Introduction. AE Black 1 * {

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1 (2000) 24, 1119±1130 ß 2000 Macmillan Publishers Ltd All rights reserved 0307±0565/00 $ Critical evaluation of energy intake using the Goldberg cut-off for energy intake:basal metabolic rate. A practical guide to its calculation, use and limitations 1 * { 1 MRC Dunn Nutrition Centre, Downham's Lane, Milton Road, Cambridge CB4 1XJ, UK OBJECTIVES: To re-state the principles underlying the Goldberg cut-off for identifying under-reporters of energy intake, re-examine the physiological principles and update the values to be substituted into the equation for calculating the cut-off, and to examine its use and limitations. RESULTS: New values are suggested for each element of the Goldberg equation. The physical activity level (PAL) for comparison with energy intake: basal metabolic rate (EI:BMR) should be selected to re ect the population under study; the PAL value of 1.55 x BMR is not necessarily the value of choice. The suggested value for average withinsubject variation in energy intake is 23% (unchanged), but other sources of variation are increased in the light of new data. For within-subject variation in measured and estimated BMR, 4% and 8.5% respectively are suggested (previously 2.5% and 8%), and for total between-subject variation in PAL, the suggested value is 15% (previously 12.5%). The effect of these changes is to widen the con dence limits and reduce the sensitivity of the cut-off. CONCLUSIONS: The Goldberg cut-off can be used to evaluate the mean population bias in reported energy intake, but information on the activity or lifestyle of the population is needed to choose a suitable PAL energy requirement for comparison. Sensitivity for identifying under-reporters at the individual level is limited. In epidemiological studies information on home, leisure and occupational activity is essential in order to assign subjects to low, medium or high PAL levels before calculating the cut-offs. In small studies, it is desirable to measure energy expenditure, or to calculate individual energy requirements, and to compare energy intake directly with energy expenditure. (2000) 24, 1119±1130 Keywords: diet record; energy intake; validity Introduction Over the decades from the thirties to the seventies, the `validity' of dietary assessment techniques has been tested by comparing one method with another, but without any means of testing which, if any, method was valid. In the eighties and nineties, the advent of biomarkers of intake, including urinary nitrogen excretion and doubly labelled water measures of energy expenditure, uncovered the frequent bias to the under-estimation of food intake. It is now generally acknowledged that mis-reporting of intake is a serious problem in studies of nutrition and health. The phenomenon under investigation is the marked bias to the under-estimation of food intake when the information is derived from self-reported dietary assessments. The word that has been, and still is, widely used for this phenomenon is `under-reporting'. *Correspondence:, c=o MRC Human Nutrition Research Downhams Lane, Milton Road, Cambridge CB4 1XJ, UK. {Present address: 9 Birch Close, Cambridge, CB4 1XN, UK. Received 15 December 1999; revised 2 May 2000; accepted 10 May 2000 This unfortunately carries connotations of causality, suggesting it is due only to conscious or unconscious omission of food items from the report. However, under-eating (dieting) is also a signi cant contributor to the systematic bias 1±5 and over-reporting=overeating has also been demonstrated. 6,7 The phenomenon is therefore more accurately described as `providing diet reports of poor validity'. This is a cumbersome phrase and therefore the word `underreporting' is still used in this paper, but should not be taken to imply causality. Poor validity of dietary assessments has implications for the interpretation of studies of diet and health. Under-estimation of nutrient intake will in ate estimates of the incidence of inadequate intake. 8 Between-subject differences in reporting, whether quantitative or qualitative, will create variable bias that may distort the relationships between nutrient intake and health. That conclusions can be altered by the inclusion or exclusion of `low energy reporters' has been demonstrated. 9,10 A key problem is how to identify those individuals who have provided data of poor validity. Until that

2 1120 can be done, the characteristics of the bias cannot be fully explored and it is not possible to determine how to minimize the effects in the analysis of dietary data. A number of authors have used the Goldberg cut-off 11 to identify `low energy reporters' and to explore their characteristics or to examine the effects of low energy reporting on the data and the conclusions to be drawn from it. However, the technique has not always been fully understood or correctly applied. The papers by Goldberg, Black and colleagues were not written as a practical guide to a tool for investigating underreporting. Thus the information needed to apply the concept was not easy to extract, and the key statistics for the derivation of the equation were relegated to the Appendix. Misinterpretations of the concepts have included using a cut-off calculated for mean intake to identify individual under-reporters, confusing the cut-off for `habitual' intake with that for low intake obtained by chance, and interpreting the illustrative tables of cut-off values based on a physical activity level (PAL) value of 1.55 as recommendations for universal application. The present paper re-states the principles underlying the Goldberg cut-off, re-examines each parameter in the equation and the values to be inserted, and provides guidance for its application and comments on its usefulness and its limitations. The principles of the Goldberg cut-off The validation of reported energy intake (EI rep ) rests on the fundamental equation Energy intake EI ˆenergy expenditure EE changes in body stores and the assumption that at the group level changes in body stores can be ignored, and therefore EI ˆ EE 1 Good agreement between EI and EE when weight is stable has been demonstrated in an experimental situation. 12 Thirty-two healthy subjects were fed on a liquid formula until weight had been stable for 14 days. At stable weight EI was kcal=day and doubly labelled water (DLW) EE was kcal=day. Agreement in a natural situation was provided by two studies in which EI was observed by researchers and EE measured by DLW. The mean difference between EI and EE in 14 hospitalized elderly women was 1% and in 10 men eating ad libitum in a metabolic facility it was 5%. 13 ± 15 The absolute energy requirements of individuals vary with age, sex and body size, but, since the basal metabolic rate (BMR) also varies with age, sex and body size, energy requirements may be stated in general terms as multiples of BMR. 16 This expression of energy requirement (EE:BMR) has also come to be known as the physical activity level (PAL).* Thus Eq. (1) may rewritten as EI:BMR ˆ PAL and the mean reported energy intake (EI rep ) in any study may be expressed as EI rep :BMR and compared with the expected PAL for that population. However, absolute agreement cannot be expected since there are errors in the measurement of all elements of this equation. The mean EI rep :BMR could derive from valid diet assessments but still be less than (or greater than) PAL. The con dence limits of the agreement between EI rep :BMR and PAL must be determined. The equation for calculating the con dence limits (cut-off) is given below. The full statistical derivation was given in the original paper. 11 In that paper, since the energy expenditures in the studies reviewed were unknown, only the lower con dence limit below which the reported mean energy intake was unlikely to represent valid data was calculated. It was calculated assuming a sedentary lifestyle. However, if the activity of a population is known such that it is possible to assign an appropriate mean PAL, then the upper con dence limit may also be calculated. To determine whether a given mean value of EI rep :BMR in n subjects is acceptable, the following must be satis ed: EI rep :BMR > PAL exp s:d: min S=100 p n EI rep :BMR < PAL exp s:d: max S=100 p n where PAL is the mean PAL for the population under study, s.d. min is 2 for the 95% or 3 for the 99% lower con dence limit, s.d. max is 2 for the 95% or 3 for the 99% upper con dence limit, and n is the number of subjects in the study. S is the factor that takes account of the variation in intake, BMR and energy requirements, and is given by r CV 2 wei S ˆ CV 2 wb d CV2 tp where CV wei is the within-subject coef cient of variation in energy intake, d is the number of days *EE: BMR is also called average daily metabolic rate (ADMR).

3 of diet assessment, CV wb is the coef cient of variation of repeated BMR measurements or the precision of estimated compared with measured BMR, and CV tp is the total variation in PAL. CV tp is the coef cient of variation derived from the mean and standard deviation of a study and includes true between-subject variation, an element of within-subject variation and methodological errors. In the original paper 11 this term was designated CV b (between-subject variation), but it has been changed to CV t in the present paper to conform to the nomenclature used in a subsequent paper examining within- and between subject variation in repeat DLW measurements. 17 This equation calculates the con dence limits (cutoffs) that determine whether the mean reported energy intake is plausible as a valid measure of food intake even if chance has produced a dataset with a high proportion of days of genuinely low (or high) intake. This was termed CUT-OFF 2 in the original paper. In order to maximize the sensitivity and speci city of the cut-off, 7 each element of the equation should be substituted with values appropriate for the speci c study being evaluated. Factors for substitution into the Goldberg equation Energy expenditure of the population under study: the PAL In the original paper that examined the reported energy intake from 37 different studies, 18 EI rep :BMR was compared with a PAL value of 1.55, the WHO value for `light' activity. 16 The reasons for choosing this value were several. First, the energy expenditure or physical activity in the studies examined was unknown; second, choosing too high a value might have exaggerated the extent of under-reporting; third, the WHO recommendations for energy requirements 16 were well known and widely used; and fourth, calorimetry and early DLW work con rmed the value of 1.55BMR for light activity as a probable minimum energy requirement for a normally active but sedentary population (not sick, disabled or frail elderly). 11 A subsequent review of 74 DLW studies comprising 574 free-living subjects 19 indicated that 1.55 was a conservative value and that many groups have higher energy expenditures. That review suggested the following broad categories: `The relationship between lifestyle, activity and PAL suggested by the data can be summarised as follows: Chair-bound or bed-bound 1.2 Seated work with no option of moving around and little or no strenuous leisure activity 1.4 ± 1.5 Seated work with discretion and requirement to move around but little or no strenuous leisure activity 1.6 ± 1.7 Standing work (eg housewife, shop assistant) 1.8 ± 1.9 Signi cant amounts of sport or strenuous leisure activity(30±60min4±5timesperweek) 0.3 Strenuous work or highly active leisure 2.0 ± 2.4 The data provide little evidence to quantify the energy cost of manual occupations, but the range 2.0 ± 2.4 is suggested as the maximum for a sustainable lifestyle.' Other PAL values for broad categories are given by the FAO=WHO=UNU Expert Consultation on Energy and Protein Requirement 16 and UK report on Dietary Reference Values. 20 The WHO report also includes tables showing the factorial derivation of the PAL values. The DLW data from 574 free-living subjects 19 were also used to derive mean energy expenditures by 1121 Table 1 Energy expenditure in different age and sex groups Age (y) Tee (MJ=day) BMR (MJ=day) PAL Age group (y) n Mean (s.d.) Mean s.d. CVt Mean s.d. CVt Mean s.d. CVt Females 1 ± ± ± ± ± ± ± Males 1 ± ± ± ± ± ± ± Adapted from Black et al,

4 1122 age and sex. These are shown in Table 1. However, it should be noted that these subjects were predominantly white collar in origin and very few were identi ed as having manual occupations. The higher PAL for the age group 18 ± 29 probably re ected more active leisure pursuits, and the lower values in the older individuals, less active leisure. Values for speci c groups in the population could be obtained from individual DLW studies. An appendix to the above paper 19 listed all 74 studies with details of the subjects and the measured energy expenditures expressed both as mega joules and as PAL. However, such gures should be used with caution since most DLW studies have been conducted on small numbers of selected subjects. There exists therefore a body of data to suggest suitable PAL values for various population groups. However, knowledge of the activities of the population is required, suggesting that a questionnaire on home, leisure and occupational activity should be included routinely in all dietary surveys. Between-subject variation in physical activity (CV tp ) CV tp is the coef cient of variation of total variation in energy requirement expenditure calculated from the mean and standard deviation of a study. This gure, although referred to as `between-subject' variation and notated as CV b in the original paper, 11 includes within- (CV w ) and between- (CV b ) subject variation and methodological errors. This is the value to be used when the group mean EI rep :BMR is being evaluated. It is also the value to use when individual EI rep :BMR are being evaluated, since the EI rep :BMR is being evaluated against a population mean that includes both between and within-subject variation. The value of 12.5% used in the original paper was taken from the FAO=WHO=UNU report on energy and protein requirements 16 (Section 4.5, p 44), but only one study 21 was cited to support this value. The text stated `...the measurement of total energy expenditure over a week indicates that the inter-individual variability of expenditure, in a speci ed group, has a coef cient of variation (CV) of about 12.5% on a body weight basis. 21 ' Accumulated data from doubly labelled water studies (Figure 1) showed a wide range of PAL, from about 1.2 to over 2.0, at all ages. 19 The CV tp from these data are shown in Table 1. They ranged from 9.5 to 23.8% in different age ± sex groups with a pooled mean of 15.4%. Examination of studies with repeat DLW measurements found the values to be little reduced when CV tp was calculated using subject means derived from more than one DLW measurement. 17 A round gure of 15% is suggested as a suitable average value to substitute into the Goldberg equation, but other values may be used if deemed more appropriate for any given study. Within-subject daily variation in energy intake (CV wel ) The within-subject day to day variation in food intake (CV wei ) of individuals is large. Figure 2 shows the energy intake of one individual who maintained a weighed diet record every sixth day for 1 y. 22 The zero line indicates the mean intake averaged over the year, ie the `habitual' intake. Average intakes from 1, 3 or 7 successive days of recording are plotted as the differences from zero. The ne dotted line shows the daily differences and demonstrates the enormous variability from day to day. The 3 day and 7 day average differences show the values obtained from alternative short-term measurements. In the year-long study of dietitians from which these data were taken, 23 the within-subject variation in daily energy intake ranged from 10 to 50% in individuals, with a pooled mean of 26%. 24 Table 2 show the CV w from other studies reviewed by Bingham 25 and Nelson and colleagues. 26 These ranged from 14 to 45% with a pooled mean of 23%. Thus 23% is suggested as a suitable average value to substitute into the Goldberg equation, but a speci c Figure 1 Distribution of individual PAL values from 574 freeliving individuals in whom energy expenditure was measured by doubly labelled water. (From Black et al, ) Figure 2 Daily energy expenditure of one individual undertaking a weighed diet record every sixth day for 1 y, illustrating the variation obtained from 1, 3 and 7 d averages. Mean intakes for 1, 3 and 7 successive days of recording are shown as the difference between that intake and the year-long average (habitual) intake. (From Black, )

5 Table 2 (A) Adapted from Bingham Pooled within-subject variation (CV w ) in energy intake (B) Adapted from Nelson et al, Reference CV w (%) Subject group CV w (%) Men Women Pregnant women Elderly men Elderly women Children 1 ± 4 y, male Children 1 ± 4 y, female (men) Children 5 ± 10 y, male (women) Children 5 ± 10 y, female Children 11 ± 17 y, male Children 11 ± 17 y, female Children 13 ± 15 y, male (women) Children 13 ± 15 y, female (men) Callmer 20 (unpublished) Pooled mean CV w 23 Pooled mean CV w 23 CV wei may be calculated for any given study according to the formula s X n Pooled mean CV w ˆ CV 2 i =n where CV i is the CV calculated for each subject from the number of days of dietary assessment available for that subject, and n is the number of subjects. Variation in basal metabolic rate (CV wb ) Since energy requirement and intake are expressed as multiples of the BMR, between-subject variation is taken into account and only within-subject variation need be considered in calculating the con dence iˆ1 limits. This includes both measurement error and variation with time on repeated BMR measurements. The classical BMR measurement is made at least 12 h after the last meal, after an overnight sleep in the place of measurement, completely at rest and at thermoneutral temperature (about 23 C). However, practicalities often dictate that these conditions are not fully met, in which case the measurement is called resting metabolic rate (RMR). Either the subject has to walk from the place of sleep to the place of measurement (in-patient RMR) or sleeps at home and travels fasted to the place of measurement in the morning (outpatient RMR), in both of which situations the measurement is done after a rest period of usually 30 ± 60 min. For the out-patient RMR, researchers do not have control over the size of the last meal, length of fast or activity in the previous 24 h all of which in uence the RMR. Other factors which may in uence RMR and which differ between studies are the nature of the equipment used (face mask, nose clip, ventilated hood, ventilated tent, or whole body calorimeter), subject familiarity with the equipment and procedures, amount of activity on the morning of the measurement, or length of rest period before the measurement, all of which may increase the withinsubject variation over that obtained under the wellstandardized conditions of the classical BMR. Under well-standardized conditions CV wb for measured BMR averages about 2.5%. This value was derived from 279 subjects measured in a whole body calorimeter on successive nights on which the antecedent diet and activity had been the same. 27 However, in community studies subjects are sampled from the population and variation includes that due to antecedent diet and activity, natural weight uctuations, menstrual cycle and methodological error. Table 3 shows results from studies that have speci cally investigated within-subject variation in BMR; the majority were conducted in young men. The mean 1123 Table 3 Within subject variation in measured BMR Subjects n Age (y) Method a Repeats CV w Reference Males 8 21 ± 23 Cal BMR 4 occasions Males 5 18 ± 21 HB BMR 1=week5 weeks 3.6 b 56 Females ± 53 7 ± Mayo data Males 23 consecutive mornings 3.5 (quoted in Soares et al 56 ) Males 6 19 ± 50 DB BMR 21 week apart 4.0 b 57 Males 23 DB RMR 21 day apart 3.4 b week apart 3.2 b Males ± 23 HB BMR 3 measures in duplicate 3 ± 36 months apart and 5 weeks later Males 6 19 ± 24 HB BMR=RMR 2 per day14 days 4.6 b 60 Males 9 19 ± 36 VH RMR 3 times over 2 y 4.0 b 61 Males ± 62 5 ± 155 per person 4.0 Early physiologists quoted in Henry et al 61 Males ± 41 VH RMR 4 ± 6 replicates Non-obese=obese females 22=32 20 ± 45 VH RMR 4 ± 6 replicates Pre-pubertal girls 19 6 ± 10 VH 1 RMR, 2 BMR over 12 weeks a BH ˆ Hartmann Braun metabolator; VH ˆ ventilated hood; Cal ˆ whole body calorimeter; DB ˆ Douglas bag; BMR ˆ classic BMR subjects sleeps in the place where the measurement is to be made; RMR ˆ resting metabolic rate, subject brought to place of measurement on the morning of measurement. b Value known to include measurement error or data recalculated to include measurement error.

6 1124 Table 4 Within-subject variation in measured BMR from DLW studies Repeats Time DWt b Mean age Subjects n (n) span a (kg) (range)(y) CV w Reference Women, pre-pregnant and 6 weeks pregnant Mothers at 4, 8 and 12 weeks of lactation Women free-living. Winter and summer Men in metabolic facility for over- and under-feeding study. First and last measurements of the study at same weight Young men `from around the university' Men, light occupations, range of leisure activity. Summer and winter Adolescents in a metabolic facility. Baseline EI (days 0 ± 8) 1.61 BMR. Overfeeding EI (days 9 ± 22) 2.45BMR. Adult men, metabolic facility, imposed totally sedentary conditions, no ambulation Healthy older men and women not obese, weight stable. Supervised endurance training for 8 weeks Obese boys (30 ± 40% body fat). 2 weeks baseline. 4 weeks imposed exercise Sedentary adults, free-living. 40 weeks imposed training for half-marathon Adults, free-living. 4 ± 5 weeks sedentary baseline, 9 weeks imposed exercise. 8F weeks (24 ± 32) 10 F 3 8 weeks (26 ± 40) 6 F 2 6 ± 7 months (28 ± 39) 8 M 2 7 months (19 ± 49) 17 M 2 (n ˆ 7) 3 (n ˆ 10) 11 weeks (18 ± 30) 9 M 2 6 months 38 (25 ± 54) 6M 3F 2 22 days (13 ± 19) M 2 21 days 23.2 (20 ± 29) 6M weeks (56 ± 78) 5F 63.2 (56 ± 65) 10 M 2 6 weeks (10 ± 11) 8M 2 or 4 40 weeks F 3M 2F (32 ± 41) 2 11 weeks (27 ± 33) a Approximate time interval between rst and last measurement. b Maximum weight minimum weight, ie mean weight change whether gained or lost. Estimated basal metabolic rate (CV wb ). CV wb from these 11 studies was 3.9%. Table 4 shows the within-subject variation in BMR obtained from DLW studies with repeat measurements. The mean value from all 12 studies was 5.5%, but the mean from free-living individuals without imposed study conditions was 4.7% ( rst six studies in Table 4). The mean of these studies and those in Table 3 was 4.2%. A round gure of 4% is suggested as a suitable average value to substitute into the Goldberg equation, but other values may be used if deemed more appropriate for any given study. Where no measured BMR or RMR is available, it may be calculated from weight and height or weight alone. Many equations are available and each gives a slightly different result. 28 The equations based on the largest body of data and most widely used in dietary assessment are those of Scho eld, 29 modi ed to provide equations for age range 60 ± 74 and >74 y. 20 The value for CV wb used in the original paper 11 was a single value of 8%. However, speci c values for the different age ± sex groups of the Scho- eld equations are available 29 (Tables Al.l and A1.2, pp 31 ± 32 of Scho eld's paper). The standard errors for predictions of BMR expressed as a percentage of the mean BMR (ie coef cients of variation for predictions) from these tables are summarized in Table 5. Table 5 Standard errors for predictions of BMR expressed as a percentage of the mean BMR (ie coef cients of variation for predictions) Age group Male Female Under 3 y ± 10 y ± 18 y ± 30 y ± 60 y Over 60 y Mean 3 ± 18 y Mean 18 ± 60 y Mean 3 ± 60 y Adapted from Scho eld et al, A gure of 8.5% is suggested as a suitable average value to substitute into the Goldberg equation, but other values may be used for studies of speci c age ± sex groups. The validity of the Scho eld equations for obese subjects The body of data on which the Scho eld equations were based contained few subjects of very high body weight (>83 kg for men or >76 kg for women), and thus probably few of very high body mass index (BMI). Since adipose tissue has a lower metabolic

7 rate than lean tissue, the BMR of obese persons may be over-estimated by the equations and thus the extent of under-reporting as measured by EI:BMR estimated associated with obesity also over-estimated. This question has been examined using the database compiled for the review of DLW energy expenditure. 19 Figure 3 shows the plot of (BMR estimated BMR measured ) against ln(bmi) for males and females aged 18 y and over. The data have been tted with a quadratic equation. For males there was no signi cant difference between BMR est and BMR meas associated with BMI (P ˆ 0.4). For females the difference was highly signi cant (P ˆ ), but only for women with BMI>35 kg=m 2 was it of practical importance. Table 6 shows the mean difference by BMI category for both sexes. Since the proportion of the population with BMI>35 kg=m 2 is approximately 5 ± 7% only (Health Survey for England 1997, Department of Health press release), it is unlikely that using estimated BMR has caused the association between under-reporting and high BMI to have been exaggerated. The association appears extremely robust; it has been found in many studies including, for example, the USA CSFII 85 ± 86 survey 3 and the UK NDNS survey. 31 Limited data on the large under-reporting in the massively obese suggest that even adjusting the BMR would not transfer these persons into the category of acceptable reporters. In 324 women with measurements of EI and DLW EE, 18 had BM35 kg=m 2. In these women mean EI:EE was 0.74 (s.d. 0.33), compared with 0.78 (s.d. 0.25) for those with BMI 25 ± 34 kg=m 2 and 0.87 (s.d. 0.22) in those with BMI<25 kg=m 2. Using the revised Goldberg cut-offs to evaluate reported energy intake There are several scenarios for the evaluation of reported energy intake (EI rep ). 1. Evaluating the group mean EI rep. (A) In a small survey where n<100. (B) In a large survey where n> Evaluating EI rep at the individual level. (A) Where intake has been measured by short records or recalls of <14 days. (B) Where intake has been measured by methods presumed to measure `habitual' intake, ie records of 14 days or more, diet history (DH) or food frequency questionnaire (FFQ) Figure 3 Difference between estimated and measured BMR plotted against 1n(BMI) and tted with a quadratic equation.

8 1126 Table 6 Mean difference BMR estimated BMR measured (MJ) according to category of BMI as measured in DLW studies Males Females BMI n Mean s.d. P a n Mean s.d. P a < ± > a Paired t-test. Each of these situations will be considered in turn. Evaluating group mean EI rep :BMR in small studies (n<100) Table 7 illustrates how the CL (cut-offs) for group mean intake vary depending on the factors substituted in the equation. Standard factors for CV tp and CV wei have been used and the CV wb for estimated BMR. The effects of varying the PAL value chosen as the yardstick, the number of subjects in the study and the number of days of dietary assessment are shown. A number of key points are: The choice of the PAL value used as the yardstick for comparison with EI:BMR is important. If 1.55 was used for a population with a true mean energy expenditure of 1.80, then bias might be undetected or the degree of bias under-estimated. Bias is detected more readily in large studies, since the con dence limits (CL) become markedly narrower as n increases, the 95% CL range decreased from 0.38 with n ˆ 10 to 0.04 with n ˆ Increasing the days of dietary assessment (d) from 1 to 4 substantially improved chances of detecting bias. The 95% CL range decreased from 0.57 with d ˆ l to 0.41 with d ˆ 4. However, the CL for mean EI rep :BMR were not substantially improved by increasing d above 4 days. Evaluating group mean EI rep :BMR in large studies (n>100) When evaluating mean intake of larger groups there are further considerations which impact on the study design: Table 7 shows that changing n had a small effect on the con dence limits above n ˆ l00. For larger studies the CL were barely different from the PAL chosen for comparison. Studies with n>100 therefore constitute a special case in which the mean El rep :BMR may be compared directly with the PAL. Table 7 also shows that changing d, the number of days of dietary assessment, was of little importance compared with changing n, the number of subjects, in determining ability to detect bias in the mean intake. With n>100, increasing d had only a small effect on the CL and a negligible effect when n>500. Thus the number of days of assessment (d) is irrelevant to the evaluation of group mean intake in very large studies (n>500). Evaluating EI rep :BMR at the individual level in studies with short records or recalls (d< 14) The sensitivity and speci city of the Goldberg cut-off for identifying invalid reports at the individual level was investigated in depth in a previous paper. 7 That analysis was based on subjects from DLW studies. The sensitivity of the Goldberg cut-off was assessed Table 7 The effect of changing the values substituted into the Goldberg equation on the calculated con dence limits (cut-offs): evaluating group mean EI:BMR PAL CV wb CV tp CV wei d (days) n S Lower 95% CL Upper 95% CL 95% CL range Effect of changing PAL Effect of changing n Effect of changing d in a small study Effect of changing d in a large study

9 Table 8 The effect of changing the values substituted into the Goldberg equation on the calculated con dence limits (cut-offs): evaluating EI:BMR at the individual level 1127 PAL CV wb CV tp CV wei d (days) n S Lower 95% CL Upper 95% CL 95% CL range Effect of changing d Effect of changing PAL Effect of measuring BMR as the proportion of subjects classi ed as underreporters (UR) by EI:EE who were correctly classi ed as UR by EI:BMR. Some conclusions from that analysis are summarized below, but the original paper should be consulted for full understanding of the issues. The evaluation of individual intake is a special case of the Goldberg cut-off in which n ˆ 1. Table 8 shows how, in this special case, the CL alter according to changing the variables in the equation. Con dence limits were substantially wider when n ˆ 1 compared with larger n. The 95% CL ranges were 1.09 ± 1.88 in Table 8 compared with 0.04 ± 0.57 in Table 7. Thus the ability of the Goldberg cut-off to detect bias (invalid reports) at the individual level is limited. The ability to detect invalid reports improved as the number of days of dietary assessment (d) increased up to 7 days; the 95% CL range dropped from 1.88 for d ˆ 1 to 1.23 for d ˆ 7. The improvement beyond 7 days was smaller. (This should not be taken to indicate that 7 days is necessarily a desirable maximum, since the precision of the estimate of individual intake improves with p d and more days might be desirable in a given study) If 1.55 is a conservative estimate of normal activity, then using a higher PAL for comparison might improve the identi cation of under-reporters. However, using higher PAL values raised both the lower and the upper CL. In the previous paper 7 it was shown that if a single PAL and cut-off were used to evaluate the intake of all subjects in a study, about 20% were misclassi ed, and that the proportion of subjects misclassi ed remained largely unaltered whether the PAL chosen was 1.55 or higher. As the PAL value was raised, more underreporters were identi ed but more `acceptable' reporters were misclassi ed as under-reporters. The sensitivity was only improved if individuals were assigned to low, medium or high levels of activity 7 and three different cut-offs calculated for the PAL values assigned to each level. The ability to detect invalid reports may be marginally improved if a measured BMR is available. In small studies where the integrity of individual data is more important, measuring BMR may avoid misclassi cations 32 which could distort analysis. Evaluating EI rep :BMR at the individual level for long records, diet history or food frequency questionnaire Evaluation of EI rep :BMR when `habitual' intake is assumed to have been measured is another special case of the Goldberg equation. If the number of days (d ) is taken as in nity, the expression for CV wei disappears, but the errors due to CV wb and CV tp remain. The CL for `habitual' intake therefore are virtually the same as those for records of 21 days. In the original paper, 11 it was suggested that `habitual' intake be compared with an EI:BMR of 1.35, the lowest value compatible with a normally active lifestyle. This was referred to as CUT-OFF 1 and reported intakes for `habitual' intake below this value were to be rejected. However, this strategy ignored the errors associated with CV wb and CV tp and is inconsistent with the use of the Goldberg cut-off as described above. It has also contributed to confusion in the use of the Goldberg cut-off. CUT-OFF 1 should be abandoned. Limitations of the Goldberg cut-off There are considerable limitations to the Goldberg cut-off. It has poor sensitivity for de ning invalid reports at the individual level. The CL are wide and only extreme degrees of mis-reporting can be identi- ed. It can make no distinction between varying degrees of mis-reporting.

10 1128 The major limitation is that the use of EI:BMR for evaluating energy intake depends on knowledge of energy requirements or energy expenditure. For good reasons the original paper 11 assumed a sedentary lifestyle and used a PAL of 1.55 as the yardstick for evaluating mean energy intake. Since that publication numerous authors have used the cut-off for 1.55 and n ˆ l to identify individual low energy reporters (LERs). This strategy has enabled the characteristics of LERs to be explored and enhanced understanding of bias in reporting. However, analysis of accumulated data from DLW studies has shown, rst, that 1.55 is a conservative value for normally active populations and, second, that a blanket cut-off for 1.55 identi es only about 50% of those de ned as under-reporters by EI:EE ratio. 7 It ignores those who may have under-reported from high energy requirements, but whose reported intake has not fallen below this conservative cut-off. The conclusions from studies of LERs (reviewed by Macdiarmid and Blundell 33 ) need con rmation by further studies which identify reports of poor validity across the whole range of energy expenditures=requirements. This implies including measurements of activity or energy expenditure as routine in dietary surveys. At the group level knowledge of activity is needed to assign an appropriate mean PAL value for that population before the presence and=or degree of bias can be determined. At the individual level knowledge of activity or energy expenditure is needed to classify subjects into low, medium or high activity and to assign an appropriate PAL value to each level. It should be noted, however, that this strategy still applies blanket cut-offs across ranges of energy requirements, albeit three narrow ranges rather than one broad range. The direct measurement of energy expenditure is a more desirable aim. Unfortunately, techniques for reporting or measuring activity or energy expenditure are also subject to substantial errors. Detailed discussion of these are outside the scope of this paper. The following paragraphs merely indicate some starting points in the literature. For large-scale studies, a questionnaire that is simple, easy to administer and easy to analyse is required. Most physical activity questionnaires have been primarily designed to document high intensity exercise or activity suf cient to raise the heart rate. However, much of the variation between subjects comes from differences in time spent sitting, standing and moving around Ð activities that are dif cult to quantify. A questionnaire that elicits the pattern of the general lifestyle, the occupational activity, and the leisure activity is required. One widely used simple questionnaire providing three activity `scores' is the Baecke index 34 although its usefulness for the particular purpose of evaluating energy intake has yet to be determined. Activity diaries provide more detailed information, but are exceedingly demanding of subject co-operation. They are therefore unsuitable for large studies or those in which compliance may be low. One comparison of a simple questionnaire and a detailed activity diary has suggested that the former might be adequate for large-scale studies. 35 On the other hand, a validation of a physical activity recall against DLW energy expenditure and by heart rate monitoring found good agreement at the group but not at the individual level. 36 In small studies, simple questionnaires may lack suf cient precision for classi cation of subjects. It is more desirable to obtain a measure of energy expenditure. Methods for measuring energy expenditure include factorial estimates derived from physical activity diaries, or the proxy measurements of movement by accelerometer or heart rate monitoring. A triaxial accelerometer has been developed that is claimed to account for 65% of the variation in PAL 37,38 and may be suitable, in combination with measurement of BMR, to validate energy intake. Heart rate monitoring with individual calibration has also been used to measure energy expenditure in a community-based medium-sized epidemiological study. 39 The latter workers are also using heart rate monitoring to validate a physical activity questionnaire which might eventually prove useful as a routine tool. Conclusions and practical recommendations 1 Reported nutrient intake data should not be accepted without critical examination for misreporting and consideration of the possible in uences of mis-reporting on the conclusions drawn. 2 Dietary studies should include an internal validation procedure, whether by measurement of energy expenditure or urinary nitrogen excretion. Validation of energy intake is more appropriate when energy or macronutrients are the parameters of interest. Urinary nitrogen excretion is to be preferred when protein or micronutrients that are highly correlated with protein intake are the parameters of interest. 3 Since obesity and dieting have been identi ed as having the most robust association with mis-reporting, weight and height should be recorded and a short questionnaire to elicit information on weight consciousness and dieting behaviour also included. 4 At the group level the Goldberg cut-off can be used to determine the probable degree of overall bias to reported energy intake in a study. Comparison should be made with a PAL value appropriate to the study population based on information about physical activity or lifestyle. 5 At the individual level, the use of a single cut-off for EI:BMR applied to all subjects is inappropriate as it fails to identify under-reporters among those with high energy requirements. A short questionnaire on home, leisure and occupational activity such that subjects may be assigned to low, medium

11 or high PAL for calculating the Goldberg cut-off should be included as routine. A short questionnaire is probably adequate and logistically acceptable for large-scale studies. 6 In small studies (n<100) it is desirable to obtain a measure of energy expenditure. Detailed activity diaries for factorial calculation of energy expenditure, or possibly the use of triaxial accelerometers or heart rate monitoring are possibilities. If these measurements are obtained, energy intake may be compared directly with energy expenditure and the Goldberg cut-off is irrelevant. Acknowledgements I acknowledge with thanks and appreciation the long collaboration and contributions to my own work and thinking of many colleagues but particularly Drs Andrew Prentice, Andy Coward, Susan Jebb, Gail Goldberg, Peter Murgatroyd and Tim Cole at the Dunn Nutrition Centre and Dr Barbara Livingstone at the University of Ulster at Coleraine. References 1 Braam L, Ocke MC, Buenode Mesquita HB, Seidell JC. 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