Relation of Triglyceride Levels, Fasting and Nonfasting, to Fatal and Nonfatal Coronary Heart Disease FREE

THE ROLE OF blood triglyceride (TG) levels in predicting coronary heart disease (CHD) independently of established major risk factors (total cholesterol level, low-density lipoprotein cholesterol (LDL-C) level, blood pressure, and smoking) remains unclear. In addition, it remains to be determined whether fasting or nonfasting levels are more informative for CHD risk. A National Institutes of Health Consensus Conference in 1993 on these issues concluded the following: TG levels are weak predictors at best for CHD when other risk factors are taken into account, and, on the basis of scant available data, nonfasting TG levels may be more important than fasting TG levels.¹

Research since then has not further clarified these issues. Assessment of any clinical association of TG levels with CHD is difficult because of high intraindividual variability in TG levels, which can result in sizable regression dilution,² skewness of population TG distributions, and high correlation of TG levels with those of other lipids, especially the inverse correlation with high-density lipoprotein cholesterol (HDL-C) levels. Although nearly all studies have shown that TG levels are a significant univariate predictor of CHD, only a few have shown that TG levels are predictive of CHD after adjustment for LDL-C and HDL-C levels and other risk factors. The Adult Treatment Panel III³ recently recommended treatment strategies for CHD prevention for people with high (200-499 mg/dL [2.26-5.63 mmol/L]) or very high (≥500 mg/dL [≥5.64 mmol/L]) fasting TG levels.

Previous work on the Multiple Risk Factor Intervention Trial (MRFIT) data⁴ used the control (no intervention) group to examine whether fasting TG levels are a significant predictor of fatal or nonfatal myocardial infarction independent of the ratio of fasting total cholesterol to HDL-C level. Since no study has compared the predictive values of fasting vs nonfasting TG levels in the same cohort, the National Institutes of Health Consensus Panel question on their comparative value in assessing CHD risk remains unanswered. This work explores that question and quantifies the association of fasting and nonfasting TG levels with incidence of nonfatal plus fatal CHD and with CHD mortality after adjustment for major CHD risk factors and regression dilution.

The design and methods of the MRFIT have been detailed.^5- 7 Briefly, the MRFIT was a randomized primary prevention trial of CHD mortality among men aged 35 to 57 years at increased risk but without clinical evidence of definite CHD at baseline. Intervention consisted of dietary advice to lower blood cholesterol level, smoking cessation counseling, and stepped-care hypertension medication. During 1973 through 1975, 22 clinical centers in 18 US cities screened 361 662 men for eligibility.⁷ After 3 screening visits, 12 866 men were randomized into the trial. Of these, 2863 (from 17 clinical centers) had TG levels measured along with total serum cholesterol from a serum sample taken at the initial screen (screen 1) with no fasting required; we refer to this as the "nonfasting" TG value, even though time since the last meal was not recorded. At screen 2, all participants had TG levels and other lipid fractions measured from a fasting (≥12 hours) plasma sample. This study concerns the 2863 men with both nonfasting and fasting TG levels measured at screens 1 and 2, respectively.

Both fasting and nonfasting TG levels were determined by the central MRFIT laboratory. Nonfasting TG level was determined from a serum sample taken at screen 1 with no fasting requirement. Fasting total cholesterol, TG, HDL-C, and LDL-C levels were determined from a plasma sample taken at screen 2,⁸ on average 44 days after screen 1. Serum levels of uric acid and glucose were also determined at screen 2. Systolic blood pressure is defined as the average of 4 random-zero manometer readings, ie, readings 2 and 3 at each of screen 2 and screen 3 (the randomization visit, on average 19 days after screen 2). Height (inches), weight (pounds), self-reported alcohol use (average number of alcoholic drinks per week), years of education, marital status, working status, and approximate total family income were recorded at screen 3. Race and self-reported current smoking level (average number of cigarettes smoked per day) were recorded at screen 1.

Baseline for computing time to CHD death was the randomization date. Mortality during the active intervention phase of MRFIT (through February 28, 1982) was verified by clinical staff and coded by means of the International Classification of Diseases, Ninth Revision (ICD-9).^7,9 Posttrial mortality through December 31, 1990, was determined by matching identifying information, provided by each participant during screening, with National Death Index records.^10- 12 Death certificates were obtained to ascertain underlying cause of death and were coded independently by 2 nosologists; a third nosologist adjudicated any disagreements. Death dates and corresponding ICD-9 or ICD-10¹³ causes from January 1, 1991, through December 31, 1999, were obtained by means of the National Death Index–Plus service. Coronary heart disease death was defined by ICD-9 codes 410 to 414 or 429.2 and by ICD-10 codes I-20 to I-25. Nonfatal CHD was defined as either a clinical myocardial infarction or a significant serial electrocardiogram change indicative of myocardial infarction,¹⁴ with follow-up through trial end (February 28, 1982).

Baseline characteristics were summarized separately for those with high or very high TG level (≥200 mg/dL [2.26 mmol/L]) on the basis of Adult Treatment Panel III guidelines³ vs those with lower levels. Univariate regression analysis of each of fasting and nonfasting TG levels on each characteristic was used to examine correlates of TG levels. Cumulative event probability curves for time to CHD death and for time to nonfatal or fatal CHD were calculated for those with TG level of 200 mg/dL or more and compared with those with TG levels less than 200 mg/dL for fasting and for nonfasting levels separately. Univariate and multivariate proportional hazards regression models¹⁵ were carried out for CHD death, with stratification by clinical center. Triglyceride levels were examined in separate models for fasting and nonfasting. Triglyceride levels were highly skewed to the right; therefore, models considered both natural log–transformed TG levels and dichotomized levels: 200 or more vs less than 200 mg/dL. Potential confounders considered in multivariate models were randomized treatment group, age, body mass index (calculated as weight in kilograms divided by the square of height in meters), race (African American or not), education (years), marital status (married or not), family income, part-time worker, alcoholic drinks per week, cigarettes smoked per day, systolic blood pressure, and fasting HDL-C, LDL-C, glucose, and uric acid levels; prespecified interactions were tested. We verified the proportional hazards regression assumption for each predictor with the test of Grambsch and Therneau.¹⁶ Similar proportional hazards models were considered for the combined nonfatal plus fatal CHD end point. The P values shown are 2-tailed, without adjustment for multiple comparisons.

Effects of regression dilution were assessed on the basis of data for all MRFIT participants who had annual visit 2 fasting TG levels and baseline risk factor values (N = 11 634). Following the methods of Clarke et al,² we derived a "nonparametric" regression dilution factor by (1) grouping all observations into screen 2 fasting TG quintiles; (2) computing within-group means of screen 2 fasting TGs and within-group means of annual visit 2 fasting TG levels; and (3) computing ratios of group-group differences in these means. A "parametric" regression dilution factor was computed as the inverse of the correlation between the screen 2 and annual visit 2 fasting TG levels. We then estimated the true TG–CHD mortality association by inflating our TG regression coefficients by each of the estimated regression dilution factors.

Of the 2863 men with both TG values, 54 had missing baseline data (46 for glucose, an additional 6 for HDL-C, and 2 for alcohol use) and were excluded. The remaining 2809 are the basis for our analysis. Mean TG levels were 186.7 mg/dL (2.11 mmol/L) (SD, 133.5 mg/dL [1.51 mmol/L]) for fasting and 284.3 mg/dL (3.21 mmol/L) (SD, 193.3 mg/dL [2.18 mmol/L]) for nonfasting levels. Of the 2809 men, 874 (31%) had fasting TG levels of 200 mg/dL or more and 1724 (61%) had nonfasting TG levels of 200 mg/dL or more. The percentages with TG level of 500 mg/dL or more were 3% with fasting levels and 11% with nonfasting levels. Baseline characteristics for these men are shown in Table 1 by TG level for fasting and nonfasting conditions. The associations of each baseline characteristic with fasting TG level were similar to those of each characteristic with nonfasting TG level, except for age, alcohol use, and smoking. Age was negatively associated, and alcohol and smoking were positively associated, with natural log fasting TG level (P<.001, <.001, and .007, respectively), but none of the three was significantly associated with natural log nonfasting TG level (P = .42, .82, and .68, respectively). The Pearson correlation coefficient for fasting-nonfasting TG level was 0.67 on the original scale and 0.65 on the natural log scale.

Median follow-up from randomization through December 31, 1999, was 25.4 years, with 328 CHD deaths. Among the 874 men with fasting TG levels of 200 mg/dL or more, there were 117 CHD deaths (13.4%); for those with fasting TG levels less than 200 mg/dL, there were 211 CHD deaths (10.9%). Among the 1724 men with nonfasting TG levels of 200 mg/dL or more, there were 226 CHD deaths (13.1%); for the 1085 men with nonfasting TG levels less than 200 mg/dL, 102 died of CHD (9.4%). Plots of cumulative mortality for those with TG levels of 200 mg/dL or more and less than 200 mg/dL (Figure 1 and Figure 2) showed an increasing separation in risk from about year 7 for fasting and year 10 for nonfasting.

Results from univariate- and multivariate-adjusted proportional hazards regression models are shown in Table 2. Both fasting and nonfasting TG levels were significant predictors of CHD mortality in unadjusted and in multivariate-adjusted (excluding HDL-C level) analyses, where hazard ratios (HRs) were similar for fasting and nonfasting TG levels. The strength of the relationships decreased with adjustment for HDL-C level, but HRs for 200 mg/dL or more compared with less than 200 mg/dL were still similar for fasting (HR, 1.24; P = .09) and nonfasting (HR, 1.26; P = .07) TG levels.

In models with TG levels entered on a natural log (ln) scale, both fasting and nonfasting TG levels were again equally strong unadjusted and multivariate-adjusted (excluding HDL-C level) predictors of CHD mortality, and were weaker when further adjusted for HDL-C level. Multivariate-adjusted (excluding HDL-C level) HRs for CHD mortality for a 1-ln higher TG level (eg, 250 vs 92 mg/dL) were 1.43 (P = .001) for fasting and 1.37 (P = .002) for nonfasting TG level. After adjustment for all confounding variables including HDL-C level, HRs dropped to 1.24 (P = .09) for fasting and 1.23 (P = .06) for nonfasting TG level. Education, income, employment status, and marital status were not significant predictors at level .05 in any of these models and were dropped. The interactions of TG level with HDL-C level, LDL-C level, systolic blood pressure, and number of cigarettes smoked per day were also considered and were not statistically significant.

There were 175 combined nonfatal or fatal CHD events, with median follow-up through February 28, 1982, of 7.6 years. Among the 874 men with fasting TG levels of 200 mg/dL or more, there were 73 nonfatal or fatal CHD events (8.4%); for the 1935 with fasting TG level less than 200 mg/dL, 5.3% had a nonfatal or fatal CHD event. Among the 1724 men with nonfasting TG levels of 200 mg/dL or more, there were 125 nonfatal or fatal CHD events (7.3%); for the 1085 men with nonfasting TG level less than 200 mg/dL, 50 (4.6%) experienced a nonfatal or fatal CHD event. Plots of cumulative mortality for those with fasting and nonfasting TG levels of 200 mg/dL or more and less than 200 mg/dL showed an increasing separation in risk from about year 1 (not shown).

Results from univariate and multivariate analyses are summarized in Table 3. Like the results for CHD mortality, both fasting and nonfasting TG levels were significant predictors of nonfatal or fatal CHD in unadjusted and multivariate-adjusted (excluding HDL-C level) analyses. Associations of TG level of 200 mg/dL or more with nonfatal or fatal CHD were also significant after adjustment for HDL-C level: the HR was 1.64 (P = .004) for fasting and 1.46 (P = .03) for nonfasting. The multivariate-adjusted HR for a 1-ln higher TG level was 1.56 (P = .008) for fasting TG level and 1.29 (P = .09) for nonfasting TG level. In general, HRs were larger for fatal or nonfatal CHD in 8 years than for fatal CHD in 25 years for both fasting and nonfasting TG levels. Education, income, employment status, and marital status were not significant predictors at level .05 in any of these models and were dropped.

To assess whether the apparently stronger relationship of TG with nonfatal or fatal CHD in 8 years compared with CHD death in 25 years was due to the more proximal measurement of TG to the event, the 25-year follow-up period was divided into 3 periods with approximately equal numbers of CHD deaths: less than 12, 12 to 17, and 18 or more years after randomization, with 98, 103, and 127 deaths, respectively. Multivariate-adjusted HRs for CHD mortality in the first 12 years associated with a TG level of 200 mg/dL or more were 1.45 (95% confidence interval [CI], 0.92-2.28; P = .11) for fasting TG level and 1.08 (95% CI, 0.69-1.69; P = .73) for nonfasting TG level. In contrast, HRs for deaths after 18 years were 0.93 (95% CI, 0.61-1.41) for fasting TG level and 1.60 (95% CI, 1.05-2.45) for nonfasting TG level, indicating that nonfasting TG level may be more predictive of long-term mortality than fasting TG level. The HRs for deaths during 12 to 17 years were 1.51 (95% CI, 0.98-2.33) for fasting TG level and 1.14 (95% CI, 0.74-1.77) for nonfasting TG level. P values corresponding to the interaction of natural log–transformed TG level and follow-up time were .99 for fasting TG level and .78 for nonfasting TG level. Taking these results together, then, there is no strong evidence that TG levels are more predictive of early (more proximal) CHD deaths than later deaths.

Analyses were also carried out classifying participants by both fasting and nonfasting TG levels simultaneously (Table 4). Most of the men with nonfasting TG level of 500 mg/dL or more also had a fasting TG level of 200 mg/dL or more (81%), and these men were at significantly increased risk of CHD death compared with men with both fasting and nonfasting TG levels less than 200 mg/dL (HR, 1.59; P = .03). There were only 62 men with nonfasting TG level of 500 mg/dL or more and fasting TG level less than 200 mg/dL (19% of those with nonfasting TG level of 500 mg/dL or more and 2% overall), among whom 5 CHD deaths occurred. Half of the men (1404 [50%]) had nonfasting TG levels of 200 to 499 mg/dL (2.26-5.63 mmol/L). For this subgroup, HRs were similar for those with high or with low fasting TG level: with fasting TG level of 200 mg/dL or more, the HR (relative to men with both fasting and nonfasting TG level <200 mg/dL) was 1.35 (P = .08), and for men with fasting TG level less than 200 mg/dL, the HR was 1.25 (P = .12). This group of men with nonfasting TG level between 200 and 499 mg/dL was at higher risk of CHD death regardless of their fasting TG level.

These associations of TG levels with CHD mortality and incidence are likely underestimated because of regression dilution. We first estimated the regression to the mean effect in the larger MRFIT cohort (N = 11 634) "nonparametrically" by computing mean screen 2 fasting TG levels and mean annual visit 2 fasting TG level within screen 2 quintiles. Ratios (screen 2 to annual visit 2) of quintile-quintile differences in means ranged from 1.28 to 1.56 (Table 5). This indicated that an estimated regression coefficient for TG should be multiplied by up to 1.56 (inflated by up to 56%) to estimate the magnitude of the true association. The "parametric" estimate (inverse of the correlation coefficient between screen 2 fasting TG level and annual visit 2 fasting TG) was 1.00/0.64 = 1.56, indicating again that TG regression coefficients should be inflated by up to 56%.

This regression to the mean effect in TGs consequently affects the estimated levels of risk for CHD incidence and death, a phenomenon known as regression dilution bias.² The problem is possibly compounded by the selection (from the men screened for MRFIT) of high-risk men—according to high cholesterol level—for randomization into the MRFIT cohort. The estimated multivariate regression coefficient for a 1-ln higher fasting TG level was 0.213 (corresponding to a HR of 1.24; Table 3), so that the regression dilution–adjusted HR was exp(0.213×1.56) = 1.39. Assuming the same regression dilution effect for nonfasting TG levels gives us similar results from the nonfasting TG regression coefficient of 0.207: exp(0.207×1.56) = 1.38.

The estimated risk of fatal CHD during 25 years was similar for fasting TG level of 200 mg/dL or more vs less than 200 mg/dL when compared with the risk of nonfasting TG level of 200 mg/dL or more vs less than 200 mg/dL. This was true as well as for the risk associated with a 1-ln higher fasting TG level compared with nonfasting TG level. Associations of TG level with nonfatal or fatal CHD in 8 years were also similar for fasting and nonfasting TG levels and on both the dichotomized and the natural log scales. These associations tended to be stronger than for fatal CHD in 25 years, particularly for fasting TG level. However, there were no consistent trends indicating that the association with TG level, fasting or nonfasting, varied during follow-up; thus, the differences observed between these two end points could result from sampling variability. The estimated attenuation in the associations (with CHD mortality in particular) was substantial because of regression dilution. This attenuation is likely greater for nonfasting than fasting TG level because the intraindividual variability in nonfasting TG measurements would be expected to be higher.

On the basis of Adult Treatment Panel III classification, the prevalence of hypertriglyceridemia (TG level ≥200 mg/dL) was double when nonfasting TG levels were used compared with the use of fasting TG levels (fasting prevalence, 31%; nonfasting prevalence, 61%), yet both were associated with a similar increase in risk of CHD. These two observations indicate that nonfasting TG may be useful for risk stratification. This was further supported by bivariate analyses of fasting and nonfasting TG, which, although limited by the number of events and therefore reduced power, suggested that elevated nonfasting TG may be associated with increased risk of CHD even among men with fasting TG level less than 200 mg/dL. Use of a nonfasting TG level may be particularly appropriate for clinical trials where collection of a fasting blood sample is infeasible.

No prospective studies, to our knowledge, have compared in the same cohort the strength of the association of fasting and nonfasting TG level with CHD risk. In a small case-control study, Patsch et al¹⁷ found a stronger association of nonfasting than fasting TG level with coronary artery disease based on angiography. Comparisons of results across studies is hazardous because of variation in event definition and differential control of confounding factors. However, in a meta-analysis¹⁸ of 17 prospective studies, an 89-mg/dL (1-mmol/L) higher fasting TG level was associated with a 14% increased risk of cardiovascular disease after adjustment for HDL-C level and other risk factors. By comparison, in recent prospective studies of nonfasting TG level, stronger associations with CHD were noted. Stampfer et al¹⁹ reported a 40% higher risk of myocardial infarction associated with a 100-mg/dL (1.13-mmol/L) higher nonfasting TG level. Iso et al²⁰ reported a 26% (P = .09) increased risk of CHD associated with an 89-mg/dL (1-mmol/L) higher nonfasting TG level. Schaefer et al²¹ examined fasting plasma levels of TGs and remnantlike particles TG and their associations with carotid artery stenosis by ultrasound in the Framingham Offspring Study. No multivariate associations were found in men for TGs or remnantlike particles TG, whereas strong multivariate associations were found in women (P<.001 for TGs and remnantlike particles TG).

The possible mechanisms by which TGs could affect CHD have been reviewed.^22- 24 Ginsberg²³ noted that much of the evidence about TG and cardiovascular disease comes from studies in which TG level was measured in participants who were fasting. He noted that postprandial TG level and chylomicron remnants are predictive of cardiovascular disease^25- 26 and speculated that the role of TG in atherogenesis might be more readily demonstrated with the use of nonfasting blood levels. Roche and Gibney²⁷ and Yu and Cooper²⁸ detailed the metabolic pathways for processing postprandial lipoproteins. Yu and Cooper described the lipolyzation of chylomicron TGs, allowing the delivery of free fatty acids to peripheral tissues. They described the roles of liver and liver receptors, the removal of chylomicron remnants from the blood, and the potential detrimental effects of chylomicron remnants on arterial walls.

A limitation of our study is that we did not determine time since last meal for the nonfasting TG measurement. Furthermore, some of the differences between fasting and nonfasting TG levels could be due to serum vs plasma differences, and/or to regression to the mean (men at screen 1 were selected for further screening on the basis of elevated total serum cholesterol level, which is positively correlated with TG levels). In our cohort of 2809, these screen 1 serum nonfasting values were on average 25% higher than screen 2 plasma fasting values. Previous work in the entire MRFIT cohort (12 866 men) showed on average 4% higher serum than plasma TG levels from the same fasting blood sample.²⁹ It is a reasonable inference that further differences were predominantly due to nonfasting vs fasting state, even though we had no data on time of last meal for nonfasting samples.

In conclusion, this work is, to our knowledge, the first direct comparison—from the same study cohort—of the prognostic importance of fasting and nonfasting TG levels for fatal and nonfatal CHD. Fasting and nonfasting TG levels were equivalently predictive both of 25-year CHD mortality and of 8-year fatal or nonfatal CHD on an unadjusted or multivariate-adjusted basis; adjustment for regression dilution made the predictiveness stronger. Taken together with the higher prevalence of hypertriglyceridemia with the use of nonfasting TG level, our results indicate that nonfasting TG level may be useful for risk stratification. Although multiple measurements may be required for accurate classification, nonfasting readings may provide information about risk that is not apparent from fasting readings alone.

Corresponding author and reprints: Lynn E. Eberly, PhD, Division of Biostatistics, School of Public Health, University of Minnesota, 420 Delaware St SE, MMC 303, Minneapolis, MN 55455-0378 (e-mail: lynn@biostat.umn.edu).