Jane E Ferrie
email: j.ferrie@public-health.ucl.ac.uk

Mika Kivimäki

Archana Singh-Manoux

Alison Shortt

Pekka Martikainen

Jenny Head

Michael Marmot

David Gimeno

Roberto De Vogli

Marko Elovainio

Martin J Shipley

Non-response to baseline, non-response to follow-up and mortality in the Whitehall II cohort Running head: Non-response and mortality

Keywords: non-response to follow-up, partial response, socioeconomic inequalities, all-cause mortality, occupational cohort, white-collar

Background: Little is known about the associations between non-response to follow-up surveys and mortality, or differences in these associations by socioeconomic position in studies with repeat data collections.

), and all-cause mortality to 2006. Odd-numbered phases included a medical examination in addition to a questionnaire.

Results: Non-response to baseline and to follow-up phases that included a medical examination was associated with a doubling of the mortality hazard in analyses adjusted for age and sex. Compared with complete responders, responders who missed one or more phases, but completed the last possible phase before they died, had a 38% excess risk of mortality. However, those who missed one or more phases including the last possible phase before death had an excess risk of 127%. There was no evidence that these associations differed by socioeconomic position.

Conclusion:

In studies with repeat data collections non-response to follow-up is associated with the same doubling of the mortality risk as non-response to baseline; an association that is not modified by socioeconomic position.

Most observational epidemiology studies rely on participation of the target population to obtain data on risk exposures and health outcomes. If non-response at baseline is non-random, findings from such studies may be biased. In prospective studies with repeat data collections the problem of bias introduced by non-response to the baseline survey may be compounded by non-response to follow-up surveys.

Many studies have compared mortality risks between responders and non-responders to baseline, with most, [START_REF] Dawber | An approach to longitudinal studies in a community: the Framingham Study[END_REF][START_REF] Doll | Mortality in relation to smoking: Ten years' observations of British Doctors[END_REF][START_REF] Wilhelmsen | A comparison between participants and nonparticipants in a primary preventive trial[END_REF][START_REF] Heilbrun | The effects of nonresponse in a prospective study of cancer[END_REF][START_REF] Walker | Non-participation and mortality in a prospective study of cardiovascular disease[END_REF][START_REF] Benfante | Response bias in the Honolulu Heart Program[END_REF][START_REF] Heilbrun | The effects of non-response in a prospective study of cancer: 15-year follow-up[END_REF][START_REF] Strandberg | Mortality in participants and non-participants of a multifactorial prevention study of cardiovascular diseases: a 28 year follow up of the Helsinki Businessmen Study[END_REF][START_REF] Berglund | Long-term outcome of the Malmo preventive project: mortality and cardiovascular morbidity[END_REF][START_REF] Hara | Comparison of cause-specific mortality between respondents and nonrespondents in a population-based prospective study: ten-year follow-up of JPHC Study Cohort I. Japan Public Health Center[END_REF][START_REF] Barchielli | Nine-year follow-up of a survey on smoking habits in Florence (Italy): higher mortality among non-responders[END_REF][START_REF] Jousilahti | Total and cause specific mortality among participants and non-participants of population based health surveys: a comprehensive follow up of 54 372 Finnish men and women[END_REF][START_REF] Mattila | Adolescent survey non-response and later risk of death. A prospective cohort study of 78,609 persons with 11-year follow-up[END_REF][START_REF] Harald | Non-participation and mortality in different socioeconomic groups: the FINRISK population surveys in 1972-92[END_REF] but not all, [START_REF] Paul | A longitudinal study of coronary heart disease[END_REF] providing strong evidence of higher rates of mortality among non-responders. However, it is considered that observational epidemiologists have paid insufficient attention to response bias in cohort studies. [START_REF] Greenland | Response and follow-up bias in cohort studies[END_REF] Unresolved issues include associations between partial response or non-response to follow-up and mortality in studies with successive data collections, and differences in these associations by socioeconomic position. Socioeconomic position is important for two reasons: one, as a confounder in aetiological studies and two, as an exposure in studies of social inequalities in health. [START_REF] Krieger | Why epidemiologists cannot afford to ignore poverty[END_REF] If the association between non-response and mortality were to differ across the socioeconomic strata residual confounding might remain a problem for aetiological studies, even after adjustment for socioeconomic position, and analyses of socioeconomic inequalities might be biased. This paper uses data from the Whitehall II study which was set up to explain socioeconomic gradients in health. 18;19 The paper has two main aims: 1) to examine associations between non-response or partial response to repeat surveys subsequent to baseline and mortality; and 2) to examine associations between non-response and mortality by socioeconomic position.

METHODS

The target population for Whitehall II was all London-based office staff aged 35-55 working in 20 civil service departments in 1985 (n=14,121). Baseline screening (Phase 1, 1985-1988) involved a medical examination and a self-administered questionnaire. Subsequent to baseline, odd-numbered data collection phases have included both a medical examination and a questionnaire, while even-numbered phases have been questionnaire only (Box 1).

Response status: Phase 1: Of the target population 10,308 people (3,413 women and 6,895 men) were responders to baseline. Invitations to participate in follow-up phases were extended only to these baseline responders. Phases 2 and 4: Responders at these Phases completed a full questionnaire. From Phase 4, response was augmented in the last resort through a short-form telephone interview consisting of selected questions from the questionnaire; such participants were categorized as partial (telephone) responders. Phases 3 and 5: Responders at these Phases attended the medical examination; partial (questionnaire) responders completed a questionnaire, but did not attend the medical examination; and partial (telephone) responders completed the short-form telephone interview only. We created a single, time dependent, response status summary measure using all available data from Phases 1 to 5. The measure has four categories which indicate the cumulative response status at each of the five phases.

Response status summary measure: (1) completed all possible phases (all five phases, or all phases prior to death if this occurred before Phase 5): (2) responders to Phase 1 who subsequently were nonresponders at one or more phases, but responded at their last possible phase i.e. the last phase they could have attended up to that point in time: (3) responders to Phase 1 who subsequently were nonresponders at one or more phases, including their last possible phase; and (4) baseline (Phase 1) nonresponders.

All-cause mortality: Mortality data were obtained from the National Health Services Central Registry. The analyses of non-response to follow-up and partial-response to follow-up include the 10,308 responders to baseline from all the 20 departments that make up the Whitehall II study. Of these 99.9% (10,297) have been followed for mortality. Mortality data were available for non-responders to baseline in 14 of the 20 departments that make up the Whitehall II study. Effective tracing of non-responders to baseline in the 6 remaining departments was impossible due to insufficient information.

Covariates: Information on age, sex and employment grade at baseline was derived from the Phase 1 questionnaire for responders and from Civil Service department files for non-responders. Civil Service employment grade was classified into high (senior administrators), intermediate (executives, professionals and technical staff) and low (clerical and office support staff). As of August 1992 the salary range among high grade employees was £25 330 -£87 620 (Euro 32 583 -112 680, US$ 44 428 -153 622) and among low grade employees £7 387 -£11 917 (Euro 9 498 -15 322, US$ 12 949 -20 895). 18;19 Statistical methods Associations between response status and mortality are presented as mortality rates per 1,000 person years, or hazard ratios (HRs) and 95% CIs determined using Cox proportional hazard models with followup as the time-scale and responders as the reference category. Person years of follow up were computed as follows: for responders, the start of follow-up (Phase 1) was their date of screening while for subsequent phases it was the date of first participation in that phase. For non-responders, the start of follow-up (Phase 1) was taken as the date that screening began for those who worked in the same building, while for subsequent phases the midpoint date for the responders was used. Participants were followed for mortality up to 31 st July 2006 or their date of death (8%), embarkation (1%) or deregistration with a health authority (1%). For all analyses that included the non-responders at Phase 1, an indicator variable was included in the models to distinguish and adjust for any mortality differences between responders in the 14 departments that provided mortality data for the non-responders and the 6 departments that did not. As there was no strong evidence of sex differences, pooled estimates are presented. The cumulative effect of non-response across phases was assessed using the response status summary measure updated at each phase and fitted as a time dependent variable. The estimates are presented as hazard ratios (HR) and 95% CIs for the total cohort and separately by employment grade category.

RESULTS

The age and sex-adjusted, all-cause mortality hazard for Phase 1 non-responders is double that for Phase 1 responders (Table 1). Non-response to follow-up at Phases 3 and 5 is associated with an increased hazard similar to that associated with Phase 1 non-response, but non-responders to Phases 2 and 4 (questionnaire-only phases) have lower HRs. Partial response: There was no evidence that the HR for mortality among partial (questionnaire) responders at Phase 3 (medical screening phase) was different from that for Phase 3 non-responders.

At Phase 4 (questionnaire only) there was no evidence that the HR for mortality among partial (telephone) responders, HR (95% CIs) 1.15 (0.79-1.67), was different to that for Phase 4 responders.

At Phase 5 (medical screening phase) the HR for mortality among partial (telephone) responders (2.08, 1.45-2.98), did not differ from that for partial (questionnaire) responders (1.89, 1.38-2.58), and both were no different to that for Phase 5 non-responders (2.23, 1.83-2.71). Adjustment of the HRs for nonresponders and partial responders for employment grade gave estimates only marginally below those adjusted for age and sex.

Non-response: Table 2 presents associations between the response status summary measure and mortality. Due to similar HRs for mortality between partial responders and non-responders at medical examination phases, and between partial (telephone) responders and responders at questionnaire phases, partial responders and non-responders are combined at Phases 3 and 5, while partial responders and responders are combined at Phase 4. Compared to participants who completed all possible follow-up surveys, the HR for mortality among non-responders at Phase 1 was 2.77 (2.32-3.31) -Table 2, upper section. The HR was 2.27 (1.94-2.67) among those who were non-responders at one or more subsequent phases and non-responders at their last possible phase, but 1.38 (1.09-1.74) among those who were non-responders at subsequent surveys but responders at their last possible phase. Adjustment for employment grade had little effect on these findings and analyses of associations between the response status summary measure and mortality within each of the three employment grade categories provided no evidence that associations differ by socioeconomic position -Table 2, lower section.

Non-response to last survey: As indicated above, missing one or more follow-up surveys was associated with an excess mortality risk of only 38% (p=0.007) provided the last possible survey had been completed. To quantify the additional mortality risk associated with non-response to the last possible survey we undertook further analyses in which responders who missed one or more phases including the last possible phase were compared with those who either completed all their possible phases or who missed one or more phases but completed their last possible phase. This comparison showed that the additional mortality risk associated with missing the last possible phase was over 100%, HR 2.02 (1.73 -2.36), p<0.001.

DISCUSSION

The key finding of this paper is that non-response to clinical follow-up was associated with the same doubling of the mortality risk as non-response to baseline. Non-response to one or more follow-up surveys was associated with a 1.4-fold excess risk among those who responded to their last possible survey, but a 2.3-fold excess risk among those who missed their last possible survey. We also found that excess mortality risk is negligible among partial responders to questionnaire only surveys, but similar to that for non-responders for participants who miss a medical examination but respond partially through questionnaire completion. In addition, the study examined non-response by socioeconomic position and found no evidence that socioeconomic position modifies the association between non-response and mortality. Our findings that non-responders to baseline had a mortality hazard double that for responders is in line with findings from previous studies. 2;5;7-14;20-26 Although a number of studies have reported somewhat higher levels of non-response in low socioeconomic groups, 5;6;23;27-33 few studies have examined whether the effects of non-response differ across the socioeconomic strata. 29;30;33 In the present study, mortality risks among non-responders were little attenuated by adjustment for socioeconomic position measured by employment grade. Similarly, analyses within grade provided no evidence of differences in the associations between non-response and mortality. Recent findings from the population-based FINRISK surveys similarly observed little attenuation of the association between non-response to baseline and mortality on adjustment for socioeconomic position, and a two-fold higher mortality risk in non-responders to baseline in every socioeconomic category. [START_REF] Harald | Non-participation and mortality in different socioeconomic groups: the FINRISK population surveys in 1972-92[END_REF] Studies that have examined morbidity have similarly provided little evidence that the bias introduced by non-response to baseline differs across socioeconomic strata. 29;33 It is implausible that there is a direct association between non-response and mortality, rather it is an association driven by a variety of common causes, for example ill-health, caring and accessibility. Some of these, such as measureable ill-health at baseline, were available for participants in the Whitehall II study, but not for non-responders to baseline, and many others were not measured. Consequently we have chosen to present our analyses adjusted only for age, sex and grade, which were available for everyone included in the analyses. Participants in the Whitehall II study were white-collar civil servants at study entry and covered a wide range of grades with a salary difference of over 10-fold between the top and bottom of the hierarchy. In this respect they are representative of the majority of employees in Western countries. Nonetheless, this work needs to be repeated in general population samples as the generalisability of our findings will be limited by the healthy worker effect at baseline which may produce an underestimation of associations between non-response and mortality.

Our findings from a study with repeat data collections show the excess mortality associated with nonresponse to previous follow-up surveys is approximately 1.4-fold if the most recent possible survey is completed, but over 2-fold if it is missed. When collecting questionnaire data only, the health-related selection introduced by the inclusion of partial responders who only complete a short telephone interview is similar to that for those who complete a full questionnaire. This means their inclusion in analyses, while increasing the power of the study, will not reduce the bias associated with complete non-response to follow-up at that phase. However, when data collection includes both a medical and a questionnaire, partial responders who miss the medical examination are similar in health status to those who are complete non-responders. Inclusion in analyses of questionnaire or telephone interview data for these partial responders would thus both increase power and reduce the health-selection bias associated with non-response to follow-up. Our findings regarding partial responders are plausible. Attendance at a medical examination requires much more time and effort than completing a questionnaire or short-form telephone interview at home. Thus, non-response to surveys that include a medical examination outside the home is likely to capture a wider range of the unmeasured common causes of non-response and mortality. Additionally, any association between non-response and mortality may be strengthened further if non-responders miss a diagnosis or treatment that might have been brought to attention by the medical examination.

With respect to studies designed to examine inequalities in health good news is provided by evidence that socioeconomic position does not modify the association between non-response and mortality.

Ethical approval: Ethical approval for the Whitehall II study was obtained from the University College London Medical School committee on the ethics of human research.

Key messages:

In common with non-response to baseline, mortality among non-responders to clinical follow-up is double that of responders. Non-response to follow-up surveys prior to the most recent survey is associated with a lower excess mortality risk.

Excess mortality risk is negligible among partial responders to questionnaire only surveys, but similar to that for non-responders for participants who miss a medical examination (but have partial response via questionnaire). Socioeconomic position does not modify the association between non-response and mortality. * maximum response rates -include questionnaire only and short-form telephone interviews

  Sources of financial support:The Whitehall II study has been supported by grants from the Medical Research Council; British Heart Foundation; Health and Safety Executive; Department of Health; National Heart Lung and Blood Institute (HL36310), US, NIH: National Institute on Aging (AG13196), US, NIH;

Acknowledgements: We thank all participating Civil Service departments and their welfare, personnel, and establishment officers; the Occupational Health and Safety Agency; the Council of Civil Service Unions; all participating civil servants; and all members of the Whitehall II study team. In addition we would like to acknowledge the very constructive contributions of the reviewer.


Ph.D Markus Jokela
email: markus.jokela@helsinki.fi

Ph.D Marko Elovainio

Ph.D Archana Singh-Manoux

Ph.D Mika Kivimäki

Longitudinal Survey of Youth

Keywords: Cognitive ability, Cognitive epidemiology, Intelligence, Mortality, NLSY IQ=intelligence quotient, SES=socioeconomic status, OR=odds ratio, CI=confidence interval, RR=relative risk

General cognitive ability, assessed by cognitive tests and often referred to as intelligence quotient (IQ), reflects a general capacity of abstract reasoning and learning potential [START_REF] Deary | The impact of childhood intelligence on later life: following up the Scottish mental surveys of 1932 and 1947[END_REF][START_REF] Plomin | Genetics and general cognitive ability (g)[END_REF][START_REF] Gottfredson | Why g matters: The complexity of everyday life[END_REF].

An increasing number of studies have shown that high IQ in childhood or early adulthood predicts low mortality risk over the life course [START_REF] Hart | Childhood IQ, social class, deprivation, and their relationships with mortality and morbidity risk in later life: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF][START_REF] Whalley | Longitudinal cohort study of childhood IQ and survival up to age 76[END_REF][START_REF] Pearce | Childhood IQ and deaths up to middle age: The Newcastle Thousand Families Study[END_REF][START_REF] Batty | IQ in early adulthood and later cancer risk: cohort study of one million Swedish men[END_REF][START_REF] Hemmingsson | The association between cognitive ability measured at ages 18-20 and coronary heart disease in middle age among men: A prospective study using the Swedish 1969 conscription cohort[END_REF][START_REF] Kuh | Childhood cognitive ability and deaths up until middle age: a post-war birth cohort study[END_REF][START_REF] Martin | Childhood cognitive performance and risk of mortality: a prospective cohort study of gifted individuals[END_REF][START_REF] Hart | Childhood IQ and cardiovascular disease in adulthood: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF][START_REF] Batty | Does IQ explain socioeconomic inequalities in health? Evidence from a population based cohort study in the west of Scotland[END_REF][START_REF] Deary | Cognitive epidemiology[END_REF][START_REF] Batty | Premorbid (early life) IQ and later mortality risk: systematic review[END_REF][START_REF] Gottfredson | Intelligence predicts health and longevity, but why?[END_REF]. However, the mechanisms underlying the association between IQ and mortality remain uncertain [START_REF] Batty | Premorbid (early life) IQ and later mortality risk: systematic review[END_REF][START_REF] Gottfredson | Intelligence predicts health and longevity, but why?[END_REF]. One possibility is that IQ predicts mortality because IQ is related to social and economic circumstances predicting mortality. An alternative scenario suggests that the effect of these circumstances on mortality is largely accounted for by IQ.

It is well established that low IQ in childhood or adolescence predicts low socioeconomic status (SES) later in life [START_REF] Rowe | Herrnstein's syllogism: Genetic and shared environmental influences on IQ, education, and income[END_REF][START_REF] Strenze | Intelligence and socioeconomic success: A meta-analytic review of longitudinal research[END_REF] and that low SES, in turn, predicts mortality risk over the life course [START_REF] Smith | Lifetime socioeconomic position and mortality: prospective observational study[END_REF][START_REF] Mustard | Gender differences in socioeconomic inequality in mortality[END_REF]. These associations give rise to two competing hypotheses of IQ, SES, and mortality. One the one hand, the effect of IQ on mortality may be mediated via socioeconomic circumstances. This hypothesis has received provisional support, as education and occupational status have been found to explain the IQ-mortality association in part, although not completely [START_REF] Batty | Premorbid (early life) IQ and later mortality risk: systematic review[END_REF][START_REF] Gottfredson | Intelligence predicts health and longevity, but why?[END_REF]. On the other hand, it has been hypothesized that IQ may explain health disparities associated with SES [START_REF] Gottfredson | Intelligence: is it the epidemiologists' elusive "fundamental cause" of social class inequalities in health?[END_REF] by determining both SES and mortality risk, so that the association between SES and mortality is actually explained by IQ. This hypothesis has also received provisional support, as IQ has been found to account for the association between SES and mortality in part, although not completely [START_REF] Batty | Does IQ explain socioeconomic inequalities in health? Evidence from a population based cohort study in the west of Scotland[END_REF].

The study of IQ, SES, and mortality is important in understanding the mechanisms explaining the IQ-mortality association and for understanding the origins of the socioeconomic gradient in health and mortality [START_REF] Singh-Manoux | The role of cognitive ability (intelligence) in explaining the association between socioeconomic position and health: Evidence from the Whitehall II prospective cohort study[END_REF]. However, previous studies of IQ and mortality have included only limited assessments of SES: In most studies, adulthood SES has been assessed only in one point in time, which may be an insufficient method of evaluating the effect of IQ on lifetime SES and the influence of SES on mortality.

In the current study, we examined whether IQ assessed at the age 16 to 23 predicted death before the age of 47 in an ethnically heterogeneous sample of men and women participating in the U.S. National Longitudinal Survey of Youth [START_REF]NLSY79 User's guide: A Guide to the 1979-2004 National Longitudinal Survey of Youth data[END_REF]. The study design allowed us to examine whether the influence of IQ on mortality was mediated by lifecourse SES, i.e., education and household income, assessed 19 times over the follow-up period, and whether the association between SES and mortality was accounted for by IQ.

We also examined whether education level of the participants' parents accounted for part of the IQ-mortality association.

MATERIALS AND METHODS

Participants

The participants were 5682 men and 5639 women (total n=11321) participating in the representative U.S. National Longitudinal Survey of Youth [START_REF]NLSY79 User's guide: A Guide to the 1979-2004 National Longitudinal Survey of Youth data[END_REF]. In this study, a national sample of individuals born between 1957 and 1964 has been followed annually since 1979 and biannually after 1994 with most recent data available on the follow-up in 2004. The original sample (n=12686) consists of three subsamples aged 15 to 22 years at baseline in 1979: a representative sample of noninstitutionalized civilian youths (n=6111); a supplemental sample designed to oversample civilian Hispanic, black, and economically disadvantaged nonblack/non-Hispanic youths (n=5295); and a military sample (n=1280). Details of the sampling process have been reported elsewhere (22; http://www.bls.gov/nls/nlsy79.htm). Due to funding constraints, the number of interviewed military sample and supplemental sample members were limited after years 1984 and 1990, respectively. Here we included all the 11321 participants who had completed a cognitive test administered in 1980 and had data on covariates and on study participation status on at least one of the 19 subsequent follow-up interviews between 1981 and 2004.

Appropriate longitudinal sampling weights taking into account sex, race/ethnicity, year of birth, sample type, and location were applied in the survival analyses in order to adjust for differential probabilities of selection into the sample and for attrition. Thus, the sample yielded representative estimates for the U.S. population born between years 1957 and 1964. The study was approved by institutional review boards of the institutions conducting the surveys and informed consent was obtained complying with Federal law and the policies of the U.S. office of Management and Budget.

Mortality

Follow-ups have been carried out annually between 1979 and 1994 and bi-annually between 1994 and 2004, so study participation data allowed us to determine the mortality status of the participants with an accuracy of one year before 1994 and with two years after 1994. For participants who have not been interviewed in follow-ups, the reason for noninterview has been recorded (e.g., refusal, unknown location, or deceased). Participants have been recorded as deceased only if the death of the participant has been possible to confirm via a death certificate, an obituary, or other national death records. As we did not have exact data on participant's date of death, age at death was determined to be the age the participant would have been in the follow-up period in which the participant was first determined to be deceased. However, some individuals permanently lost to follow-up at some point of the study may have deceased without being recorded as such in the data. For non-deceased participants who were lost to follow-up, the end of study period was determined by the participant's age at the latest available interview. Table 1 shows the deaths and censored participants by study year. Of the participants who had data on IQ and other covariates, 360 were recorded as deceased by the end of the follow-up period. Data on the cause of death were not available, so the focus of this study was on all-cause mortality only. Given that the study has been set up to examine social and economic rather than epidemiological research questions, there were no repeated-measure data of medical risk factors and health behaviors associated with mortality that could have been incorporated in the present analyses.

IQ

In 1980 the then 16 to 23 year-old participants were administered the Armed Services Vocational Aptitude Battery (ASVAB), which consists of 10 subtests: general science, arithmetic reasoning, word knowledge, paragraph comprehension, numerical operations, coding speed, auto and shop information, mathematics knowledge, mechanical comprehension, and electronics information [START_REF]NLSY79 User's guide: A Guide to the 1979-2004 National Longitudinal Survey of Youth data[END_REF]. Four of the subtests comprise the Armed Forces Qualification Test (AFQT) which includes only the more general, less vocationallyspecific tests assessing general cognitive ability, and is calculated from the raw scores by the formula AFQT=arithmetic reasoning + word knowledge + paragraph comprehension + (mathematics knowledge/2). The scale was standardized (M=0, SD=1) in order to facilitate the interpretation of the statistical estimates.

SES indicators

Participant's SES was assessed on the basis of highest completed education and household income at each follow-up phase. The participants reported their highest completed grade on a 20-point scale (range from 0=no education to 20=8 th year of college or more; mean=12.7, SD=2.4). Household income was assessed on the basis of total net household income from all income sources of the participant and her/his spouse in past calendar year, reported by the participants in each follow-up. The scale was transformed by square root transformation and standardized (mean=0, SD=1).

Covariates

Other characteristics included age, sex, racial/ethnic background, baseline health status, parental education, and marital status.

Racial/ethnic background. In the National Longitudinal Survey of Youth the participants have been categorized into three ethnic/racial groups (Hispanic, Black, and White/other) based on self-reported data (protocol described in section 4.32 of the study user's guide, ref. [START_REF]NLSY79 User's guide: A Guide to the 1979-2004 National Longitudinal Survey of Youth data[END_REF]. This categorization was used in the present analyses. Of the participants included in the study, 1736 were Hispanic, 2777 were Blacks, and 6808 were White/other. Baseline health status. At baseline the participants reported whether they had a health problem that limited the kind or the amount of work they were able to do. The participants were mostly healthy at study inception, with 4.8% of the participants reporting a limiting health problem.

Parental education was assessed at baseline in 1979 on the basis of education level of the more educated parent of the participant (range from 0=no education to 20=8 th year of college or more; mean=11.7, SD=3.4).

Marital status. Given that marital status is associated with mortality risk [START_REF] Seeman | Social network ties and mortality among the elderly in the Alameda County Study[END_REF] and that income level was assessed on the basis of household income (see above), which is naturally related to marital status, marital status was included as a covariate. Marital status was reported by the participants in each follow-up, and was coded into three categories: never married, married, and separated/divorced/widowed.

Statistical analysis

The associations between IQ and completed education was assessed with 19 separate linear regressions predicting completed education level at each follow-up phase, adjusted for sex, birth year, and race/ethnicity, and using cross-sectional sampling weights. The association between IQ and household income was assessed in the same way, additionally adjusted for marital status. These analyses allowed us to examine whether the association between IQ and SES indicators changed over the follow-up period. The results were illustrated by plotting the standardized regression coefficients of IQ by follow-up phase.

As a preliminary analysis of IQ and mortality, we categorized the participants into low (IQ score below 1 SD of mean), average (IQ within 1 SD of mean) and high (IQ above 1 SD of mean) IQ groups, calculated sample hazard functions, i.e., the probability of dying at a given age, and plotted the corresponding mortality functions, i.e., the probability of dying by a given age, for each group (the 95% confidence intervals were calculated using Greenwood's approximation). This analysis provided crude sample mortality rates for the three IQ groups without sample weights or adjustment for covariates.

The association between IQ, SES, and mortality was then assessed with discrete-time survival analysis [START_REF] Singer | Applied longitudinal data analysis: modeling change and event occurrence[END_REF]. This analysis allowed us to take into account differences in followup period between the participants and to model time-dependent effects, e.g., life-course SES. Discrete-time rather than continuous-time survival analysis was applied due to the rather coarse measure of timing of death (see above). The members of the oldest cohort were 47 years old at the final follow-up phase, so survival analysis allowed us to estimate mortality up to this age. Age was used as the measure of time and given that mortality rate tends to accelerate with age we included linear and quadratic terms of age in all models.

Survival analysis was carried out in four steps. First we assessed how background covariates, SES, and IQ predicted mortality risk when these were examined in separate models. We then entered all covariates in a single model in order to assess the independent effects of IQ and SES. The interaction effect between IQ and age was tested in order to assess whether the influence of IQ on mortality was constant over age or whether this association strengthened or weakened with age. The results of the survival analysis were illustrated by calculating the predicted probability of death by the age of 47 by IQ level (from 2 SD below the mean to 2 SD above the mean).

As a sensitivity analysis, we examined the associations between the four AFQT subscales and mortality separately. We also assessed whether the association between IQ and mortality was moderated by sociodemographic background covariates, i.e., sex, race/ethnicity, and parental education, by testing interaction effects between these covariates and IQ in overall model. These interaction effects were further explored by fitting the survival models separately by sex, race/ethnicity, and levels of parental education.

The participants reported their education, household income and current marital status in annual or biannual follow-up interviews between 1981 and 2004, which resulted in 19 assessment times. These data allowed us to construct these three covariates as timevarying, i.e., having different values across the follow-up period. All missing values in these variables were replaced by data from the previous year. Other covariates were used as time-invariant, i.e., having a constant value over the follow-up period. Appropriate longitudinal sampling weights were used in the analyses. In addition, some of the participants were from the same households, so not all observations were independent.

Robust estimator with household clustering instead of maximum likelihood was therefore used to provide appropriate estimates of standard errors.

Finally, we carried out an attrition analysis by refitting the final survival analysis model of IQ and mortality but by having age at censoring rather than mortality as the outcome. This model was fitted by excluding the deceased participants and by applying the same sampling weights as in predicting mortality. No association between IQ and censoring suggest loss to follow-up to be an unlikely source of bias. Underestimation is possible if IQ is associated with censoring in a similar way as it is associated with mortality. The risk of overestimation increases if the IQ-censoring association is in the opposite direction than the IQ-mortality association.

RESULTS

IQ as a predictor of SES indicators

First we assessed how IQ predicted education and household income over the follow-up period. As shown by the plotted regression coefficients of IQ in Figure 1, the association between IQ and education increased from =0.51 to =0.69 (all p-values<0.001) between years 1981 and 2004. The coefficients between IQ and household income, in turn, increased from =0.19 to =0.35 (all p-values<0.001). These patterns indicated that IQ predicted SES more strongly in late than in early adulthood.

[FIGURE 1

HERE, WIDTH 1 COLUMN] IQ and SES indicators as predictors of mortality

In assessing the association between IQ and mortality, we first calculated and plotted crude mortality rates by IQ groups (Figure 2). Individuals with high IQ had a lower mortality rate than those with low IQ, and by the age of 47 the mortality functions had reached the values of 9%, 6%, and 3% in the low, average, and high IQ groups, respectively, suggesting a three-fold mortality differential between those with above-average IQ and those with below-average IQ.

[FIGURE 2

HERE, WIDTH 1 COLUMN]

We then modeled the associations between IQ, SES, and mortality using discrete-time survival analysis. Model 1 of Table 2 shows the association between background variables and mortality. Model 2 includes SES indicators and marital status in addition to background variables. Although participant's own education predicted decreased mortality when household income was not included in the model (OR=0.88, CI=0.83, 0.93, P<0.001; data not shown in the table), only income was significant when both SES indicators were entered in a single model (model 2). Models 3 and 4 repeated models 1 and 2 but included IQ as a covariate. When sex, birth year, race/ethnicity, and baseline health status were controlled, IQ predicted decreased mortality risk (OR=0.71, CI=0.62, 0.82, P<0.001; data not shown) and this association was little changed after including parental education in the model (OR=0.78; model 3). The relation between IQ and mortality was attenuated when adjusting for education and marital status in addition to background variables (OR=0.91, CI=0.75, 1.11, P=0.36; data not shown), and disappeared almost completely when further adjusted for household income (OR=0.99; model 4). Adjusting for marital status (but not for SES indicators) did not substantially alter the association between IQ and mortality (OR=0.82, CI=0.70, 0.97, P=0.02; data not shown), indicating that the influence of IQ on mortality risk was largely accounted for by life-course socioeconomic circumstances.

In order to illustrate the effect magnitude of IQ on mortality in terms of percentage points, we calculated the predicted probabilities of death by the age of 47 by IQ level (table 3), adjusted for background variables and parental education (step 1), for education and marital status (step 2), and for household income (step 3). The right-most column of table 3 reproduces the OR:s reported in the paragraph above and in table 2. Given the low absolute mortality probabilities, the OR:s closely equaled to relative risk ratios, i.e., the relative change in probability of death by the age of 47 associated with 1 SD change in IQ.

Sensitivity analyses

We examined interaction effects between age and IQ in order to evaluate whether the influence of IQ on mortality was dependent on age. None of these interaction effects were significant, suggesting that the association between IQ and mortality risk was constant over age. As a sensitivity analysis of the IQ measure, we examined whether different IQ subscales predicted mortality risk differently. Fitting model 3 of table 2 using standardized IQ subscales (Mean=0, SD=1) rather than the global score indicated that, in separate models, mortality was similarly predicted by arithmetic reasoning (OR=0.84, CI=0.72, 0.99, P=0.04), word knowledge (OR=0.88, CI=0.75, 1.02, P=0.09), paragraph comprehension (OR=0.80, CI=0.70, 0.92, P=0.04), and mathematics knowledge (OR=0.85, CI=0.75, 0.97, P=0.02), suggesting that the IQ-mortality association held across cognitive domains.

Next we examined the association between IQ and mortality separately within different ethnic/racial groups and within men and women. Race/ethnicity × IQ interactions were not statistically significant (p values>0.89), suggesting no differences between racial/ethic groups. IQ predicted mortality in a similar fashion in Hispanic (OR=0.67, CI=0.48, 0.92, P=0.012), Black (OR=0.71, CI=0.57, 0.88, P=0.002), and White/other (OR=0.83, CI=0.66, 1.03, P=0.09) participants. In all groups, adjusting for marital status, education, and household income attenuated the relation between IQ and mortality to a nonsignificant level (Hispanic: OR=0.85, CI=0.57, 1.28, P=0.44; Black: OR=0.96, CI=0.75, 1.24, P=0.75; White/other: OR=1.02, CI=0.88, 1.33, P=0.88) indicating that socioeconomic circumstances mediated the association in all ethnic/racial groups.

The association between IQ and mortality was also similar in men (OR=0.77, CI=0.64, 0.94, P=0.009) and in women (OR=0.78, CI=0.58, 1.05, P=0.11), and this association disappeared when socioeconomic status was controlled for (men: OR=0.99, CI=0.78, 1.26, P=0.94; women: OR=1.00, CI=0.71, 1.39, P=0.98).

Finally, we examined whether the association between IQ and mortality was moderated by parental education by including IQ × parental education interaction term in model 1 of table 2 (data not shown). This interaction effect was statistically significant (OR=0.95, CI=0.91, 1.00, P=0.05) and suggested that IQ predicted mortality more strongly in participants with higher parental education. To illustrate this we fitted the survival analysis model separately by three levels of parental education (less than completed high school, completed high school, more than high school

Attrition analysis

We performed attrition analysis by fitting survival analysis model predicting age at censoring. In interpreting these results, it needs to be kept in mind that certain subsamples were excluded by study design in 1984 and 1990 and that the model was fitted using sample weights correcting for attrition (see method section). Adjusted for age and age squared, drop-out was predicted by birth year (OR=1.89, CI=1.83, 1.95, P<0.001), Black (OR=0.85, CI=0.82, 0.89, P<0.001) and Hispanic (OR=0.88, CI=0.84, 0.93, P<0.001) ethnicity, parental education (OR=0.99, CI=0.98, 1.00, P=0.02), health limitations (OR=0.91, CI=0.85, 0.97, P=0.01), IQ (OR=0.97, CI=0.95, 1.00, P=0.05), marital status (married: OR=1.04, CI=0.98, 1.11, P=0.24; divorced: OR=1.11, CI=1.04, 1.18, P=0.002), and education (OR=1.02, CI=1.01, 1.03, P<0.001) but not by household income (OR=1.00, CI=0.98, 1.02, P=0.95). Thus, the attrition effect of IQ in our analyses was small and, if anything, may have attenuated the estimated association between IQ and mortality, as some deceased participants with low IQ (and hence with high mortality risk) might have been lost to follow-up rather than having been recorded as deceased in the dataset.

DISCUSSION

The present study provides evidence for the role of IQ in predicting early mortality in the United States. Independently of age, sex, race/ethnicity, baseline health limitations, and parental education, high IQ in young adulthood predicted low mortality risk by the age of 47 years: An IQ one SD above the mean was associated with an approximately 22% decrease in the odds of death. This association was completely mediated by life-course socioeconomic circumstances, i.e., educational level and household income, which were predicted by IQ in the expected way. The IQ-mortality association was similar in Hispanic, Black, and White/other participants and also in men and women.

The main limitation of the study is that the mortality data were based on study participation data, so that individuals permanently lost to follow-up had to be censored at the age of last available interview, because we did not have mortality data for them after drop-out. Although survival analysis takes censoring into account in estimation (i.e., that not all participants are followed through the same period of time), selective attrition may still have biased the mortality estimates. However, we believe that a major attrition bias in this study is unlikely. If anything, the weak associations between IQ and attrition might have served to bias our estimates conservatively.

Another potential limitation of this study is that IQ was assessed in adolescence or young adulthood when the participants had already went through years of education. Thus, we cannot exclude the possibility of reverse causality between IQ and SES, education in particular. The association between IQ and SES grew stronger with age, suggesting that at least part of the association flowed from IQ to SES rather than the reverse. Namely, the strengthening association between IQ and later SES would be difficult to explain by reverse causality whereas it is plausible that the influence of IQ on SES becomes stronger with age as individuals have more opportunities to attain higher SES.

According to U.S. mortality statistics [START_REF]Web-based Injury Statistics Query and Reporting System[END_REF] covering the present study period, the five leading causes of death between ages 16 and 47 were unintentional injuries (32% of all deaths), cancer (18%), cardiovascular disease (14%), suicide (12%) and homicide (9%).

Previous studies have associated IQ with unintentional injuries [START_REF] Osler | Cognitive function in childhood and early adulthood and injuries later in life: the Metropolit 1953 male birth cohort[END_REF][START_REF] Lawlor | Associations between childhood intelligence and hospital admissions for unintentional injuries in adulthood: the Aberdeen Children of the 1950s cohort study[END_REF], suicidal behavior [START_REF] Gunnell | Is low IQ associated with an increased risk of developing suicidal thoughts? : A cohort study based on an 18-month follow-up of the national psychiatric morbidity survey[END_REF], homicide risk [START_REF] Batty | Is low IQ related to risk of death by homicide? Testing a hypothesis using data from the Vietnam Experience Study[END_REF], and cardiovascular disease [START_REF] Hemmingsson | The association between cognitive ability measured at ages 18-20 and coronary heart disease in middle age among men: A prospective study using the Swedish 1969 conscription cohort[END_REF][START_REF] Hart | Childhood IQ and cardiovascular disease in adulthood: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF] whereas the IQ-cancer association has received mixed support (4,7). We did not have cause-specific mortality data, but given the age of the study sample it seems reasonable to assume that a large part of the deaths in the sample were due to non-intentional injuries. This would also be in agreement with the observation that the IQ-mortality association was mediated by SES, because low SES is associated with exposure to hazardous environments and increased risk of non-intentional injuries [START_REF] Turner | Socio-economic distribution of environmental risk factors for childhood injury[END_REF][START_REF] Cubbin | Socioeconomic status and injury mortality: individual and neighbourhood determinants[END_REF][START_REF] Whitlock | Motor vehicle driver injury and socioeconomic status: a cohort study with prospective and retrospective driver injuries[END_REF][START_REF] Gottfredson | Innovation, fatal accidents, and the evolution of general intelligence[END_REF]. Furthermore, IQ and SES are both known to be associated with health behaviors [START_REF] Gottfredson | Intelligence: is it the epidemiologists' elusive "fundamental cause" of social class inequalities in health?[END_REF][START_REF] Hemmingsson | How does IQ affect onset of smoking and cessation of smoking-Linking the Swedish 1969 Conscription Cohort to the Swedish Survey of Living Conditions[END_REF][START_REF] Batty | Mental ability across childhood in relation to risk factors for premature mortality in adult life: the 1970 British Cohort Study[END_REF][START_REF] Batty | Childhood IQ in relation to risk factors for premature mortality in middle-aged persons: the Aberdeen Children of the 1950s study[END_REF][START_REF] Adler | Socioeconomic status and health. The challenge of the gradient[END_REF][START_REF] Anderson | Toward understanding the association of socioeconomic status and health: a new challenge for the biopsychosocial approach[END_REF][START_REF] Hanson | Socioeconomic status and health behaviors in adolescence: a review of the literature[END_REF], so part of the SES-mediated influence of IQ on mortality may reflect SES differences in health behaviors.

The present findings add to the cognitive epidemiology literature [START_REF] Deary | Cognitive epidemiology[END_REF] by showing that the association between IQ and mortality generalizes to populations not studied previously.

Previous studies in the United States [START_REF] Martin | Childhood cognitive performance and risk of mortality: a prospective cohort study of gifted individuals[END_REF] have not been based on representative samples, and nearly all earlier studies have included only white participants [START_REF] Batty | Premorbid (early life) IQ and later mortality risk: systematic review[END_REF]. In the present sample IQ predicted early mortality with a similar effect size as has been observed in previous studies [START_REF] Hart | Childhood IQ, social class, deprivation, and their relationships with mortality and morbidity risk in later life: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF][START_REF] Whalley | Longitudinal cohort study of childhood IQ and survival up to age 76[END_REF][START_REF] Pearce | Childhood IQ and deaths up to middle age: The Newcastle Thousand Families Study[END_REF][START_REF] Batty | IQ in early adulthood and later cancer risk: cohort study of one million Swedish men[END_REF][START_REF] Hemmingsson | The association between cognitive ability measured at ages 18-20 and coronary heart disease in middle age among men: A prospective study using the Swedish 1969 conscription cohort[END_REF][START_REF] Kuh | Childhood cognitive ability and deaths up until middle age: a post-war birth cohort study[END_REF][START_REF] Martin | Childhood cognitive performance and risk of mortality: a prospective cohort study of gifted individuals[END_REF][START_REF] Hart | Childhood IQ and cardiovascular disease in adulthood: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF][START_REF] Batty | Does IQ explain socioeconomic inequalities in health? Evidence from a population based cohort study in the west of Scotland[END_REF][START_REF] Deary | Cognitive epidemiology[END_REF][START_REF] Batty | Premorbid (early life) IQ and later mortality risk: systematic review[END_REF], suggesting that IQ predicts mortality in a similar fashion in the United States as in European countries, across different racial/ethnic groups, and early and later mortality.

Previous research has found SES to account for the IQ-mortality association only in part whereas we found that life-course socioeconomic circumstances accounted for the IQmortality association completely. There are at least three possible reasons for this difference. First, the differing findings might reflect differences between countries. This explanation is implausible, however, because IQ predicts SES in a similar fashion in the United States as in other western countries [START_REF] Strenze | Intelligence and socioeconomic success: A meta-analytic review of longitudinal research[END_REF]. Second, we assessed life-course educational level and household income with 19 assessment times, which provides a more accurate measure of SES and therefore a stronger test of the mediator hypothesis than used in most previous studies. Third, the present study assessed only early mortality up to midlife, whereas some previous studies [START_REF] Hart | Childhood IQ, social class, deprivation, and their relationships with mortality and morbidity risk in later life: prospective observational study linking the Scottish Mental Survey 1932 and the Midspan studies[END_REF][START_REF] Whalley | Longitudinal cohort study of childhood IQ and survival up to age 76[END_REF] have assessed mortality up to older ages. It is possible that socioeconomic status is more important in mediating early mortality risk than mortality at later ages.

The association between IQ and mortality risk was dependent on parental education. This association was clear in individuals with high parental education but absent in those whose parents had low education. The absence of IQ-mortality association in participants from less educated households may reflect their increased exposure to other mortality risk factors that may override the influence of IQ. Furthermore, behavior genetic studies have shown that genetic factors may be less influential in determining IQ differences in individuals from families with low rather than with high SES [START_REF] Rowe | Genetic and environmental influences on vocabulary IQ: parental education level as moderator[END_REF][START_REF] Turkheimer | Socioeconomic status modifies heritability of IQ in young children[END_REF][START_REF] Harden | Genotype by environment interaction in adolescents' cognitive aptitude[END_REF]. Thus, IQ scores in individuals with low parental SES may disproportionally reflect consequences of early environment rather than their genetic potential influencing the extent to which people are successful in selecting healthy environments and life styles after childhood. Replications of the IQ × parental SES interaction effect and genetically informative studies are needed to assess this hypothesis.

Social disparities in mortality are an important public health issue, but factors explaining such disparities have remained unclear. We found a stronger association between income and mortality than between education and mortality suggesting that material factors are important in social disparities [START_REF] Smith | Education and occupational social class: which is the more important indicator of mortality risk?[END_REF]. Our findings do not support the hypothesis [START_REF] Batty | Does IQ explain socioeconomic inequalities in health? Evidence from a population based cohort study in the west of Scotland[END_REF][START_REF] Gottfredson | Intelligence: is it the epidemiologists' elusive "fundamental cause" of social class inequalities in health?[END_REF] that socioeconomic disparities in early mortality would reflect IQ differences associated with SES. Adjusting for IQ did not alter the robust association between socioeconomic circumstances and mortality risk, implying that SES was an independent predictor of mortality even when IQ was taken into account. This suggests that reducing socioeconomic inequalities should reduce the mortality differentials associated with IQ differences.

In conclusion, our study of early mortality in the United States suggests that IQ may be one of the factors contributing to early death via its predictive association with lifetime socioeconomic circumstances. SES, in turn, is associated with mortality risk in a manner not confounded by IQ differences associated with SES. Step 1 adjusts for sex, birth year, race/ethnicity, baseline health, and parental education

Step 2 adjusts for step 1 + marital status and own education

Step 3 adjusts for step 2 + household income See table 2 for statistical details

Figure 1 .Figure 2 .

 12 FIGURES

Table 1 .

 1 Life table of the sample by study years.

		Risk set	Deaths	Censored
	Year
	1980	11321	0	0
	1981	11321	10	0
	1982	11301	13	18
	1983	11270	13	62
	1984	11195	9	951
	1985	10235	12	39
	1986	10184	14	32
	1987	10138	13	41
	1988	10084	16	52
	1989	10016	12	67
	1990	9937	9	1382
	1991	8546	12	44
	1992	8490	11	39
	1993	8440	17	68
	1994	8355	23	90
	1996	8242	35	118
	1998	8089	28	276
	2000	7785	35	300
	2002	7450	31	371
	2004	7048	47	7001
	Note: Risk set includes participants who
	were alive and for whom data on mortality
	status were available. Participants were
	censored due to study drop-out and funding
	constraints limiting the number of
	participants in 1984 and 1990.

Table 2 .

 2 Predicting mortality risk by IQ and sociodemographic covariates. Odds ratios (95% confidence intervals) from nested discrete-time survival analyses (n=11321).

				Model 1	Model 2		Model 3	Model 4
				OR	95% CI	OR	95% CI	OR	95% CI	OR	95% CI
	Gender	0.44*** 0.33, 0.58 0.46*** 0.35, 0.61 0.44*** 0.33, 0.58 0.46***	0.35, 0.61
	Birth year	1.08*	1.01, 1.15	1.04	0.98, 1.12	1.03	0.96, 1.10	1.04	0.97, 1.12
	Race/ethnicity
	Black	1.62*** 1.25, 2.09	0.99	0.75, 1.31	1.28	0.96, 1.71	0.98	0.72, 1.33
	Hispanic		0.93	0.63, 1.39	0.82	0.55, 1.22	0.84	0.57, 1.25	0.81	0.54, 1.21
	White/other		1.00		1.00		1.00		1.00
	Parent's education	0.93*** 0.89, 0.96	0.96 *	0.92, 1.01 0.96* 0.91, 1.00	0.96	0.92, 1.01
	Baseline health limitation	1.80*	1.07, 3.04	1.61	0.94, 2.74 1.74* 1.03, 2.94	1.61	0.94, 2.73
	Marital status
	Never married				1.00				1.00
	Married				0.36*** 0.25, 0.53			0.36***	0.25, 0.53
	Divorced/Widowed				0.67* 0.47, 0.98			0.68*	0.47, 0.98
	Own education				0.95	0.89, 1.02			0.95	0.88, 1.03
	Household income z				0.66*** 0.55, 0.78			0.66***	0.55, 0.79
	IQ	z						0.78** 0.66, 0.91	0.99	0.81, 1.20
	Note: *** p<0.001, ** p<0.01, * p<0.05; z standardized ORs (Mean=0, SD=1)
	All models adjust for Age and Age	2 for the passage of time

Table 3 .

 3 The probability of death (%) by the age of 47 by IQ level. Predictions from discretetime survival analysis models (n=11321). Given the relatively low absolute percentage points, the OR:s also provide an approximate relative risk of death by the age of 47 associated with 1 standard deviation change in IQ, i.e., RR p(+1 SD)/ p(Mean).

				IQ
		-2 SD	-1 SD	Mean	+1 SD	+2 SD	OR
	Step 1	9.0	7.1	5.5	4.3	3.4	0.78
	Step 2	5.7	5.2	4.8	4.4	4.0	0.91
	Step 3	4.7	4.6	4.5	4.5	4.4	0.99
	Note:

Acknowledgements: ASM was supported by a 'EURYI' award from the European Science Foundation and a 'Chaire d'excellence' award from the French Ministry of Research. MK was supported by the Academy of Finland (grants 124322 and 117604).


Brian M Hays

Divita Gupta

Théo Guillaume

Omar Abdelkader Khedaoui

Ilsa R Cooke

Franck Thibault

François Lique

Ian R Sims
email: ian.sims@univ-rennes1.fr

Theo Guillaume

Collisional excitation of HNC by He found to be stronger than for structural isomer HCN in experiments at the low temperatures of interstellar space

observed abundance ratio of these molecules as a thermometer, [START_REF] Hacar | HCN-to-HNC intensity ratio: a new chemical thermometer for the molecular ISM[END_REF] but is not sufficient to explain the enhanced abundance of HNC in certain environments.

The determination of molecular abundances from observational spectra requires knowledge of the population of the rotational energy levels of the interstellar molecules. In the ISM, the density is generally not high enough to maintain local thermodynamic equilibrium conditions. As a consequence, the population of the molecular levels is controlled by the competition between collisional (de-)excitation and radiative processes.

Given the similar dipole moments and rotational constants for HCN and HNC, it had been assumed that the excitation conditions (i.e. the populations of the rotational states) were almost identical, [START_REF] Graninger | The HNC/HCN ratio in star-forming regions[END_REF] and these were computed on the basis of only HCN collisional rate coefficients. The HNC/HCN line intensity ratio then gives, as a first approximation, the HNC/HCN abundance ratio. Because the HNC lines are much stronger than those of HCN in the cold ISM, the HNC/HCN abundance ratio was found to be much greater than 1, in strong disagreement with existing astrochemical models. [START_REF] Hirota | Abundances of HCN and HNC in Dark Cloud Cores[END_REF] To explain this, Lique and co-workers proposed that HCN and HNC do not undergo identical collisional excitation, contrary to what had been widely assumed. [START_REF] Vera | The rotational excitation of the HCN and HNC molecules by H₂ revisited[END_REF][START_REF] Sarrasin | The rotational excitation of HCN and HNC by He: new insights on the HCN/HNC abundance ratio in molecular clouds[END_REF] They obtained high level ab initio potential energy surfaces (PESs) for HCN and HNC with both He and H2, the dominant colliders in the cold ISM, [START_REF] Sarrasin | The rotational excitation of HCN and HNC by He: new insights on the HCN/HNC abundance ratio in molecular clouds[END_REF][START_REF] Denis-Alpizar | A new ab initio potential energy surface for the collisional excitation of HCN by para-and ortho-H₂[END_REF][START_REF] Dumouchel | The rotational excitation of HCN and HNC by He: temperature dependence of the collisional rate coefficients[END_REF][START_REF] Dumouchel | The rotational excitation of the interstellar HNC by para-and ortho-H₂[END_REF] and performed close-coupling scattering calculations to generate collisional rate coefficients for rotational energy transfer (RET). They observed strong differences between the two isomers and attributed this to a change in anisotropy between the two potentials. When these newly calculated rate coefficients, now differing strongly between these two isomers, are used in conjunction with observational data, significantly lower abundance ratios are found (∼1-2), compatible with the predictions of astrochemical models. However, these calculations and their important astrochemical consequences have not yet been supported by any experimental measurements.

While HCN and HNC are readily identified in the ISM, few laboratory experiments are available to confirm the theoretically supported kinetics or RET rates for these molecules. In fact, no experiments of this kind have been performed on HNC, presumably owing to its instability in the laboratory making it impossible to obtain pure samples. While vibrational spectroscopy has been used to examine these isomers in collisional or photodissociative environments, [START_REF] Wilhelm | Is Photolytic Production a Viable Source of HCN and HNC in Astrophysical Environments? A Laboratory-based Feasibility Study of Methyl Cyanoformate[END_REF][START_REF] Wilhelm | Collisional Energy Transfer from Vibrationally Excited Hydrogen Isocyanide[END_REF] only recently has rotational spectroscopy become sensitive enough to allow for the detection of these species through transient experiments. [START_REF] Prozument | Photodissociation transition states characterized by chirped pulse millimeter wave spectroscopy[END_REF][START_REF] Prozument | A new approach toward transition state spectroscopy[END_REF] Here we describe an experiment to investigate the collisional properties of HCN and HNC with He at low temperatures, combining simultaneous in situ generation of HCN and HNC by laser photolysis in continuous uniform supersonic flows of cold He with time-resolved chirped pulse Fourier transform millimeter wave (CPFTmmW) spectroscopy. [START_REF] Brown | A broadband Fourier transform microwave spectrometer based on chirped pulse excitation[END_REF][START_REF] Park | Design and evaluation of a pulsed-jet chirped-pulse millimeter-wave spectrometer for the 70-102 GHz region[END_REF] We measure the pressure broadening cross sections for these two systems down to 10 K and thereby verify the theoretically predicted difference in collisional excitation on an absolute scale.

Results

Experimental protocols

The experiments were performed in a recently developed CPUF (Chirped Pulse in Uniform supersonic Flow) apparatus. The CPUF technique was first developed by Suits and coworkers [START_REF] Abeysekera | A chirped-pulse Fourier-transform microwave/pulsed uniform flow spectrometer. II. Performance and applications for reaction dynamics[END_REF] and is based on the cold collisional environment provided by a CRESU (Cinétique de Réaction en Ecoulement Supersonique Uniforme or Reaction Kinetics in Uniform Supersonic Flow) apparatus, [START_REF] Rowe | Study of the reactions N₂ + +2N₂→N₄ + +N₂ and O₂ + +2O₂→O₄ + +O₂ from 20 to 160 K by the CRESU technique[END_REF][START_REF] Sims | Rate constants for the reactions of CN with hydrocarbons at low and ultra-low temperatures[END_REF] combined with in-situ production of reactive or ACCEPTED MANUSCRIPT / CLEAN COPY transient species by pulsed laser photolysis and CPFTmmW detection. The CPUF apparatus used here differs from the previous version of Suits and co-workers principally in its use of continuous rather than pulsed uniform supersonic flows, giving more precisely defined flow conditions and enabling higher repetition rate pulsed photolysis (100 Hz in this case). It also employs a higher power CPFTmmW spectrometer which is fully described in Hays et al. [START_REF] Hays | Design and performance of an E-band chirped pulse spectrometer for kinetics applications: OCS -He pressure broadening[END_REF] Fig. 1a shows a schematic diagram representing the essential features of the experiment. Briefly, the cold core of a uniform supersonic flow of He containing trace amounts of vinyl cyanide at temperatures ranging from 10-70 K is illuminated with an ultraviolet pulsed excimer laser beam at 193 nm. This causes photodissociation of the vinyl cyanide, yielding photoproducts including HCN and HNC. [START_REF] Prozument | Photodissociation transition states characterized by chirped pulse millimeter wave spectroscopy[END_REF] The initially rotationally and translationally hot photoproducts undergo rapid thermalization by frequent collisions with the relatively dense, cold He. The thermalized HCN and HNC are then excited and detected in a time-resolved fashion on the j=1-0 transition by a CPFTmmW spectrometer arranged perpendicularly to the flow, yielding the pressure broadening cross-sections as a function of delay time after the excimer laser pulse. At a given time delay after the photolysis laser fires, a short (140-200 ns) pulse of amplified mm-wave radiation is broadcast into the flow from the transmitter, after which the molecular signal (the free induction decay, or FID) is collected and amplified by the receiver and recorded over ∼1 μs. After background subtraction these molecular FIDs can be fitted in the time-domain to yield the so-called T2 relaxation time for each molecule as a function of time delay after its creation by laser photolysis. This delay corresponds to the position in the flow where the molecules were generated, as can be seen in Fig. 1b which shows a typical 2-D density map of the uniform supersonic flow (obtained by impact pressure measurements) above Fig. 1c which displays the measured T2 relaxation times for HCN and HNC as a function of time after the laser shot. Further details may be found in the Methods section.

Time domain data analysis

The rate of decay of the FID was fitted to obtain the pressure broadening cross sections in these experiments. As described in a recent study, [START_REF] Hays | Design and performance of an E-band chirped pulse spectrometer for kinetics applications: OCS -He pressure broadening[END_REF] both the time domain and the frequency domain contain the same information, but the effects of interference from hyperfine splitting are more easily accounted for in the time domain. Examples of these fits are shown in Fig. 2 for HCN and HNC. Fig. 2 panels b andd show the FFTs (Fast Fourier Transforms) of both the data and the time domain fit, to show how well the FID decay is captured by the model. A time domain Voigt model was used to fit this data, which also accounts for the hyperfine splitting of HCN that imposes a beat frequency along with the FID. This model is discussed in the Supplementary Material.

The fitted exponential decays are shown in Fig. 2 panel e for HCN and HNC, without any frequency components. It is easily seen that the decay times (T2) for each of these molecules is different, despite being produced under the same conditions. This is entirely due to the difference in pressure broadening cross section of these two molecules. The T2 decay times are converted into pressure broadening cross sections σ(T) using the precisely known density and temperature of the uniform supersonic flow as described in the Supplementary Materials (Supplementary Eq. 2). Only molecules isolated within the cold uniform flow were included in the pressure broadening data, which could be easily distinguished from molecules produced in the nozzle by time dependent changes of the decay rate at the end of the collected flow data as shown in Fig. 1c.

ACCEPTED MANUSCRIPT / CLEAN COPY

Theoretical calculations

Several HCN-He and HNC-He potential energy surfaces (PESs) have been published (e.g. [START_REF] Sarrasin | The rotational excitation of HCN and HNC by He: new insights on the HCN/HNC abundance ratio in molecular clouds[END_REF] ). However, it is well known that scattering calculations under cold conditions are highly sensitive to the PES accuracy. Hence, using state of the art ab initio methods, we computed highly accurate two-dimensional HCN-He and HNC-He PESs with HCN and HNC intramolecular distances fixed. Theoretical pressure broadening cross sections and rate coefficients for the j = 0 and 1 levels of HCN and HNC were obtained by performing quantum dynamics 28 on these PESs using the MOLSCAT code. [START_REF] Hutson | MOLSCAT computer code, version 14[END_REF] Because of the low temperatures considered in this work, a full close-coupling approach was employed to compute the related cross sections. [START_REF] Shafer | Quantum scattering theory of rotational relaxation and spectral line shapes in H₂-He gas mixtures[END_REF] More details on these calculations can be found in the Methods section and in the Supplementary section 'Theoretical calculations'.

Sarrasin et al. 2010 and references therein

The experimentally determined pressure broadening cross sections are compared against those found from scattering calculations from the ab initio PESs used for each molecule in Fig. 3. The results are also reported in Supplementary Table 1.

Discussion

The experimental and theoretical pressure broadening cross sections from this work are compared together and against previously available experimental data in Fig. 3. Importantly, these cross sections are presented on an absolute scale, directly calibrating the magnitude of cross sections used in these calculations. The agreement between the experimental data and the scattering calculations from this work over the temperature range 10-70 K is very good. The j=1-0 transition for HCN in He has been experimentally observed on several occasions above 130 K [START_REF] Rohart | HCN absorption line shapes studied by millimeter wave coherent transients: speed dependent effects and collision interaction potential[END_REF] and by Ronningen and de Lucia between 1.3 K and 6.2 K. [START_REF] Ronningen | Helium induced pressure broadening and shifting of HCN hyperfine transitions between 1.3 and 20[END_REF] At higher temperatures, the current calculations agree very well with previous experiments. The highest temperature data from Ronningen and de Lucia are just below the lowest temperature at which the current experiments were performed (10 K). However, it can be seen from Fig. 3 that the calculations predict the cross section to rise as the temperature drops, and that this is confirmed by the current experiments down to 10 K. Other studies have called into question the absolute accuracy of the collisional cooling technique used by Ronningen and de Lucia especially at very low temperatures. [START_REF] Ronningen | Helium induced pressure broadening and shifting of HCN hyperfine transitions between 1.3 and 20[END_REF][START_REF] Thachuk | Linewidths and shifts of very low temperature CO in He: A challenge for theory or experiment?[END_REF] It should be emphasized, though, that the temperature ranges of the two sets of measurements do not overlap. For the unstable HNC isomer, this is the only pressure broadening study to the authors' knowledge, and so no data were available for comparison. The current experimental cross sections are in very good agreement with the calculated values, with an average RMS deviation of only 18 % and 25 % for the HCN-He and HNC-He cross sections respectively.

The rising trend with decreasing temperature is well reproduced as is the overall magnitude for HNC-He, which is significantly higher than for HCN-He. Pressure broadening cross sections typically do not offer the most sensitive test of PESs and scattering calculations due to the summation of both inelastic and elastic contributions, [START_REF] Ben-Reuven | Impact Broadening of Microwave Spectra[END_REF] followed by integration over a thermal distribution of energies to give the temperaturedependent cross sections measured and calculated here. All this averaging normally reduces the sensitivity of these measurements to differences in PESs, though the effect of averaging is diminished at low temperatures.

What is so striking about this study is that even after this blurring of the microcanonical processes and the ACCEPTED MANUSCRIPT / CLEAN COPY apparent similarities between HCN and HNC, the difference in these systems is still readily apparent from the measurements.

The larger pressure broadening cross sections observed for HNC-He versus HCN-He, especially at very low temperatures, are partly due to the greater well depth for the HNC-He PES versus the HCN-He PES (46.09 versus 30.10 cm -1 ). Another contributing factor can be understood by examining the state-to-state rate coefficients for inelastic rotational energy transfer out of the j = 0 and 1 levels, which constitute the main contribution to the measured pressure-broadening cross sections. They have been explicitly calculated by Sarrasin et al., [START_REF] Sarrasin | The rotational excitation of HCN and HNC by He: new insights on the HCN/HNC abundance ratio in molecular clouds[END_REF] and are displayed for a temperature of 10 K in Supplementary Fig. 2. As noted by Sarrasin et al., the HCN-He system with its less anisotropic potential appears to resemble more closely the case of an "almost homonuclear" diatomic molecule -atom system than does HNC-He, resulting in a propensity for even Δj transitions which is not observed for HNC-He. Such behavior was predicted long ago by McCurdy and Millar [START_REF] Mccurdy | Interference effects in rotational state distributions: Propensity and inverse propensity[END_REF] in terms of semiclassical theory, as being due to destructive interference in the case of odd Δj transitions. At these low temperatures, the rate coefficients drop strongly with Δj according to the usual exponential gap law, as can be seen for the HNC-He case in Fig. S2, where by far the largest state-to-state rate coefficients are for Δj = +1 and (when possible) Δj = -1. However, in the "almost homonuclear" HCN-He system, the even Δj propensity rule results in a strong suppression of these Δj = ±1 rate coefficients, resulting in a much lower overall pressure broadening cross section. This effect is particularly important at the lowest temperatures where higher positive Δj rate coefficients, corresponding to endoergic processes, are also suppressed. In space, these same processes compete with spontaneous emission from j=1-0, the most commonly monitored transition in astronomical observations of these molecules. [START_REF] Vera | The rotational excitation of the HCN and HNC molecules by H₂ revisited[END_REF] The strong difference in inelastic cross sections for j=1-0 between HCN and HNC with He -almost four times -is expected to preferentially pump into the j=1 level for HNC.

The difference between the pressure broadening cross sections for HCN and HNC validates the explanation proposed by Lique and coworkers [START_REF] Vera | The rotational excitation of the HCN and HNC molecules by H₂ revisited[END_REF] for the HNC/HCN abundance ratio, at least in the case of one of the main colliders in dark clouds, He atoms. This is further strengthened by the good agreement on an absolute scale between the scattering calculations and experimental measurements. The inelastic rate coefficients calculated previously by Lique and co-workers for these systems [START_REF] Sarrasin | The rotational excitation of HCN and HNC by He: new insights on the HCN/HNC abundance ratio in molecular clouds[END_REF][START_REF] Dumouchel | The rotational excitation of HCN and HNC by He: temperature dependence of the collisional rate coefficients[END_REF] agree very well with those computed here, differing by less than 5% on the overlapping temperature range. This confirms the validation of these previous results. Both theory and experiment now demonstrate that the assumption of identical RET rate coefficients for the two isomers is incorrect. Similar effects of propensity rules were also predicted for collisions of HCN and HNC with the other main collider, H2, but only for the para (j=0) level. [START_REF] Vera | The rotational excitation of the HCN and HNC molecules by H₂ revisited[END_REF] The magnitudes and differences were found to be similar to HCN and HNC with He. Current predictions using these collisional data still find that HNC can be more abundant in some of the coldest regions, hinting at unexplored chemical routes to this overabundance. [START_REF] Riaz | Chemical tracers in proto-brown dwarfs: CN, HCN, and HNC observations[END_REF] Comparison of the isoelectronic systems HCN and N2H + with He have revealed different propensity rules even in such similar cases. [START_REF] Pursell | Collision-induced double resonance studies of HN + 2 and HCN[END_REF] Our experiment is the only one to definitively find a difference in an isomeric collisional system, but other such examples may be present in the ISM. Several other cyanide/isocyanide species have been explored in theoretical studies. [START_REF] Vera | Cyanide/isocyanide abundances in the interstellar medium -II. Inelastic rate coefficients of Al and Mg compounds[END_REF][START_REF] Vera | Cyanide/isocyanide abundances in the interstellar medium -III. The excitation of Al and Mg compounds[END_REF][START_REF] Vera | Cyanides/isocyanides abundances in the interstellar medium -IV. Temperature dependence of SiCN/SiNC rate coefficients and astrophysical applications[END_REF][START_REF] Senent | Cyanide/isocyanide abundances in the interstellar medium -I. Theoretical spectroscopic characterization[END_REF][START_REF] Bop | Isomerism Effects in the Collisional Excitation of Cyanoacetylene by Molecular Hydrogen[END_REF] Abundance ratios of longer chain cyano/isocyano molecules have been determined in recent observations, [START_REF] Cernicharo | Discovery of HC₄NC in TMC-1: A study of the isomers of HC₃N, HC₅N, and HC₇N[END_REF][START_REF] Xue | Detection of Interstellar HC₄NC and an Investigation of Isocyanopolyyne Chemistry under TMC-1 Conditions[END_REF] and these may also be susceptible to different excitation conditions in the same way as for HCN and HNC with He. The pressure broadening of HCN and HNC ACCEPTED MANUSCRIPT / CLEAN COPY at low temperatures may also be of interest for observations of cold planetary atmospheres, such as that of Neptune. [START_REF] Rezac | New determination of the HCN profile in the stratosphere of Neptune from millimeter-wave spectroscopy[END_REF] The wide variety of HNC/HCN abundance ratios observed in comets (ranging from ∼ 0.03 to 0.3) remains puzzling and the formation mechanism of HNC is still unclear. [START_REF] Mumma | The Chemical Composition of Comets-Emerging Taxonomies and Natal Heritage[END_REF][START_REF] Lis | Hydrogen Isocyanide in Comet 73P/Schwassmann-Wachmann (Fragment B)[END_REF] Our study demonstrates the importance of specific calculations and measurements of collisional excitation for key astrophysically relevant systems.

The predictions made from high level scattering calculations are often compared against systems that are amenable for experimental measurement, usually employing laser excitation for detection, sometimes combined with velocity map imaging and other sophisticated techniques as Stark deceleration. [START_REF] De Jongh | Imaging the onset of the resonance regime in low-energy NO-He collisions[END_REF] While such experiments permit exquisite comparisons between theory and experiment, they are limited to a number of 'model' systems such as NO or OH with established laser detection schemes. The use of a chirped pulse spectrometer allows for a much wider variety of molecules to be probed. The cooling provided by the CRESU technique pushes the available energies into the regime where quantum effects are important, and the fact that the flow density and temperature are precisely known permits absolute measurements, further testing ab initio theories.

Conclusions

This study reports the only experiments to date showing that collisional excitation differs between structural isomers at interstellar temperatures. HCN and HNC have been formed together in situ in cold He supersonic flows, and a strong difference has been observed in the pressure broadening cross sections for the two isomers, especially at the lowest temperatures. Theoretical calculations of the pressure broadening parameters for HCN and HNC with He have been made to compare against these results, and excellent agreement has been observed on an absolute scale. This confirms previous theoretical arguments of Lique and coworkers regarding the importance of differences in PESs of structural isomers with He, and it is to be expected that other systems should display this too. This is one of the few experiments to specifically address the HCN to HNC abundance in the interstellar medium, with a combination of experimental techniques that allow for both isomers to be probed explicitly. The region that the spectrometer probes is indicated. The map is completed by a computational fluid dynamics simulation for the flow within the nozzle using Ansys Fluent, capped at a maximum density of 1 × 10 17 cm -3 (otherwise the much higher density in the throat of the nozzle would compress the scale of the whole map). c, Measurements of the T2 decay time as a function of time after the laser shot. The time axis is scaled to the distance from the nozzle exit in b, as under uniform conditions the molecules moving through the detection zone will be formed at a given position that can be correlated to time. The shaded areas around the T2 decay times represent the 95% confidence intervals for fitting the Voigt profile at each point along the uniform flow. Each T2 decay time results from fitting of an FID which is an average of 10 5 laser shots. The difference in T2 decay times for the two isomers can be clearly seen. [START_REF] Rohart | HCN absorption line shapes studied by millimeter wave coherent transients: speed dependent effects and collision interaction potential[END_REF] The new experimental data and scattering calculations display very good agreement over the temperature range 10-70 K. Both the species show a rising trend in pressure broadening cross sections with decreasing temperature, with the cross sections for HNC being higher than for HCN over this temperature range. This reflects a stronger collisional excitation of HNC than HCN by He (and by implication H2), especially at the low temperatures found in dense interstellar clouds. This strongly affects the observed HNC/HCN abundance ratio and helps explain anomalously high values that have previously been reported. 1. Conditions within the flows were precisely controlled using mass flow controllers (Brooks SLA5851S) and through pressure measurements using capacitance manometers (Brooks XacTorr CMX0). The temperature and density of the uniform flow from each de Laval nozzle were characterized by impact pressure measurements through the Rayleigh relations in isentropic flow. [START_REF] Dupeyrat | Design and testing of axisymmetric nozzles for ionmolecule reaction studies between 20 °K and 160[END_REF] HCN and HNC were prepared by pulsed excimer laser photolysis of vinyl cyanide (acrylonitrile, Sigma Aldrich >99% purity) at 193 nm (Coherent LPX Pro 210 F) at a 100 Hz repetition rate with a laser fluence in the observation region of typically 30 mJ cm -2 . This produces both molecules at relatively high abundance. [START_REF] Prozument | Photodissociation transition states characterized by chirped pulse millimeter wave spectroscopy[END_REF] Vinyl cyanide was entrained in the flow at low densities (∼0.1% of the total), via a Coriolis liquid flow controller (Bronkhorst mini CORI-FLOW ML120V00) and a Controlled Evaporative Mixer (CEM, Bronkhorst W-202A-222-K)

as described previously. [START_REF] Gupta | Low Temperature Kinetics of the Reaction Between Methanol and the CN Radical[END_REF] The laser photolysis products are rotationally cooled by collisions to low temperatures under the CRESU conditions, providing a cold sample to probe spectroscopically. The laser was aligned along the center of the flow and constrained in size to specifically interact with molecules only in the isentropic core and not the surrounding warmer boundary layers. [START_REF] Hearne | A novel Ka-band chirped-pulse spectrometer used in the determination of pressure broadening coefficients of astrochemical molecules[END_REF] The ground vibrational state j=1-0 transitions of HCN and HNC were probed in the CRESU flows using an E-band CP-FTmmW spectrometer which has been recently described in detail. [START_REF] Hays | Design and performance of an E-band chirped pulse spectrometer for kinetics applications: OCS -He pressure broadening[END_REF] Here, it was adapted for spectroscopy under CRESU conditions in a specially constructed vacuum chamber.

The spectrometer operates between 60-90 GHz but was able to extend the range to observe HNC. The rectangular feedhorns were mounted inside the vacuum chamber perpendicularly to the uniform flow, and situated 10 or 16 cm downstream from the exit of the de Laval nozzles to probe the uniform portion of the cold flow.

The arbitrary waveform generator (Keysight M8195A) and digitizer (Teledyne SP Devices ADQ7DC-PCIE) used in these experiments were the same as in previous works [START_REF] Hays | Design and performance of an E-band chirped pulse spectrometer for kinetics applications: OCS -He pressure broadening[END_REF][START_REF] Hearne | A novel Ka-band chirped-pulse spectrometer used in the determination of pressure broadening coefficients of astrochemical molecules[END_REF] but were reconfigured for time resolved acquisition following pulsed laser photolysis in the uniform flow. 400 µs pulse train sequences were used with 2.5 µs long pulse/collection cycles, resulting in 160 individual pulse -FID (Free Induction Decay) cycles per train, with the laser firing after the first half of each sequence. The first 80 cycles of each sequence were averaged together to produce an averaged background without any molecular signals. This background was subtracted from each cycle after the laser fired, allowing for the molecular FID to be differentiated from the reflections inherent in the system. When performing background subtractions, a slight phase shift [START_REF] Park | Design and evaluation of a pulsed-jet chirped-pulse millimeter-wave spectrometer for the 70-102 GHz region[END_REF] of each cycle was performed to increase rejection of reflections. When the SNR on a single FID was too low for fitting in the time domain, 2-4 sequential cycles were horizontally averaged before fitting, increasing the SNR of a scan but blurring the time resolution within the pulse sequence after the laser fired. This combines aspects of both segmented averaging [START_REF] Neill | Segmented chirped-pulse Fourier transform submillimeter spectroscopy for broadband gas analysis[END_REF] and FastFrame® 24 in a custom acquisition scheme. A total of 10 5 laser shot -pulse train sequences were averaged together at 100 Hz, taking about 15 minutes per experiment. At 70 K four of these experiments were averaged together due to lower population in the probed levels.

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FigureFig. 1 |

 1 Figure Legends/Captions (for main text figures)

Fig. 2 |

 2 Fig. 2 | Examples of fits to FIDs for HCN and HNC in He at 16 K. a, Time domain experimental data (black) and the fit (red) obtained using Supplementary Eq. 1 for HCN. The inset shows a zoomed version for 5 ns. b, FFTs of the data and the fit for HCN shown in panel a. c, Time domain experimental data (black) and the fit (blue) obtained for HNC using a time domain Voigt single frequency function as in Hays et al. 27 d, FFTs of the data and the fit for HNC shown in panel c. e, demodulated decay profiles of fitted data for HCN (red) and HNC (blue). Panels b and d show the FFT of the time domain fits which represent the FFTs of the original data well, highlighting the robustness of fitting the time-domain data. The demodulated decay profiles for each molecule shows the fitted exponential decays of the frequency components for the molecules. The decay times (T2), recorded under similar conditions, are clearly different, with the decay being faster for HNC than for HCN, reflecting the different He pressure broadening cross-sections. All data displayed result from the averaging of FIDs recorded from 10 5 laser shots.

Fig. 3 |

 3 Fig.3| Theoretical and experimental pressure broadening cross sections for HCN and HNC with He. Results from this study are compared to those from other experiments, including empirical fits to HC 15 N data. Error bars for the current experiments correspond to 95% confidence intervals from the averaging of the measured T2 decay times combined with contributions from estimated possible systematic errors in the flow density and temperature of 10% (see Supplementary Section 1A for further details). Error bars from other work correspond to 2σ when available. Previous data for HCN are from Ronningen and de Lucia 2005[START_REF] Ronningen | Helium induced pressure broadening and shifting of HCN hyperfine transitions between 1.3 and 20[END_REF] and Rohart and Kaghat 2010.[START_REF] Rohart | HCN absorption line shapes studied by millimeter wave coherent transients: speed dependent effects and collision interaction potential[END_REF] The new experimental data and scattering calculations display very good agreement over the temperature range 10-70 K. Both the species show a rising trend in pressure broadening cross sections with decreasing temperature, with the cross sections for HNC being higher than for HCN over this temperature range. This reflects a stronger collisional excitation of HNC than HCN by He (and by implication H2), especially at the low temperatures found in dense interstellar clouds. This strongly affects the observed HNC/HCN abundance ratio and helps explain anomalously high values that have previously been reported.

  10 17 cm -3 , each temperature condition having a specific density that varies by less than 10% over the probing region. Measured temperatures and densities are given in Supplementary Table

Acknowledgments

The authors would like to thank Jonathan Courbe, Jonathan Thiévin, Didier Biet, Ewen Gallou, and Alexandre Dapp for technical support, as well as André Canosa for the design of some of the Laval nozzles used in the project. The authors acknowledge funding from the European Union's Horizon 2020 research and innovation programme under the European Research Council (ERC) grant agreement 695724-CRESUCHIRP (I.R.S.) and under the Marie Skłodowska-Curie grant agreement 845165-MIRAGE (I.R.C. and I.R.S.). The authors are also grateful for support from the European Regional Development Fund, the Region of Brittany and Rennes Metropole. This work was supported by the French National Programme "Physique et Chimie du Milieu Interstellaire" (PCMI) of CNRS/INSU with INC/INP co-funded by CEA and CNES. ACCEPTED MANUSCRIPT / CLEAN COPY

Data availability

The source data for all the Figures in the main article and in the Supplementary Information, as well as the source data for the experimental and theoretical cross sections are publicly available in the Zenodo data repository 55 at https://doi.org/10.5281/zenodo.6256226

Code availability

The codes used to analyse the experimental data and generate the theoretical data presented in this study are publicly available in the the Zenodo data repository 55 at https://doi.org/10.5281/zenodo.6256226, or via the cited references.

Competing Interests Statement:

The authors declare no competing interests.

Methods

Experimental

He buffer gas (Air Liquide 99.995%) is flowed through convergent-divergent de Laval nozzles, which were aerodynamically shaped to form cold supersonic uniform flows. Four de Laval nozzles were used to produce five temperatures around 10, 16, 19, 30, and 70 K in uniform flows with densities varying between 0.2-1 ×

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Theoretical

The PESs were calculated in the supermolecular approach by means of the explicit correlated coupled cluster method including single, double, and perturbative triple excitations [CCSD(T)-F12a] [START_REF] Knizia | Simplified CCSD(T)-F12 methods: Theory and benchmarks[END_REF] with an augmented correlation-consistent polarized valence triple-ζ [aug-cc-pVTZ] basis set [START_REF] Kendall | Electron-Affinities of the 1st-Row Atoms Revisited -Systematic Basis-Sets and Wave-Functions[END_REF] (hereafter CCSD(T)-F12/aug-cc-pVTZ).

After fitting of the interaction energy, close coupling scattering calculations were performed on these PESs to find the pressure broadening cross sections. More details on these calculations can be found in Supplementary section 2 'Theoretical calculations'.