78) at T2.Descriptive statistics for the frequency measures are presented in Table T1: r ≥ .47, p < .001; T2: r ≥ .41, p < .001). Moderate correlations were also observed within reporters across T1 and T2 . Game frequency was operationalized as child reports of hours gamingModerate correlations were observed across the three frequency measures at each time point . Experimenters clearly informed children what was meant by “cooperative” and “competitive” gaming: Children were asked to think about the times that they play video games, and to rate the frequency with which, when playing, they play a game where they have to “work together with others; that the game is cooperative” and “play against others; that the game is competitive”.Whereas third-party review boards provide information about whether a game contains violent content, the extent to which games are played cooperatively or competitively is not. Violent content is therefore objectively observable based on a game’s content and design, whereas competitive and cooperative play is more difficult to observe. Following Przybylski and Mishkin , cooperat-tests were used to determine whether children’s psychosocial health and gaming frequency changed from T1 to T2, independent t-tests were used to determine whether there were gender differences on all variables at both timepoints, and correlations were calculated. To investigate our hypotheses, three sets of structural path models were estimated with the lavaan package  externalizing problems, (H2) internalizing problems, (H3) peer problems, and (H4) overall psychosocial health. Figure T1 influencing gaming frequency. These same models were used to explore the relationship between gaming and changes in attention problems and in prosocial behavior. Our second and third models targeted those children who regularly played video games, defined as children who played for more than 1 h per week In the first models, cross-lagged panel models were estimated to test H1-H4; that is, whether gaming at oblems, H peer prooblems, H peer proT1 and T2. Peer problems and total difficulties decreased from T1 to T2  = 2.09, p = .038; total difficulties: MT1 = 7.15 SDT1 = 5.33 vs. MT2 = 6.47 SDT2 = 4.99, t(173) = 2.29, p = .023). Regarding gaming frequency, while children did not report an increase in the number of discrete sessions per week that they played video games  = −1.18, p = 0.238), children’s hours gaming per week increased according to both parents  = −4.15, p < .001) and children  = −2.61, p = .01). Table Table T1 and T2 as having more hyperactivity problems, less prosocial behavior, and more overall difficulties. Parents also reported boys as having more conduct problems at T2 than girls, but this was not observed at T1. According to parents and children, boys played video games more frequently than girls at both T1 and T2. Finally, the popularity of violent video games increased from T1 (n = 47) to T2 (n = 64), t(177) = −2.69, p = .008.Gender differences were observed at both time points for both the SDQ and gaming frequency Table . RegardiT1 was associated with changes in psychosocial health. Contrary to our expectations, (H2) gaming frequency predicted an increase in internalizing problems from T1 to T2 . Also contrary to our expectations, gaming was unrelated to (H1) externalizing problems—β4 = .092, p = .125; (H3) peer problems—β4 = .040, p = .516; and (H4) total difficulties—β4 = .039, p = .413. Regarding hyperactivity/inattention and regarding prosocial behavior, gaming frequency was not associated with changes in these variables;—β4 = −.053, p = .255, and β4 = −.022, p = .727 respectively. None of the psychosocial health measures at T1 were associated with changes in gaming from T1 to T2  thus no selection effects were observed. Stability paths for these models ranged from 0.54 to .79 with p ≤ .001 for β1(SDQ across timepoints), and from 0.3 to .31 with p ≤ .001 for β2(gaming frequency across timepoints), indicating that the SDQ and gaming frequency were relatively stable across time points, with gaming however showing more variability from T1 to T2.Figure T1 β4. In both the conduct problems and the prosocial behavior models, no associations were observed for either violent gaming  or the interaction term We next explored whether changes in conduct problems and in prosocial behavior would be associated with violent gaming among those children who played games at see Fig. . ViolentT1 5 and (1) cooperative gaming and (2) competitive gaming—β6 and β8 respectively. Neither cooperative nor competitive gaming at T1 was associated with changes in prosocial behavior ; moreover no significant interaction was observed between cooperative gaming and gaming frequency on prosocial behavior . Yet, a significant interaction was observed between competitive gaming and hours gaming . To interpret this interaction, simple slopes analyses were conducted, and the regions of significance were identified using the Johnson-Neyman Technique . Thus, hours gaming at T1 predicted less prosocial behavior at T2 only for those who both gamed more on average than their peers and who tended to play video games competitively.Finally, we simultaneously investigated whether changes in prosocial behavior would be (H5a) positively associated with cooperative gaming and (H5b) negatively associated with competitive gaming among those children who played games at see Fig. . CooperaThe goal of the present study was to provide insight into the potential influences of playing video games on children’s psychosocial development. Despite the importance of this topic, few longitudinal studies have been conducted in this field. Moreover, the little research among pre-adolescent children has predominantly focused on gaming and children’s externalizing problems. Moreover, while cooperative and competitive gaming have become a recent focus of attention, no studies have yet examined their potential influences while taking into consideration the naturalistic way they occur, that is, often in tandem. Thus, in contrast to past work, this study employed a longitudinal design, recruited pre-adolescent children, examined children’s psychosocial across multiple domains, and simultaneously explored both cooperative and competitive gaming. Contrary to our expectations that gaming would predict improvements in children’s externalizing, internalizing, peer, and overall psychosocial problems, gaming frequency was associated with increases in children’s internalizing problems, and was not associated with other changes. We also explored the potential relationship between gaming and changes in hyperactivity and inattention, and in prosocial behavior; no relationships were observed. Importantly, no selection effects were observed as well; that is, psychosocial health at the study’s first time point was not associated with changes in gaming frequency. Likewise, children’s preference for violent video games was neither associated with changes in externalizing problems nor in prosocial behavior. Finally, while neither cooperative nor competitive gaming were associated with changes in prosocial behavior, frequent competitive gaming among children who played video games for approximately eight and a half hours or more per week was associated with declines in prosocial behavior.For video games, their potential negative influence on children’s conduct is perhaps the chief concern among the public and within the scientific community. Violent video games in particular are widely seen as having a deleterious influence on children’s conduct, giving rise to aggressive behavior and discouraging prosocial behavior (Anderson et al. This outcome may have also been influenced by our operationalizing violent gaming as a dichotomous variable. This method lumped together games that were low and high in violent content. Our procedure was motivated by the young age of our sample; we expected children to have difficulties rating the intensity and realism of gaming violence. Indeed, such ratings would likely have been either uninformative or a source of bias in our sample. This is because, of the 138 games listed by children in our sample, just seven titles were rated by the Pan European Game Information board as being unsuitable for children below the age of 16. The low prevalence of highly violent gaming in our sample may have therefore made it more difficult to observe an association between violent gaming and antisocial outcomes. On the other hand, the observed outcome may offer encouragement given that children who played games with age-appropriate levels of violence did not develop anti-social tendencies when compared to their peers who played non-violent games.It is important to consider that there was a dramatic change in the prevalence of violent video game play at this study’s second wave. By this study’s second wave, the number of highly violent games listed by children tripled to 22, and the number of children listing a preference for violent video games increased by nearly 50%. One possibility for this shift is that violent content becomes increasingly of interest to children as they develop. Violent content could be an avenue for developing children to explore mature themes such as life-and-death. Another possibility is that because highly violent games are restricted to older audiences, these games may also be designed as more challenging and complex than many non-violent games. Violent games may therefore be attractive for child gamers looking for greater challenges to meet their growing abilities. Finally, children’s interest in violent games may remain constant across development, however their access to violent games may change over the course of development. A crucial factor to consider is therefore parental mediation (Nikken and Jansz Against our expectations, gaming frequency was associated with an increase in internalizing problems, such as anxiety and depressive symptoms. Notably, however, no selection effect was observed in our study, thus it was not the case that children experiencing heightened internalizing problems were more likely to play video games the following year. Internalizing problems seem like an important domain for future research particularly considering how gaming research has mostly focused on externalizing problems such as aggression. Importantly, this significant finding should also be interpreted with caution due to the small size of the observed relationship. Still, our finding is consistent with reports that excessive gaming relates to heightened levels of depressive symptoms among adolescents (Maras et al. Gaming was not associated with changes in hyperactivity and inattention. To our knowledge, this is just the second study to examine this hypothesis with a longitudinal design (see Gentile et al. Finally, habitually playing competitive video games was only associated with a decline in prosocial behavior among children who played video games competitively for about 8 or more hours per week. This was found when controlling for cooperative gaming, which often co-occurs with competitive gaming, and which has been found to promote prosocial behavior (Gentile et al. This study’s limitations are important to consider. First, as already discussed, this study examined the potential influence of video games in a broad sense. Modern video games encompass a highly diverse set of interactions (Granic et al. Video games did not seem to pose harm for most domains of children’s psychosocial development. Parents should be particularly attentive to potential increases in children’s internalizing problems as a result of video game playing. The deleterious effects of violent video game play remains a highly debated topic, with this study not lending support for the influence of violent gaming on externalizing problems or on prosocial behavior. While cooperative gaming seems unrelated to prosocial behavior in our sample, frequent gamers who also tend to play competitively may be at risk for behaving less prosocially. Finally, this field would benefit greatly from validated measures that quantify or categorize the types of social and emotional processes being activated by different games and game types, and that accurately measure the social environment of video game sessions.


Although faces are often included in the broad category of emotional visual stimuli, the affective impact of different facial expressions is not well documented. The present experiment investigated startle electromyographic responses to pictures of neutral, happy, angry, and fearful facial expressions, with a frontal face direction (directed) and at a 45° angle to the left (averted). Results showed that emotional facial expressions interact with face direction to produce startle potentiation: Greater responses were found for angry expressions, compared with fear and neutrality, with directed faces. When faces were averted, fear and neutrality produced larger responses compared with anger and happiness. These results are in line with the notion that startle is potentiated to stimuli signaling threat. That is, a forward directed angry face may signal a threat toward the observer, and a fearful face directed to the side may signal a possible threat in the environment. The Cambridge Handbook of Human Affective Neuroscience, 2012). As such, these affective stimuli have been widely used to elicit emotions in research, and most used are pictures of emotional scenes. Negative emotional scenes typically produce more intense reactions compared with neutral and positive scenes  that in humans involves a rapid closure of the eyelids. It has been well demonstrated that the amplitude of the startle eye blink varies with the emotional content of the background stimulus viewed simultaneously , and it Some studies have investigated questions related to startle responding to facial stimuli. Among these, The only startle study we are familiar with investigating the effect of gaze direction found greater responding to pictures where the model gazed directly at the observer . However, this study investigated the effect of pictures of nude females on the startle reaction in men, thereby investigating the effect of a distinct subtype of pleasant stimuli not directly comparable to the stimuli utilized in the present study. Adams and Kleck (2003) investigated behavioral data and argued that facial displays of emotions and gaze direction combines to signal approach or avoidance. They found that anger and happy expressions were more quickly recognized in combination with a direct gaze, and that fear and sad expressions were more quickly decoded with an averted gaze. Later, In an exploratory part of the experiment, we investigated the startle response to facial stimuli at early lead intervals. A few studies have investigated early lead interval responses (50–500 ms) to pictures of emotional scenes . These sIn the present study, we investigated the startle reflex during presentations of pictures displaying angry, fearful, happy, and neutral facial expressions. Faces were directed (toward the viewer) and averted (away from the viewer), and startle eliciting stimuli were presented at an early lead interval (250 ms) and at a late lead interval (3500 ms). Based on the startle literature reviewed earlier, we hypothesized that startle would be potentiated to angry faces compared with neutral and fearful at a late lead interval when faces were directed forward. In addition, based on the work of Adams and Kleck 2003, , we expeThirty people  participated in the study. All participants reported good health and did not report any hearing problems, previous serious disease, or injury. The participants were instructed not to drink caffeinated beverages or use nicotine-containing substances for 3 hr prior to the experiment. They were also told that they could withdraw from the study at any time without giving any reason. Written informed consent was obtained from all participants, who were given two lottery tickets  for their participation or course credit for an introductory psychology class.The experiment took place in an electrically and acoustically shielded chamber where the temperature was kept at about 20℃. A Bruel and Kjær 2235 Sound Level Precision Meter was used to measure the intensity of auditory stimuli. Programs for experimental control were written by the first author in Coulbourn Human Startle System HSW v. 7.500 – 00 and run on a Microsoft Windows XP based Dell PC that controlled presentation of experimental stimuli and data acquisition.Pictures of faces were selected from the Radboud Faces Database  showing four emotions: angry, fearful, happy, and neutral. Pictures of facial expressions were shown with the expresser facing the observer  and at a 45° angle to the left (averted) . Four maStartle-eliciting noise had an intensity of 95 dB (SPL), instantaneous rise time, and duration of 50 ms. The stimuli were delivered through Sennheiser HD 250 headphones. The startle-eliciting noise was presented 250 or 3500 ms after picture onset. Startle-eliciting stimulus was presented at every trial (once per picture).. Inter-electrode distance was 1.5 to 2 cm. The ground electrode was placed centrally on the forehead. The electromyography (EMG) signal was amplified by a factor of 50,000 and filtered (passing 8–1000 Hz) by a Coulbourn V75-04 bioamplifier. The signal was rectified and integrated with a Coulbourn V76-24 contour-following integrator with a 10-ms time constant, and the output was sent to the PC via a LabLinc V interface. A 12 bits A/D board was used. Sampling on each trial began 100 ms prior to onset of the startle stimulus and continued for 200 ms after onset of the stimulus. The sampling rate was 1000 Hz.Startle eyeblink electromyographic responses were recorded from the right orbicularis oculi with two sintered-pellet silver chloride AgCl miniature electrodes (4 mm diameter) filled with Microlyte electrolyte gel (Coulbourn Instruments)Participants rated the pictures for valence, arousal, and domination in the room adjacent to the experimental chamber. Participants used the mouse pointer to indicate the level of valence, arousal, and domination elicited by each picture using a visual analog scale (VAS) on a computer screen. Each participant viewed each picture for as long as they wanted before rating. The instruction read: “Please mark on the line below how you would rate the picture”; for valence, the endpoints were labeled “unpleasant” and “pleasant”; for arousal, the endpoints were labeled “relaxed” and “agitated”; and for domination, the endpoints were labeled “submissive” and “dominant.” The response range for both was from 0 to 100 mm. The program for the VAS was written in, and controlled by, MATLAB version 8.3 with Psychophysics toolbox .Figure After arrival at the laboratory, the participants sat down in a desk chair, read, and signed the Informed Consent Form. Thereafter, the participants were lead into the experimental chamber and seated in a reclining chair. The subjects were informed of the general purpose of the study and about the stimuli and procedure. They were also told that they could withdraw from the study without giving any reason at any time. The skin below the participants’ right eye was cleaned with a swab containing alcohol and pumice, and the electrodes for measurement of the startle blink EMG were attached. The headphones were attached and the experimental procedure was initiated as described in the Apparatus and Stimuli section. The door to the experimental chamber was closed during all stimulus presentation.After the startle session, the participants rated the pictures  on a computer in the room adjacent to the experimental chamber. The participant viewed each picture for as long as they wanted before rating (more details in the Apparatus and Stimuli section). After the subjective test, the experiment was over, and the participants received the lottery tickets and left.Startle blink reflexes were scored as the difference between the maximum amplitude of the EMG response in the window from 0–200 ms after noise onset, compared with the mean EMG level for the last 100 ms prior to onset of the startle eliciting noise on that trial. Startle amplitude values were T-transformed (Z-scores multiplied by 10 and add 50). Startle baseline on each trial was calculated as the mean EMG activity in the last 100 ms prior to the startle eliciting stimulus. The average for each trial includes values of zeroes for nonresponse trials. Less than 1% of startle responses was scored as nonresponses.p < .05. Significant main effects or interactions related to the hypothesis were followed-up by contrast analyses. Other main effects or interactions were followed-up by Tukey’s honest significant difference test.The design was a 2 Face Direction  by 4 Emotion  by 2 Lead intervals  within-subject design. Results were considered significant if F = 3.03, p = .092, η2 = .09, toward more positive ratings for frontal face direction. The main effect of Emotion was significant, F = 121.10, p < .001, η2 = .81. Contrast analyses revealed the following significant hierarchy from most positive to least positive: Happy > Neutral, F = 274.48, p < .001, η2 = .90, Neutral > Fearful, F = 19.29, p < .001, η2 = .40, Fearful > Angry, F = 12.03, p = .002, η2 = .29. The interaction of Face Direction by Emotion was not significant, F = 1.49, p < .22, η2 = .05.In the 2-Face Direction by 4-Emotion within-subject analysis, the main effect of Face Direction was not significant, although there was a weak trend, F < 1). The main effect of Emotion was significant, F = 30.10, p < .001. Contrast analyses revealed the following significant hierarchy from most arousal to least arousal: Fearful = Angry, Angry > Neutral, F = 69.59, p < .001, η2 = .71, Neutral = Happy. The interaction of Face Direction by Emotion was not significant (F < 1).In the 2-Face Direction by 4-Emotion within-subject analysis, the main effect of Face Direction was not significant (F < 1). The main effect of Emotion showed a trend, F = 2.64, p = .054, η2 = .08. The interaction of Face Direction by Emotion was not significant (F < 1).In the 2-Face Direction by 4-Emotion within-subject analysis, the main effect of Face Direction was not significant  = 36.75, p < .001, η2 = .56, with greater startle to long lead interval compared with short lead interval. The other main effects were not significant: Emotion: (F < 1), Face Direction: F = 2.20, p = .15, η2 = .07. The interaction of Face Direction by Emotion was significant, F = 3.86, p = .01, η2 = .12. In addition, the three-way interaction of Face Direction by Emotion by Lead interval was significant, F = 8.74, p < .001, η2 = .23  = 10.75, p = .003, η2 = .27. The same difference was found for happy faces, F = 4.65, p = .04, η2 = .14. For fearful faces, the difference between face direction was not significant, but there was a trend toward greater startle when the faces were directed to the side, F = 3.84, p = .06, η2 = .12. There was no difference between face directions for neutral expressions (F < 2).Following up the 3-way interaction, contrast analysis revealed the following differences for the long lead interval condition: Planned comparisons showed greater startle to angry facial expressions with frontal directed face compared with the same expression where the face was directed to the side, F = 6.07, p = .020, η2 = .17, and neutral, F = 4.56, p = .041, η2 = .14, facial expressions with the same face direction. Fearful facial expressions with left face direction elicited greater startle than happy, F = 4.51, p = .042, η2 = .13, and angry, F = 10.19, p = .003, η2 = .26, facial expressions with the same face direction. In addition, neutral facial expression with left face direction elicited greater startle than angry with the same direction, F = 6.23, p = .019, η2 = .18.Planned comparison contrast analysis showed greater startle to angry facial expressions with frontal face direction compared with fearful, F = 7.34, p = .011, η2 = .20. In addition, neutral facial expression elicited greater startle than fearful facial expressions when faces were directed forward, F = 6.05, p = .020, η2 = .17. With face direction to the side angry facial expression produced larger startle than fearful, F = 5.89, p = .022, η2 = .17, happy, F = 8.83, p = .006, η2 = .23, and neutral, F = 11.45, p = .002, η2 = .28, expressions.Follow-up analysis revealed fewer differences for the short lead interval condition: Planned comparison contrast analysis showed greater startle to neutral facial expression when the face was directed forward compared with the same facial expression when the face was directed to the side, In the present study, we compared the startle response to pictures of faces with a forward direction (directed) with faces directed to the side (averted). Our analyses of the long lead interval condition revealed that directed angry faces elicited greater startle than averted angry faces. Larger startle to directed faces was also evident for happy faces. For the fearful faces, there was a trend toward greater startle to the averted compared with the ones directed forward. As far as we know, this is the first study comparing startle reactions to faces directed to the side to the standard pictures of faces directed forward. These results fit well with the findings from These results could also be in line with findings from the imaging literature: Happy faces are not expected to signal threat. However, a happy face directed at you is more relevant to you than one directed to the side. As such, it could be that startle responding to facial stimuli mimics the activation of the amygdala and is in line with the “relevance detector model” , which pThe significant differences between facial expressions with face direction left also fits with both the model from Gaze direction may also have been an important factor in the present study. Indeed, pictures of faces directed to the side indirectly produced an averted gaze in comparison to facial expressions with frontal face direction and forward gaze direction. As it is impossible to disentangle the effect of face direction and gaze direction in the present study, it remains a question for further research whether face or gaze direction is the crucial factor in this regard.The results for the frontal face direction at the late lead interval revealed greater startle to angry faces, compared with fearful and neutral. These results are similar to the results of The results also revealed a significant main effect of lead interval with greater startle to the long lead interval compared with the short lead interval. This is probably caused by prepulse inhibition PPI;  reducingThe notion from In the valence data, the main effect of Emotion was significant, and contrast analyses revealed that the happy faces was rated as more positive than the neutral ones. Neutral was rated as more positive than fearful, which in turn was rated as more positive than angry. It is notable that angry faces was rated as most negative, and this could possibly explain why angry faces elicited more startle than fearful ones. However, as the startle data showed a much more complicated pattern of results, this is not likely. It is also worth noting that angry and fearful facial expressions with left face direction had a similar valence rating as their frontal-face-direction equivalents. Therefore, it is not likely that the larger startle to facial expressions with frontal face direction could be explained by valence.Analysis of the arousal data revealed a main effect of emotion. Follow-up analyses revealed more arousal to angry and fearful facial expressions compared with neutral and happy. It is not common for happy facial expressions to be equal in arousal to neutral expressions. However, it is possible that the Norwegian words used in this study describing the end points on the VAS for arousal were not identical in meaning to the English counterparts, which may explain this atypical result. There is no good Norwegian word for arousal and the word used is more akin to “activation.”As described in the previous two paragraphs, taking face direction into account, the effect of the emotional facial expressions on startle and the subjective evaluations of the same expressions diverged. That is, startle was greater to angry facial expressions with frontal directed face compared with angry facial expression with the face directed to the side. In the subjective data , there was no such difference. The same was found for happy faces. For fearful faces, there was a trend toward greater startle when the faces were directed to the side. Such a trend was not apparent in the subjective data. A similar difference between startle and subjective evaluations of valence, arousal, and dominance was reported by To conclude, the present study showed that pictures of angry facial expressions directed forward elicited greater startle responses than pictures of neutral and fearful expressions. This is in line with previous research reporting greater startle to pictures of angry faces. However, in the present study, pictures of emotional faces were also presented at an angle from the side. Responses to these pictures were different, with greater startle to fearful faces compared with angry and happy. In sum, these results lend support to the notion that startle responses to emotional faces follow the same pattern as behavioral data from


Early haemoperfusion (HP) therapy has been found to be very effective in acute paraquat (PQ) poisoning, but the effective rescue window is still uncertain. Demographic data and the type of therapies administered of 621 patients were included as confounding factors in this retrospective study. After receiver operating characteristic curve analysis and intra-group/subgroup analysis, the initiation of glucocorticoid therapy within 3 hrs of exposure with a second treatment given <21 hrs after exposure, HP initiated within 4 hrs of exposure with a second treatment given <20 hrs after exposure, the appearance of pulmonary lesions ≤8 days after exposure and six other variables were used in a multiple analysis. The strength of positivity of the PQ urine test on admission, HP initiated within 4 hrs of exposure with a second treatment given <20 hrs after exposure, the appearance of pulmonary lesions ≤8 days after exposure, and stage III AKI on admission were independent factors of survival probability. HP therapy for acute PQ poisoning was the main therapeutic intervention investigated in this study. Outcomes were time dependent, and the crucial factor was the initiation of therapy within 4 hrs of PQ poisoning followed by a second treatment within 20 hrs. Reliable prognostic factors would be helpful in guiding treatment, and early prediction of inevitable death would be important in avoiding inappropriate treatment in patients with acute PQ poisoning6.Paraquat  is a widely used herbicide in some Asian countries due to its high efficiency and relatively low cost. However, accidental PQ poisoning is a serious health problem associated with a mortality rate of 60–70%7. In particular, the PQ concentration in body fluid will reach its maximum level in the first 4–5 hrs8. However, reliance on mortality predictions of almost 20% is unlikely to be the optimal method5. PQ intoxication thus frequently causes death due to respiratory and kidney failure10. Early initiation of treatment is the most important factor in survival, and renal protection is the cornerstone of treatment11. The reduction rate of PQ concentration by haemoperfusion (HP) is 67–83% in three hours12, so HP is very effective in acute poisoning rescue13.Survival in cases of PQ poisoning have been described as dose and time dependentAs a provincial centre for PQ intoxication treatment, our facility has been treating approximately 300 cases annually. Patients are sent to our centre directly or transferred from other hospitals, with or without first aid after PQ poisoning. We assumed that the effect of HP was dependent on the rescue window and that the window was likely influenced by many factors. In this article, we performed a retrospective study to review patient survival conditions six months after PQ poisoning and to investigate our assumptions about the therapeutic window for HP therapy.There were 705 patients initially reviewed. After further screening, 84 patients were excluded: 24 were younger than 18 years of age. 23 generated negative results from serial PQ semiquantitative urine testing conducted three times on three different days, 8 were discharged within 24 hrs without any further treatment, 9 could not estimate the exact time of poisoning, and 20 were ‘lost’ and could not be followed up.Finally, a total of 621 patients were used for analysis; 327 (52.66%) of these survived 6 months after PQ poisoning. The mean age of the full group was 37.05 ± 13.27 years. Of the patients, 298 were males, and 323 were females. One hundred seventy-three (27.86%) patients were directly taken to our institute (group A), 426 (68.60%) were transferred to us after first aid (group B), and 22 (3.54%) were transferred from more than one hospital (group C). One hundred forty-two (32.47%) patients had a medical history, and 63 (14.41%) took long-term medication. Demographic data and univariate analyses comparing patient survival and patient death are shown in Table χ2 = 2.736, P = 0.036; positive results at admission and first repeat test: χ2 = 3.108, P = 0.031; all three positive results: χ2 = 3.452, P = 0.022), the arrival type , a medical history of chronic obstructive pulmonary disease (COPD) or asthma , a medication history positive for immunosuppressive drugs , serum creatinine on admission  and acute kidney injury (AKI) stage I, II or III . Glucocorticoids, immunosuppressors and HP were used a median of 7, 8 and 10 times, respectively. No significant difference was found between the survival rate and the number of repeated treatments . Serum creatinine tests were conducted 8 times/10 days (range 7–16 days). It was first tested on admission, and testing was repeated after each HP treatment. There was a significant difference between serum creatinine levels on admission and after HP , but no significant difference was found after glucocorticoid or immunosuppression therapy. In addition, no significant difference in survival rate was found between age, gender, and other demographic data.There were significant differences in the survival rate based on the number of positive PQ semiquantitative urine tests . The mean time to semiquantitative urine test after initial PQ exposure was 5.84 ± 2.37 hrs for the first test, 10.97 ± 3.45 hrs for the second test and 17.24 ± 4.06 hrs for the third test. Gastric lavage, cathartics, and emetics were used only once; the mean time to each treatment was 2.46 ± 2.18 hrs, 3.28 ± 3.02 hrs and 1.54 ± 1.17 hrs, respectively. A significant difference was found in the first appearance of AKI stage I (P = 0.036), the first appearance of pulmonary lesions (P = 0.024), the first use of glucocorticoids (P = 0.043), and the first and second uses of HP (P = 0.033) (Table P = 0.106), cathartic administration (P = 0.173), or emetic administration (P = 0.152). Additionally, no significant correlation was found between survival rate and the time to gastric lavage , cathartic administration  or emetic administration .The time after PQ poisoning was considered in further analysis. Patients reached our institute at a mean time of 5.84 ± 2.37 hrs. The mean time of groups A, B and C were found to have a significant difference . Therefore, the variables were expressed in terms of hours  and used as dummy variables  but not in the >3 hrs subgroup . In addition, a significant difference after the second HP treatment was found in the <4 hrs subgroup  and in the ≥4 hrs subgroup . No significant difference was found in the time of the second immunosuppressor treatment . After ROC curve analysis, the cut-off for the second glucocorticoid treatment was 20.82 hrs (rounded to 21 hrs) in the ≤3 hrs subgroup, and the cut-off for the second HP treatment was 19.87 hrs (rounded to 20 hrs) in the <4 hrs subgroup and 23.26 hrs (rounded to 23 hrs) in the ≥4 hrs group. The glucocorticoid variable of was divided into three groups: (i) first treatment ≤3 hrs, second treatment <21 hrs, (ii) first treatment ≤3 hrs, second treatment ≥21 hrs and (iii) first treatment >3 hrs . The HP variable was divided into four groups: (i) first treatment <4 hrs with second treatment <20 hrs, (ii) first treatment <4 hrs with second treatment ≥20 hrs, (iii) first treatment ≥4 hrs with second treatment <23 hrs and (iv) first treatment ≥4 hrs with second treatment ≥23 hrs . No significant difference between patient survival and patient death was found in either of these subgroups or in other time periods of glucocorticoid, immunosuppressor or HP treatment .Glucocorticoid therapy, immunosuppression therapy and HP were provided approximately once daily; glucocorticoids were the only therapy given before transfer to our institute. No significant difference was found between survival rate and repeated therapy administrations (number of glucocorticoid treatments: P = 0.023), semiquantitative urine PQ test level of +++ (P = 0.001), first HP <4 hrs with second HP <20 hrs (P = 0.003), and first appearance of pulmonary lesions ≤8 days (P = 0.047)  and positive PQ urine semiquantitative test. The exclusion criteria were age less than 18 years of age, lost to follow-up, suspected cases ruled out , unknown time of exposure, and discharge within 24 hrs without any further treatment. The patient flowchart and the study protocol are shown in Fig. 19. Pulmonary lesions were diagnosed by characteristic effusion and fibrosis in computed tomography (CT) images . The number of times each therapy was repeated, the frequency of outcome measurements, and the rescue window after PQ poisoning were assessed in hours.All dependent variables and subgroups were determined by univariate analysis. Subgroups contained the therapies provided each hour and the number times each therapy was repeated. The data analysis included arrival time (directly arrived or transferred), age, gender, medical history , medication history longer than 6 months , semiquantitative urine dithionite PQ test results , gastrointestinal decontamination , use of glucocorticoids and immunosuppressive drugs, HP (average reinfusion rate was 9 L/session and 300 mL/min in the postdilution mode), serum creatinine , and pulmonary lesions. Changes in serum creatinine level categorized AKI into three stages t test or a Wilcoxon rank-sum test was performed on numerical data among groups. The chi-square test was used for categorical data. Pearson correlation analysis or curve estimation was used to estimate correlations between two variables. Repeated-measures analysis of variance was used to estimate the relationship between repeated therapies and the tests. ROC curve analysis was used to evaluate the threshold value of numerical variables, which were then transformed into binary variables. A one-way analysis of variance or Kruskal-Wallis test was used to compare multiple subgroups; otherwise, subgroups were divided into dummy variables and analysed by univariate analysis. The variables with P values less than or equal to 0.10 were used in the multiple-factor analysis. Multiple logistic regression analysis was used to evaluate the relationship between dependent variables and survival probability. The level of statistical significance was set at P < 0.05. All analyses were performed using SPSS software .Student’s This study was approved by the Ethical Committee of the First Affiliated Hospital, School of Medicine, Zhejiang University. We obtained the verbal consent of all patients or their relatives by phone calls.


In addition, these concentrations of biguanides also trigger burst of reactive oxygen species production which, in combination with pleiotropic OXPHOS inhibition, can be toxic for the organism. We conclude that the beneficial effect of biguanides should probably be associated with subtler mechanism, different from the generalized inhibition of the respiratory chain.Metformin is widely prescribed as a first-choice antihyperglycemic drug for treatment of type 2 diabetes mellitus, and recent epidemiological studies showed its utility also in cancer therapy. Although it is in use since the 1970s, its molecular target, either for antihyperglycemic or antineoplastic action, remains elusive. However, the body of the research on metformin effect oscillates around mitochondrial metabolism, including the function of oxidative phosphorylation (OXPHOS) apparatus. In this study, we focused on direct inhibitory mechanism of biguanides (metformin and phenformin) on OXPHOS complexes and its functional impact, using the model of isolated brown adipose tissue mitochondria. We demonstrate that biguanides nonspecifically target the activities of all respiratory chain dehydrogenases , but only at very high concentrations (10 Metformin (dimethyl biguanide) is the most widely used frontline drug for treatment of type II diabetes mellitus , 2. At tMetformin utility was also explored in the model of heart failure where epidemiologic evidence suggests its protective effect , 13. NevIn addition to its antihyperglycemic effect, a broad range of epidemiologic studies showed that chronic metformin treatment is associated with a reduced risk of cancer –18. As wF) was identified as a site of superoxide production using NADH-linked substrates [Q) towards the molecular oxygen [F) is not fully occupied by the substrate and is more accessible to oxygen allowing superoxide generation [Respiratory chain is also a predominant source of reactive oxygen species (ROS), and targeting individual complexes may modulate ROS production and thus influence various pathological processes and/or signaling in different metabolic pathways . Using sbstrates , 33. Undr oxygen , 31. SDHneration . Mitochoneration  even in neration . ROS proneration . Since mneration .ψm) in the isolated BAT mitochondria so that the cationic biguanides are much less concentrated in mitochondrial matrix, and therefore, the determined concentrations of these drugs are very close to the actual ones required for inhibition. We compared the effects of biguanides on dehydrogenases with their typical inhibitors  using activity assays, mitochondrial oxygen consumption, and ROS production.Given the conflicting reports regarding the molecular target of biguanides in mitochondria, in this study, we addressed the direct impact of biguanides metformin and phenformin on NDH, SDH, and mGPDH, mitochondrial glycerophosphate dehydrogenases, which feed electrons into respiratory chain and were proposed to be a target of these drugs. We used mitochondria isolated from brown adipose tissue (BAT), which are advantageous for such study as they contain comparable amounts of these dehydrogenases. Furthermore, uncoupling protein 1 (UCP1) dissipates mitochondrial membrane potential (ΔWe used interscapular brown adipose tissue (BAT) of four-week-old Wistar rats kept at 22°C under 12 h/12 h light/dark cycle on a standard diet and water supply ad libitum. All animal works were approved by the institutional ethics committee and were in accordance with the EU Directive 2010/63/EU for animal experiments. Mitochondria were isolated in STE medium  and supplemented with BSA (10 mg/mL) by differential centrifugation . Qualityε340 = 6.22 mM−1·cm−1) or CoQ1 oxidoreductases . The assay medium contained 50 mM KCl, 20 mM Tris–HCl, 1 mM EDTA, 1 mg/mL BSA, 2 mM KCN, pH 7.4, and 50 μM CoQ1. The reaction was started by adding 100 μM NADH, 25 mM glycerophosphate (GP), or 25 mM succinate, respectively, and after 5–10 minutes, changes of absorbance were monitored at 30°C. Enzyme activities were expressed as pmol/s/mg protein.Activities of mitochondrial dehydrogenases were determined spectrophotometrically as NADH oxidoreductases  were measured with TPP+-selective electrode as described in [μM carbonyl cyanide-p-trifluoromethoxyphenylhydrazone (FCCP). The electrode was calibrated by stepwise addition of TPP+ (1–6 μM) before each measurement, and Δψm changes were plotted as pTPP, that is, the negative common logarithm of TPP+ concentration.The changes in mitochondrial membrane potential  supplemented either with 10 mM pyruvate plus 2 mM malate, 10 mM succinate, or 10 mM GP. Amplex UltraRed was used at the final concentration of 50 μM with horseradish peroxidase (HRP) at 1 U/mL. Fluorescence signal from the well containing all substrates and inhibitors, but not mitochondria, was subtracted as a background for every experimental condition used. Background, caused mostly by autoxidation of the dye or nonenzymatic effect of inhibitors on apparent ROS production, varied between conditions from 0 to 1 pmol/s/mg and was uniform across respective titration points. Signal was calibrated using H2O2 at the final concentration of 0–5 μM, and H2O2 stock concentration was routinely checked by measuring its absorption at 240 nm.Hydrogen peroxide production was determined fluorometrically by measuring oxidation of Amplex UltraRed (Thermo Fisher) essentially as before . Fluoresv/v) 2-mercaptoethanol, 4% (w/v) SDS, 50 mM Tris–HCl, pH 7.0, 10% (v/v) glycerol, and 0.017% (w/v) Coomassie Brilliant Blue R-250), and Tricine SDS-PAGE [w/v) polyacrylamide slab gels at room temperature. The gels were blotted onto a PVDF membrane  by semidry electrotransfer at 0.8 mA/cm2 for 1 h. Membranes were blocked in 5% nonfat dried milk dissolved in TBS  for 1 h at room temperature and incubated for 2 h with the following primary antibodies: antibody to NADH dehydrogenase (NDUFA9 subunit of complex I)—Abcam ab14713, succinate dehydrogenase complex (subunit A (SDHA) of complex II)—Abcam ab14715, Core2 subunit of complex III—Abcam ab14745, actin—Millipore MAB1501, rabbit polyclonal antibody to porin (1 : 1000) was a kind gift from Professor Vito de Pinto , and rabbit polyclonal antibody to mGPDH was custom prepared [BAT and liver homogenates were denatured at 65°C for 15 min in a sample lysis buffer  using 15–60 μg protein/mL of freshly isolated mitochondria. For measurements, 10 mM pyruvate plus 2 mM malate, 10 mM succinate, or 10 mM GP, respectively, were used. The oxygen consumption was expressed in pmol oxygen/s/mg protein.Oxygen consumption was measured at 30°C as described before  using OxThe direct mechanistic knowledge on how biguanides  influence mitochondrial function is not yet clear. While their inhibitory effect on NADH dehydrogenase  received most attention , 23, 24,ψm determination on TPP+-selective electrode , as they have several advantages for such study. First, compared to the liver, where the inhibitory action of biguanides on mGPDH was originally described , they colectrode . This meguanides  and thus50 varied from 80 to 180 mM, Figures 50 varied between 10 and 25 mM for both isolated enzyme activities and respiration of all three dehydrogenases  were the same or slightly higher than corresponding IC50 values in frozen-thawed mitochondria (spectrophotometry), thus conferring comparable concentration of biguanides on both sides of the inner mitochondrial membrane. Determined IC50 in our case is rather high in comparison with that previously reported [As a first step, we estimated direct impact of metformin on activity of each complex measured as individual dehydrogenase activity (substrate : coenzyme Q) in frozen-thawed mitochondria or in the context of electron transport chain in intact mitochondria measured as oxygen consumption using respective substrates . Metformogenases . Becausereported , 33. ThiμM of iGP-1. Interestingly, we also observed some nonspecific actions of iGP-1 on SDH  than that in the case of mGPDH  and activity of mGPDH was abolished at 50  Figures , but the Figures .50 we identified for phenformin and metformin, it is to be questioned whether their effects on respiratory chain enzymes are pharmacologically relevant and can affect mitochondrial function in tumors. As already mentioned, isolated BAT mitochondria with inherently low levels of mitochondrial membrane potential represent a model free of the effect of metformin partitioning due to its charge. Biguanides carry a positive charge, which means that they do accumulate in cells and mitochondria in dependence on membrane potential across both cellular and mitochondrial membranes. Thus, the concentration could be 100 to 300 times higher inside energized mitochondria (compared to concentration in cytosol) with the difference in mitochondrial membrane potential between 120 and 150 mV [μM range [Given the extremely high ICd 150 mV . Since tμM range , the actμM range , 47. HowμM range . One of μM range  reportedAction of pharmacologically active compounds on suppression of cancer cell proliferation may not only be achieved by decreased flow of electrons through respiratory chain and associated impairment of aerobic ATP production. Another successful antiproliferative strategy may be to increase in reactive oxygen species (ROS) production and subsequent induction of apoptosis , 50. It 2O2 upon conversion by superoxide dismutase) in fresh mitochondria from BAT, where the respiratory chain is not reduced due to uncoupling by UCP1. First, using the established inhibitors of individual dehydrogenases and their respective substrates, we established ROS production pattern characteristic for each site in BAT mitochondria , indicative of ROS production from IF site , very low levels of ROS were produced in uncoupled state, which is in agreement with the previously published data  and thise of NDH . Low leve of NDH .GP-dependent ROS production was quite high even when respiratory chain was not reduced and did not increase further after GDP addition, again in agreement with previously reported data , 58, andCompared to typical inhibitors, both metformin and phenformin induced dose-dependent increase in ROS production, but it seemed rather indiscriminative, as its pattern was highly similar for all substrates tested Figures . It sugg50 higher than 80 mM in case of metformin or ranging from 20 to 30 mM in case of the more potent phenformin. Moreover, these biguanide concentrations induce nonspecific increase of mitochondrial reactive oxygen species production. Our data suggest that biguanides do not bind to any specific sites on respiratory chain dehydrogenases and require high concentrations to be effective. Under these conditions, their effect on dehydrogenases remains therapeutically questionable. Indeed, phenformin use had to be discontinued due to induction of lactic acidosis in treated patients. As summarized in Our data demonstrate that the inhibitory effect of biguanides on OXPHOS enzymes is rather pleiotropic including the previously reported inhibition of NADH and mGPDH dehydrogenases. Drug sensitivity of respiratory chain complexes in brown adipose tissue was comparable, with IC


Nature Communications6: Article number: 6793; DOI: 10.1038/ncomms7793 (2015); Published: 04092015; Updated: 03132018In the originally published version of this Article, there were errors in the Methods section and in Supplementary Figure 2. In the Methods section entitled ‘Quality control’, and in Supplementary Figure 2, references to ‘MAF>0.0001’ were incorrectly given as ‘MAF>0.01’. These errors have now been corrected in both the PDF and HTML versions of the Article and in the Supplementary Information File.