Weight Cycling and Cancer Incidence in a Large Prospective US Cohort

Abstract

Weight cycling, which consists of repeated cycles of intentional weight loss and regain, is common among individuals who try to lose weight. Some evidence suggests that weight cycling may affect biological processes that could contribute to carcinogenesis, but whether it is associated with cancer risk is unclear. Using 62,792 men and 69,520 women enrolled in the Cancer Prevention Study II Nutrition Cohort in 1992, we examined the association between weight cycling and cancer incidence. Weight cycles were defined by using baseline questions that asked the number of times ≥10 pounds (4.54 kg) was purposely lost and later regained. Multivariable-adjusted hazard ratios and 95% confidence intervals for all cancer and 15 individual cancers were estimated by using Cox proportional hazards regression. During up to 17 years of follow-up, 15,333 men and 9,984 women developed cancer. Weight cycling was not associated with overall risk of cancer in men (hazard ratio = 0.96, 95% confidence interval: 0.83, 1.11 for ≥20 cycles vs. no weight cycles) or women (hazard ratio = 0.96, 95% confidence interval: 0.86, 1.08) in models that adjusted for body mass index and other covariates. Weight cycling was also not associated with any individual cancer investigated. These results suggest that weight cycling, independent of body weight, is unlikely to influence subsequent cancer risk.

Weight cycling, which is also referred to as “yo-yo dieting,” is the repeated intentional loss of weight followed by regain. With almost half of American adults attempting to lose weight (1) and most weight loss not maintained (2), weight cycling is very common. Weight cyclers tend to be heavier (3, 4) and to gain more weight over time (3, 5) than noncyclers. Although both weight and weight gain are known to increase risk of cancer (6–8), whether weight cycling is associated with cancer risk, independent of body weight, is largely unknown (9).

Although few studies have examined the physiological effects of weight cycling, there is some evidence that it could affect processes that contribute to carcinogenesis. Studies in mice have shown that weight cycling increases T-cell accumulation (10) and enhances inflammatory responses (11) in adipose tissue. In a study of postmenopausal women, frequent intentional weight loss was associated with lower natural killer cell cytotoxicity (12). Such changes could result in increased chronic inflammation and decreased immune surveillance for transformed cells. Animal and human studies suggest that weight cycling results in increased fat accumulation (13–15), particularly abdominal or visceral fat (16, 17); alters levels of glucose, insulin, cholesterol, and some hormones (18–20); causes impaired systemic glucose tolerance (10); and is associated with shorter telomeres in peripheral blood lymphocytes (21). However, many of these findings have not been replicated, and at least 2 other studies showed no associations between weight cycling and accumulation of fat mass (22, 23).

The epidemiologic studies of weight cycling in relation to cancer risk published to date investigated relatively few cancers, and most included only women. French et al. (24) investigated the relationship between weight cycling and several diseases in the Iowa Women's Health Study and found no associations with endometrial, breast, colon, or lung cancer. Three case-control (25–27) and 1 prospective cohort study (28) also investigated the association of weight cycling with endometrial cancer. Two of these studies had null results (25, 28), 1 found a 26% increased risk (26), and the other found no association in women who had never been obese and an almost 3-fold increased risk for women who had ever been obese (27). Two case-control studies of breast cancer found disparate results, with 1 being null (29) and the other reporting that a fluctuating weight pattern was associated with a nearly 2-fold increased risk (30). A case-control study of renal cell cancer found no association for men but a 2.3-fold increased risk for women (31), while a prospective study with only women found a 2.6-fold increased risk (32). Finally, weight cycling was not associated with risk of non-Hodgkin lymphoma, leukemia, or multiple myeloma in a case-control study of women (33). Thus, the existing evidence does not clearly indicate whether weight cycling is associated with risk of any specific cancer (9) and does not address overall cancer risk.

Because many people try to lose weight (1), understanding whether weight cycling is associated with cancer risk has important public health implications. If weight cycling is associated with increased cancer risk, then more effort should be made to encourage the avoidance of weight gain by normal weight individuals and the maintenance of weight loss by overweight and obese individuals.

In this study, we investigated the association between weight cycling and cancer incidence using the American Cancer Society Cancer Prevention Study II (CPS-II) Nutrition Cohort. Weight cycling was defined by using self-reported intentional weight loss and regain episodes collected at baseline in 1992, and cancer incidence from 1992 to 2009 was used to determine risk of all cancer and 15 specific cancers in both men and women.

METHODS

Study population

The participants of this study (86,402 men and 97,785 women) were from the CPS-II Nutrition Cohort, a prospective study of cancer incidence begun in 1992. The Nutrition Cohort, which is described in detail elsewhere (34), was formed as a subcohort of CPS-II, a prospective study of cancer mortality involving approximately 1.2 million Americans enrolled in 1982. Nutrition Cohort participants were recruited from CPS-II participants between the ages of 50 and 74 years who were living in 21 states. Participants completed a self-administered questionnaire at baseline in 1992–1993 that included questions on demographic, anthropometric, medical, and lifestyle information. Follow-up questionnaires to update exposure information and to ascertain newly diagnosed cancers were sent to living Nutrition Cohort members in 1997 and every 2 years since. The response rate on all of the follow-up questionnaires, among those cohort participants who were mailed surveys, was at least 85%. Follow-up time for this study was from the date of receipt of the baseline survey in 1992–1993 to the date of diagnosis of cancer, death, date of the last survey returned, or June 30, 2009, whichever came first. All aspects of the CPS-II Nutrition Cohort Study were approved by the Emory University Institutional Review Board (Atlanta, Georgia).

CPS-II participants were excluded from this analysis if they were lost to follow-up (3,141 men and 3,111 women), reported any prevalent cancer other than nonmelanoma skin cancer in 1982 or 1992–1993 (9,769 men and 13,094 women), self-reported a cancer that could not be verified (527 men and 354 women), or self-reported a cancer more than 6 months before the diagnosis date of that cancer (100 men and 48 women). Also excluded were participants with missing body mass index (BMI, expressed as weight (kg)/height (m)²) or BMI <18.5 in 1982 or 1992 (2,161 men and 4,174 women), missing information on number of times a participant purposely lost ≥10 pounds (4.54 kg), or regained weight that was purposefully lost (1,098 men and 1,866 women), as well as those who reported never having purposefully lost ≥10 pounds but reported ≥10 pounds as the most weight ever purposefully lost (1,070 men and 936 women), and those who reported the number of times that weight was purposefully lost that differed from the number of times weight was regained by more than 2 (2,468 men and 3,004 women). The characteristics of the individuals excluded for missing or invalid weight loss and regain information are shown in Web Table 1 available at http://aje.oxfordjournals.org/. Finally, because an undiagnosed cancer at baseline may have influenced weight patterns, the first 2 years of follow-up (1992–1994, 3,276 men and 1,678 women) were excluded from all analyses. For analyses of endometrial cancer, women who reported removal of their uterus or had unknown uterine status at baseline were also excluded (27,022 women). Likewise, for ovarian cancer analyses, women who reported removal of their ovaries or had unknown ovarian status at baseline were also excluded (15,811 women). Women who reported removal of their uterus or ovaries during follow-up were censored from analyses of these cancers at the time of receipt of the questionnaire on which the procedure was reported. Thus, the final analytical cohort for this study comprised 62,792 men and 69,520 women except for analyses of endometrial or ovarian cancer, where the analytical cohort was 42,498 or 53,709 women, respectively.

Identification of cancer cases

A total of 25,317 first cancers occurred in the analytical cohort between 1994 and 2009. Of these, 24,217 were self-reported on follow-up questionnaires and subsequently verified through medical records (n = 15,911) or linkage with state cancer registries (n = 8,306). An additional 1,100 cancers were identified through linkage to the National Death Index. For analyses of all cancers and specific cancers, all cases were the first diagnosed cancer for that individual. In situ diseases were not included for breast, prostate, colon, or rectal cancers. Aggressive prostate cancer was defined as a Gleason score ≥8, American Joint Committee on Cancer tumor stage T3 or T4, or lethal prostate cancer of unknown stage.

Characterization of weight cycling

CPS-II Nutrition Cohort participants were asked 2 questions about weight cycling on the baseline questionnaire: “How many times in your life have you purposely lost 10 pounds or more?” and “How many times in your life have you regained as much as 10 pounds that you previously had lost?” A write-in answer was required for both questions with space for a range from 0 to 99. A weight cycle was defined as a combination of both an intentional loss and a subsequent regain. Thus, someone who reported 3 weight losses and 3 subsequent regains was assigned 3 weight cycles, whereas someone who reported 3 weight losses but only 2 regains was assigned 2 weight cycles. The individuals who reported weight cycling (“weight cyclers”) were further categorized by the total number of cycles they reported: 1–4, 5–9, 10–19, and 20 or more weight cycles. Weight change pattern or trajectory was not considered in assigning weight cycling status or category.

Statistical analyses

BMI in 1982 and 1992 was calculated by using self-reported body weight on questionnaires completed at those times and height self-reported on the 1982 questionnaire. BMI at age 18 years was calculated by using weight at age 18 years recalled and reported on the 1992 questionnaire.

All statistical analyses were conducted with SAS, version 9.3, software (SAS Institute, Inc., Cary, North Carolina). Cox proportional hazards regression (35) was used to determine the hazard ratios and 95% confidence intervals for the association of weight cycling with cancer incidence. The time scale for the analysis was time in study from cohort entry in 1992–1993. P values for linear trend were estimated by modeling weight cycles as a continuous variable and were derived by use of the Wald test (36). The Cox proportional hazards assumption was tested by modeling multiplicative interaction terms between weight cycling numbers and time. The statistical significance of the interaction term was assessed by using the likelihood ratio test (36), and no deviations from proportional hazards were found. Interactions between either BMI or age at baseline and weight cycling were tested by using the likelihood ratio test in which either variable was modeled categorically. All P values are 2 sided.

Cox models were adjusted for age by stratifying on exact year of age in 1992. Additional covariates included in the multivariable-adjusted models were race (white, black, or other/missing), education (some high school, high school graduate, some college, college graduate, or missing), smoking status (nonsmoker, former <20 years, former ≥20 years, current <40 years, current ≥40 years, ever/unclassifiable, or missing), physical activity in metabolic equivalents (<8, 8 to <17.5, 17.5 to <31.5, ≥31.5/week, or missing), use of aspirin or other nonsteroidal antiinflammatory drugs (not current user, 1–14 pills/month, 15–29 pills/month, 30–59 pills/month, ≥60 pills/month, or missing/unknown), and history of diabetes (yes, no). For analyses of women, postmenopausal hormone use (never, current estrogen, current combined hormone replacement therapy, former estrogen or combined hormone replacement therapy, or ever/other/unknown) was also included in the model. Included in analyses of prostate, breast, and colon and rectal cancer were history of prostate-specific antigen testing (starting in 1997), mammography (starting in 1992), and colorectal endoscopy (starting in 1997), respectively, and all were modeled as time-dependent variables. BMI in 1992 (continuous, modeled linearly) was also included in the models as indicated. The inclusion of alcohol intake as a covariate did not change the associations and was not included in the final multivariable-adjusted model.

RESULTS

A large proportion of CPS-II Nutrition Cohort participants reported some weight cycling. More women (57%) than men (43%) weight cycled, and women were more likely to report higher numbers of weight cycles than men.

Weight cyclers tended to be younger, were more likely to be former smokers and to have a history of diabetes, and were less likely to be current smokers than noncyclers (Table 1 (men) and Table 2 (women)). Weight cyclers also had a higher BMI at age 18, in 1982, and in 1992 and gained more weight in the intervals between age 18 and 1982 and between 1982 and 1992 than did noncyclers. Women weight cyclers were also more likely to be former postmenopausal hormone users and less likely to be current postmenopausal hormone users and to have a higher energy intake and a lower alcohol consumption.

Table 1.

Age-Standardized^a Frequencies^b of Selected Characteristics of Men by Weight Cycling Status, Cancer Prevention Study II Nutrition Cohort, 1992

Variable	No. of Weight Cycles
	Noncycler (n = 36,103)		1–4 (n = 19,466)		5–9 (n = 3,936)		10–19 (n = 2,430)		≥20 (n = 857)
	Mean	%	Mean	%	Mean	%	Mean	%	Mean	%
Age in 1992, years	64.5		63.1		62.0		61.6		60.9
Race
White		97.3		97.9		98.1		98.3		98.2
Black		1.2		1.0		1.1		0.8		1.4
Other/missing		1.5		1.1		0.8		0.9		0.4
Education
<High school		8.4		6.7		5.8		5.4		6.3
High school graduate		20.2		17.5		15.0		15.5		12.3
Some college		25.2		25.9		24.6		29.1		31.8
≥College graduate		45.6		49.2		53.8		49.3		49.3
Missing		0.6		0.7		0.8		0.7		0.3
Smoking status
Never		35.0		29.7		30.0		28.3		31.2
Current		10.6		6.5		6.0		5.9		5.7
Former		53.6		63.0		63.2		65.1		62.4
Missing		0.8		0.8		0.8		0.7		0.7
Diabetes, yes		6.7		11.7		13.9		14.3		15.8
Alcohol, g/day	11.5		11.0		10.8		10.8		11.7
Energy, kcal/day	1,818		1,776		1,784		1,807		1,872
BMIc at age 18	21.3		22.4		23.3		23.6		24.6
BMIc in 1982	25.0		27.2		28.2		28.8		29.9
BMIc in 1992	25.2		27.7		29.0		29.9		31.0
Exercise, METs/hour	13.2		13.1		13.6		13.9		13.4
Wt Δ between age 18 years and 1982, poundsd	26.0		33.6		34.9		36.3		37.3
Wt Δ between 1982 and 1992, pounds	1.7		4.0		5.5		7.7		8.2

Abbreviations: BMI, body mass index; METs, physical activity in metabolic equivalent; Wt Δ, weight change.

^a Standardized to the age distribution of the men in the cohort.

^b Values are presented as percentages unless otherwise noted.

^c BMI expressed as weight (kg)/height (m)².

^d One pound = 0.454 kg.

Table 2.

Age-Standardized^a Frequencies^b of Selected Characteristics of Women by Weight Cycling Status, Cancer Prevention Study II Nutrition Cohort, 1992

Variable	No. of Weight Cycles
	Noncycler (n = 29,879)		1–4 (n = 25,559)		5–9 (n = 6,663)		10–19 (n = 4,990)		≥20 (n = 2,429)
	Mean	%	Mean	%	Mean	%	Mean	%	Mean	%
Age in 1992, years	63.0		61.7		60.7		60.2		59.7
Race
White		97.1		97.6		97.9		98.0		98.1
Black		1.3		1.4		1.4		1.5		1.1
Other/missing		1.6		1.0		0.7		0.5		0.8
Education
<High school		4.8		4.7		4.6		5.0		5.6
High school graduate		31.7		32.0		32.0		30.7		27.6
Some college		29.6		31.7		32.3		34.1		38.1
≥College graduate		33.3		30.9		30.5		29.6		27.8
Missing		0.6		0.7		0.6		0.6		0.9
Smoking status
Never		56.8		54.6		52.5		51.1		47.3
Current		9.4		7.3		6.6		6.7		7.2
Former		32.3		36.6		39.5		40.9		44.2
Missing		1.5		1.5		1.4		1.3		1.3
Diabetes, yes		4.1		6.6		8.8		10.8		12.9
PMH use in 1992
Never		42.5		42.5		42.5		42.7		42.4
Current ERT		18.8		19.4		19.3		20.6		17.8
Current CHRT		13.7		11.3		10.2		10.2		10.4
Former ERT/CHRT		17.8		19.7		21.1		19.9		22.1
Ever/other/missing		7.2		7.1		6.9		6.6		7.3
Alcohol, g/day	5.4		4.4		4.0		3.7		3.5
Energy, kcal/day	1,357		1,359		1,385		1,393		1,451
BMIc at age 18	19.9		20.9		21.7		22.3		23.0
BMIc in 1982	22.5		25.1		26.9		27.9		29.1
BMIc in 1992	23.3		26.5		28.7		29.9		31.3
Exercise, METs/hour	12.3		11.8		11.6		11.7		12.1
Wt Δ between age 18 and 1982, poundsd	15.2		24.5		30.9		33.3		36.3
Wt Δ between 1982 and 1992, pounds	5.1		8.5		10.5		11.7		13.0

Abbreviations: BMI, body mass index; CHRT, combined hormone replacement therapy; ERT, estrogen replacement therapy; METs, physical activity in metabolic equivalents; PMH, postmenopausal hormone; Wt Δ, weight change.

^a Standardized to the age distribution of the women in the cohort.

^b Values are presented as percentages unless otherwise noted.

^c BMI expressed as weight (kg)/height (m)².

^d One pound = 0.454 kg.

The associations of weight cycling with risk of all cancer and 11 individual cancers for men and women combined are shown in Table 3. The associations for men and women separately for these cancers plus 4 sex-specific cancers are shown in Web Table 2. In men and women combined and in women only, weight cycling was associated with overall cancer risk before but not after adjustment for BMI at baseline in 1992. In men alone, weight cycling was not associated with overall cancer risk in either model.

Table 3.

Association of Weight Cycling With Cancer Incidence in Men and Women Combined, Cancer Prevention Study II Nutrition Cohort, 1992–2009

Cancer Type and No. of Weight Cycles	Total No.	No. of Cases	HRa	95% CI	HRb	95% CI
All cancer
0	65,982	13,101	1.00	Referent	1.00	Referent
1–4	45,025	8,423	1.02	0.99, 1.05	0.99	0.96, 1.02
5–9	10,599	1,930	1.05	1.00, 1.10	1.00	0.95, 1.05
10–19	7,420	1,309	1.07	1.01, 1.13	1.01	0.95, 1.07
≥20	3,286	554	1.07	0.98, 1.16	0.99	0.91, 1.08
P for trend			0.02		0.93
Colon cancerc
0	65,982	872	1.00	Referent	1.00	Referent
1–4	45,025	608	1.12	1.01, 1.24	1.05	0.94, 1.17
5–9	10,599	123	1.04	0.86, 1.26	0.93	0.76, 1.14
10–19	7,420	85	1.08	0.86, 1.35	0.94	0.74, 1.19
≥20	3,286	32	0.95	0.66, 1.35	0.80	0.55, 1.16
P for trend			0.53		0.53
Rectal cancerc
0	65,982	259	1.00	Referent	1.00	Referent
1–4	45,025	146	0.89	0.73, 1.10	0.84	0.67, 1.04
5–9	10,599	44	1.21	0.88, 1.68	1.09	0.77, 1.54
≥10	10,706	29	0.82	0.56, 1.22	0.72	0.47, 1.09
P for trend			0.27		0.11
Pancreatic cancer
0	65,982	263	1.00	Referent	1.00	Referent
1–4	45,025	202	1.23	1.02, 1.49	1.09	0.89, 1.33
5–9	10,599	43	1.20	0.87, 1.67	0.98	0.69, 1.38
≥10	10,706	40	1.18	0.84, 1.67	0.90	062, 1.30
P for trend			0.31		0.65
Renal cancer
0	65,982	207	1.00	Referent	1.00	Referent
1–4	45,025	167	1.25	1.02, 1.54	1.12	0.90, 1.39
5–9	10,599	45	1.51	1.09, 2.10	1.25	0.89, 1.77
≥10	10,706	44	1.55	1.10, 2.16	1.20	0.83, 1.74
P for trend			0.05		0.68
Esophageal cancer
0	65,982	102	1.00	Referent	1.00	Referent
1–4	45,025	68	1.18	0.86, 1.61	1.15	0.83, 1.61
≥5	21,305	28	1.25	0.81, 1.92	1.21	0.75, 1.94
P for trend			0.57		0.76
Liver cancer
0	65,982	77	1.00	Referent	1.00	Referent
1–4	45,025	58	1.24	0.88, 1.76	0.99	0.68, 1.42
≥5	21,305	21	1.08	0.66, 1.78	0.69	0.40, 1.20
P for trend			0.82		0.30
Non-Hodgkin lymphoma
0	65,982	744	1.00	Referent	1.00	Referent
1–4	45,025	519	1.08	0.96, 1.21	1.01	0.89, 1.14
5–9	10,599	109	1.00	0.82, 1.23	0.90	0.73, 1.11
10–19	7,420	76	1.04	0.82, 1.32	0.91	0.70, 1.17
≥20	3,286	37	1.19	0.85, 1.66	1.00	0.70, 1.42
P for trend			0.23		0.99
Multiple myeloma
0	65,982	146	1.00	Referent	1.00	Referent
1–4	45,025	102	1.09	0.84, 1.41	1.05	0.80, 1.38
5–9	10,599	26	1.24	0.81, 1.90	1.18	0.76, 1.85
≥10	10,706	20	1.00	0.62, 1.62	0.94	0.56, 1.57
P for trend			0.19		0.31
Lung cancer
0	65,982	1,301	1.00	Referent	1.00	Referent
1–4	45,025	696	0.90	0.82, 0.99	0.94	0.85, 1.04
5–9	10,599	142	0.85	0.72, 1.02	0.91	0.76, 1.10
10–19	7,420	92	0.82	0.66, 1.01	0.89	0.71, 1.11
≥20	3,286	49	1.00	0.75, 1.33	1.11	0.82, 1.50
P for trend			0.30		0.97
Melanoma
0	65,982	813	1.00	Referent	1.00	Referent
1–4	45,025	552	1.03	0.93, 1.15	1.07	0.95, 1.20
5–9	10,599	140	1.14	0.95, 1.36	1.20	0.99, 1.46
10–19	7,420	94	1.15	0.92, 1.42	1.23	0.98, 1.55
≥20	3,286	25	0.72	0.48, 1.07	0.79	0.52, 1.19
P for trend			0.65		0.97
Stomach cancer
0	65,982	134	1.00	Referent	1.00	Referent
1–4	45,025	72	0.97	0.72, 1.30	0.98	0.72, 1.34
≥5	21,305	27	0.92	0.60, 1.42	0.95	0.59, 1.51
P for trend			0.89		0.99

Abbreviations: CI, confidence interval; HR, hazard ratio.

^a Base model: adjusted for age, race, education, smoking status, physical activity, nonsteroidal antiinflammatory drug use, history of diabetes, and postmenopausal hormone use.

^b Base model plus body mass index expressed as weight (kg)/height (m)² in 1992 (continuous).

^c Additionally adjusted for colorectal endoscopy starting in 1997 (time dependent).

The association between weight cycling and prostate cancer risk, which accounted for almost 47% of all cancers in men, was null, regardless of whether BMI was included in the model. No association was observed when only aggressive prostate cancers were examined. The associations with postmenopausal breast and endometrial cancers were similar to those for overall cancer risk in women, where a positive association was observed before but not after adjustment for BMI. There was no association between weight cycling and ovarian cancer risk either prior to or after adjustment for BMI.

Weight cycling was not statistically significantly associated with a number of other cancers, including colon, rectal, pancreatic, renal, esophageal, liver, lung, and stomach cancers, non-Hodgkin lymphoma, multiple myeloma, and melanoma, in both men and women (Table 2 and Web Table 2).

The associations with cancer risk determined by using weight cycling as a continuous variable before and after adjustment for BMI in men and women separately are shown in Figures 1 and 2. The only statistically significant association after adjustment for BMI was for esophageal cancer in women. However, this analysis included only 35 cases (20 weight cycling cases).

Figure 1.

Association of weight cycling with cancer incidence in men in the Cancer Prevention Study II Nutrition Cohort. Hazard ratios (HRs) and 95% confidence intervals (CIs) are shown for each cancer for analyses adjusted for age, race, education, smoking status, physical activity, nonsteroidal antiinflammatory drug use, and history of diabetes (A) and all these variables plus body mass index expressed as weight (kg)/height (m)² in 1992 (continuous) (B). The HR is the association per 5 weight cycles determined by using weight cycles as a continuous variable. All analyses for prostate cancer are additionally adjusted for prostate-specific antigen screening starting in 1997 (time dependent), and all analyses for colon and rectal cancer are additionally adjusted for colorectal endoscopy starting in 1997 (time dependent).

Figure 2.

Association of weight cycling with cancer incidence in women in the Cancer Prevention Study II Nutrition Cohort. Hazard ratios (HRs) and 95% confidence intervals (CIs) are shown for each cancer for analyses adjusted for age, race, education, smoking status, physical activity, nonsteroidal antiinflammatory drug use, history of diabetes, and postmenopausal hormone use (A) and all these variables plus body mass index expressed as weight (kg)/height (m)² in 1992 (continuous) (B). The HR is the association per 5 weight cycles determined by using weight cycles as a continuous variable. All analyses for breast cancer are additionally adjusted for mammography starting in 1992 (time dependent), and all analyses for colon and rectal cancer are additionally adjusted for colorectal endoscopy starting in 1997 (time dependent).

Because a recent analysis of weight cycling and endometrial cancer found a statistically significant interaction with history of obesity (BMI <30) (27), we examined associations stratified by obesity status. “Never obese” individuals were those whose BMI was <30 at age 18, in 1982, and in 1992, whereas “ever obese” individuals reported having a BMI ≥30 at 1 or more of these time points. The results of these analyses for overall cancer risk, the most common cancers in men and women (prostate and postmenopausal breast), kidney cancer in men, and endometrial cancer in women are shown in Figures 3 (men) and 4 (women). These associations, as well as those for all other cancers (Web Table 3), were null regardless of history of obesity. Although the point estimates were above 1 for some cancers among the ever obese and primarily less than or equal to 1 for the never obese, the interaction was not statistically significant for any cancer in either sex.

Figure 3.

Association of weight cycling with overall, prostate, and renal cancer risk in men stratified by obesity status in the Cancer Prevention Study II Nutrition Cohort, 1992. Obesity status: A) never obese, body mass index <30; B) ever obese, body mass index ≥30. Hazard ratios (HRs) and 95% confidence intervals (CIs) were determined in analyses for age, race, education, smoking status, physical activity, nonsteroidal antiinflammatory drug use, history of diabetes, and body mass index expressed as weight (kg)/height (m)² in 1992 (continuous).

Figure 4.

Association of weight cycling with overall, breast, and endometrial cancer risk in women stratified by obesity status in the Cancer Prevention Study II Nutrition Cohort, 1992. Obesity status: A) never obese, body mass index <30; B) ever obese, body mass index ≥30. Hazard ratios (HRs) and 95% confidence intervals (CIs) were determined in analyses for age, race, education, smoking status, physical activity, nonsteroidal antiinflammatory drug use, history of diabetes, postmenopausal hormone use, and body mass index expressed as weight (kg)/height (m)² in 1992 (continuous).

Whether the association of weight cycling with cancer risk varied by age was investigated in stratified analyses with either younger or older participants defined relative to the median age of the study population (64 years for men and 62 years for women). In older women aged ≥62 years, there was a statistically significant positive trend between number of cycles and ovarian cancer (for 1–4 cycles: hazard ratio (HR) = 1.02, 95% confidence interval (CI): 0.72, 1.44; for 5–9 cycles: HR = 1.89, 95% CI: 1.15, 3.13; for ≥10 cycles: HR = 2.07, 95% CI: 1.23, 3.50) (P for trend = 0.03), whereas the association was null in younger women aged <62 years (for 1–4 cycles: HR = 0.88, 95% CI: 0.60, 1.29; for 5–9 cycles: HR = 1.13, 95% CI: 0.66, 1.95; for ≥10 cycles: HR = 0.68, 95% CI: 0.36, 1.28) (P for trend = 0.41). However, the age-specific associations of weight cycling with ovarian cancer risk were not statistically significantly different (P for interaction = 0.12). All other associations with weight cycling in younger and older men and women were null (data not shown).

Although smoking status was controlled for in analyses of weight cycling and cancer risk, it is possible that residual confounding by smoking dose could influence our findings. Therefore, in sensitivity analyses, the study population was limited to never and long-term former (≥20 years) smokers. All associations were null and similar to those obtained with the full study population (data not shown).

DISCUSSION

This study is the first to comprehensively investigate the associations of weight cycling with overall and site-specific cancer risk in men and women. Results of this large prospective study showed that weight cycling was not associated with overall cancer risk or risk of 15 specific cancers independent of BMI in men or in women.

Our null findings agree with the results of some (24, 25, 28, 29, 33), but not all (26, 27, 30–32), previous studies of weight cycling and risk of specific cancers. Two factors that likely contribute to inconsistencies across studies include variability in how body size was accounted for in statistical models and differences in how weight cycling was defined. The importance of adjusting for BMI can be seen by comparing our results for all cancers, endometrial cancer, and renal cancer in women before and after this adjustment (Figure 2 and Web Table 2). In these analyses, weight cycling was statistically significantly associated with increased risk before adjustment for BMI but not afterwards. Potential confounding by BMI was not considered in the case-control study that found that weight cycling was associated with a statistically significantly 2.1-fold higher risk of breast cancer (30). Body size was adjusted for somewhat differently by 2 of the other case-control studies that reported significant results (26, 32). Although BMI as a continuous variable was used in our study, Trentham-Dietz et al. (26) used quartiles of BMI and Luo et al. (32) used the waist-to-hip ratio in their analyses of the associations of weight cycling and risk of endometrial or renal cancer, respectively. It is possible that residual confounding might account for the positive findings in these studies.

The lack of a standardized definition of weight cycling has led to considerable variability in the classification of weight cyclers among studies. The intentionality of the weight loss is particularly important to consider because unintentional weight loss occurs with some unhealthy behaviors such as smoking and is also an early symptom of many diseases, including several cancers. Therefore, it may be strongly related to the disease or mortality endpoint. Thus, only intentional weight loss should be considered when defining weight cyclers. However, many studies defined weight cyclers on the basis of weight patterns or variability over time, regardless of intentionality. Such studies have reported significant associations for breast (30) and renal (32) cancer. Other features of the weight cycling criteria that often differ between studies are the magnitude of the weight cycle and the period over which the weight cycling occurred. In the case-control study that found a significant association between weight cycling and risk of renal cancer in women but not men (31), a weight cycle was defined as 3 or more instances of intentional weight loss of >5 kg for women and >10 kg for men. Lindblad et al. (31) suggest that this difference may account for their disparate findings by sex. However, we defined a weight cycle as ≥10 pounds, which was similar to the >5 kg criteria for women, and our results in both men and women were null (31). Whether consideration of weight cycles of a larger magnitude would alter our findings is not known.

The possibility that the association between weight cycling and endometrial cancer risk might differ by obesity status was suggested by an Australian case-control study that found no association among never-obese women and a statistically significant 2.8-fold higher risk among ever-obese women (27). In our study, there were modest, but not statistically significant, differences between never-obese and ever-obese women and men for some cancers, including endometrial cancer. Whereas 90% of the weight cycling cases in the Australian case-control study fell into the ever-obese category, only 45% of the endometrial cancer cases in our study were ever obese. Thus, it is possible that our ability to detect an association may have been limited by the number of ever-obese individuals. Further research in populations with more obese subjects is needed to better understand the association between weight cycling and cancer risk in this group.

To our knowledge, this study is the largest and most comprehensive study of weight cycling and cancer risk conducted to date. The size of the study and the large number of weight cyclers enabled the examination of dose-response relationships. Other strengths include the prospective nature of the study and the availability of information about important covariates, which were needed to separate the influence of these factors on cancer risk from those of weight cycling.

An important limitation to this study was the lack of information on when during adulthood the weight cycling occurred. Also absent was information on the magnitude of the weight cycling or information on weight cycles of more than 10 pounds. Thus, we were unable to assess whether specific exposure periods or amplitude of gain/loss was associated with cancer risk. An additional limitation was the number of ever-obese individuals, which limited the power of our analyses in this group. Small sample sizes in analyses stratified by either obesity or age for some of the rarer cancers also limit our ability to determine whether our null results reflect a true lack of association or difference between associations. The limited number of esophageal cancer cases also precluded investigation of subtypes of this cancer. Finally, because weight cycling was self-reported, there may be some misclassification due to incorrect responses.

In summary, the results of this study indicate that weight cycling, independent of BMI, is not associated with overall cancer risk or risk of any specific cancer in either men or women. Therefore, overweight and obese individuals should be encouraged to attempt to lose weight even though the weight lost may be regained.