450 LETTERS TO THE EDITOR TRANSFUSION Vol. 34. No. 5-1994

blood collected at Miyazaki Red Cross Blood Center between June and August 1993 were screened for HTLV-I proviral DNA by polymerase chain reaction (PCR). Nested double PCR was performed on DNA extracted from buffy coats of donor blood by the method of Matsumoto et aL4 with slight modification. HTLV-I pdtax primers were used to detect at least one HTLV- I-infected cell in 200,000 noninfected cells. Although the sen- sitivity was reduced to about one-tenth since DNA from 10 samples was mixed and screened together, the method used was able to detect one HTLV-I-positive sample even when it was mixed with 20 HTLV-I-negative samples (data not shown). A total of 1101 seronegative blood components were tested for HTLV-I proviral DNA; all results were negative. This finding is in agreement with the report of Matsumoto et aL4 who tested 155 seronegative samples with PCR in Tokyo, an area that is not endemic for HTLV-I. Sato and Okochi5 reported that none of 712 patients who received blood components in Fukuoka, an HTLV-I-endemic area, had seroconverted for HTLV-I since 1990 when a modified particle agglutination test was introduced to screen blood components. Our results, as well as those of de Montalembert et al.,’ confirm that the modified particle ag- glutination test, currently used in Japan, is very effective in screening blood components to detect potential HTLV-I infec- tivity.

SEIJI KONW, MD Blood Transfusion Division

Miyazaki Medical Center 5200 Kihara Kiyotake

Miyazaki 889-16 Japan

NORIKO MISHIRO KAZUMI UMEKI, MT TOMIO KOTANI, MD

Central Laboratory for Clinical Investigation Miyazaki Medical College

Miyazaki MIEKO SETOGUCHI

TOSHIMITSU SHINGU, MD Miyazaki Red Cross Blood Center

Miyazaki SACHIYA OHTAKI, MD

Blood Transfusion Division Miyazaki Medical Center

and Central Laboratory for Clinical Investigation

Miyazaki Medical College

1.

2.

3.

4.

5 .

References De Montalembert M, Wattel E, Lefrkre F, et al. Human T-cell lymphotropic virus type I and I1 DNA amplification in seropositive and seronegative at-risk individuals. Transfusion

Saito S, Ando Y, Furuki K. et al. Detection of HTLV-I genome in seronegative infants born to HTLV-I seropositive mothers by polymerase chain reaction. Jpn J Cancer Res 1989;80:808- 12. Ehrlich GD, Glaser JB, Abbott MA, et al. Detection of anti- HTLV-I tax antibodies in HTLV-I enzyme-linked immuno- sorbent assay-negative individuals. Blood 1989;74: 1066-72. Matsumoto C, Mitsunaga S, Oguchi T, et al. Detection of hu- man T-cell leukemia type I (HTLV-I) provirus in an infected cell line and in peripheral mononuclear cells of blood donors by the nested double polymerase chain reaction method: com- parison with HTLV-I antibody tests. J Virol 1990;64:5290-4. Sato H, Okwhi K. Transfusion transmitted infection of HTLV-I and its prevention. Gann Monogr Cancer Res 1992;39: 15 1-62.

1993;33: 106- 10.

Perioperative blood transfusion and colorectal cancer recurrence

To the Editor: There are numerous factual errors in the recent review by

Vamvakas and Moore in TRANSFUSION.’ A few examples of the misrepresentations follow. First, Vamvakas and Moore claim that the confounders of age and duration of operation were not included in the final statistical model in our first ar- ticle’ on this subject. This is not true. In the culminating para- graph of the Results section, we demonstrated that after trans- fusion was added to a model that already included age and duration of operation, transfusion was the most significant predictor of recurrence, and the confounders became nonsig- nificant. This shows that the association of transfusion with earlier recurrence was extremely unlikely to be explained by these variables. Second, Vamvakas and Moore claim that the confounder of preoperative hemoglobin was not included in the statistical model demonstrating a deleterious association between transfusion and death due to colon cancer in the ar- ticle by Foster et a1.3 This is untrue, as is evident from Table 4 of that article, in which hemoglobin level was found to be nonsignificant (p = 0.53) in a model including transfusion (p = 0.05). Foster et aL3 discussed this issue extensively (p 1198). Third, Vamvakas and Moore claim that the study by Marsh et a1.4 investigating our report of an association between whole blood transfusion and cancer recurrence “pointed to plasma protein fraction as the culprit.” In fact, that study was designed to detect an effect of any type of plasma transfusion, and it found that patients receiving whole blood had a 2.47-fold in- crease in recurrence rate, a result almost identical to our origi- nal findings.

Throughout their review, Vamvakas and Moore examine individual pieces of evidence and uniformly criticize only those suggesting that transfusion is causally related to increased cancer recurrence, thus revealing a virtually complete bias against this hypothesis. In Table 1 of their review, there are actually six studies (not four) that found transfusion to be an independent predictor of cancer recurrence or death, unex- plained by known confounding factors. This, to us, is compel- ling evidence that transfusion most likely is causally related to colorectal cancer recurrence and death. The confounding factors (e.g., anemia and duration of surgery) that Vamvakas and Moore offer as potential explanations for the transfusion effect have not been shown in the literature to be significant predictors of cancer recurrence and certainly do not explain the findings in our own studies to date. Vamvakas and Moore describe a significant, pooled estimate of the transfusion-as- sociated increase in the risk of death or cancer recurrence as 37 percent and call this “small.” Even if only one-half of this association is eventually determined to be causal, we know of no oncologist who would consider an almost 20-percent in- crease in recurrence or death as small.

With a handful of exceptions, transfusion is always associ- ated with poorer outcomes in cancer surgery in dozens of studies in the literature. This sort of consistency is extremely unlikely to be explained by confounding variables yet to be identified. Indeed, the estimated increased risk of recurrence in transfused colorectal cancer patients has been shown in previous meta- analyses, uncited by Vamvakas and Moore, to be in the range of 80 p e r ~ e n t . ~ . ~ An effect of this magnitude, independent of histopathologic stage (the only uniformly accepted prognostic factor to date), is unlikely to be explained by confounding variables of little or no prognostic significance in this disease. That Vamvakas and Moore obtained a significant, but much smaller estimate of the recurrence risk associated with trans-

TRANSFUSION 1994-Vol. 34. No. 5 LETTERS TO THE EDITOR 45 1

fusion also suggests that their selection criteria were biased in favor of studies with smaller effects.

Finally, Vamvakas and Moore do not find the animal and human data a compelling reason for further study of this issue in humans. In contrast, we believe that investigations in pa- tients are essential and urgent. For one thing, preliminary ani- mal and human data suggest that white cell reduction might be useful in reducing the immunosuppressive effects of transfu- sion. The opportunities for reducing morbidity and mortality in these common life-threatening diseases may be substantial. Studies of the sort opposed by Vamvakas and Moore will be critical in determining whether these hopes can be fulfilled.

NEIL BLUMBERG, MD Transfusion Medicine Unit

Box 608 University of Rochester Medical Center

601 Elmwood Avenue Rochester, NY 14642

JOANNA M. HEAL, MRCP Hematology Unit

Department of Medicine University of Rochester Medical Center

and American Red Cross Blood Services

Rochester

1.

2.

3.

4.

5.

6.

References Vamvakas E, Moore SB. Penoperative blood transfusion and colorectal cancer recurrence: a qualitative statistical overview and meta-analysis. Transfusion 1993;33:754-65. Blumberg N, Aganval MM, Chuang C. Relation between re- currence of cancer of the colon and blood transfusion. Br Med J (Clin Res Ed) 1985;290:1037-9. Foster RS Jr, Costanza MC, Foster JC, Wanner MC, Foster CB. Adverse relationship between blood transfusions and survival after colectomy for colon cancer. Cancer 198535: 1195-201. Marsh J, Donnan FT, Hamer-Hodges DW. Association between transfusion with plasma and the recurrence of colorectal car- cinoma. Br J Surg 1990,77:623-6. Chung M, Steinmetz OK, Gordon PH. Perioperative blood transfusion and outcome after resection for colorectal cancer. Br J Surg 1993;80:427-32. Amato A, Butti A. Do patients with less advanced colorectal cancem recur more when perioperatively transfused? (abstract). J Surg Oncol 1991;48(Suppl 2):169.

The above letter was sent to Drs. Vamvakas and Moore, who offered the following reply.

The principal aim of our statistical overview1 was to present nonstatisticians with the complexities and “hidden” assump- tions of multivariate analysis, so that readers could assess the completeness of statistical control for confounding in studies that reported an association of perioperative blood transfusion with colorectal cancer recurrence. As explained in detail (pp 755-8) in our article,’ control for confounding can be achieved through multivariate analysis only if all confounders of the association under study are physically present in the final model. Confounders (e.g., admission hemoglobin and duration of sur- gery) are highly interrelated with transfusion, and the statisti- cal software package’ that was used by Blumberg et al.) does not tolerate the coexistence of interrelated variables in models that are built in a stepwise manner. In this particular statistical program, when a “more significant” variable such as transfu- sion enters the model, interrelated confounders are dropped as “nonsignificant.”’ Confounders are retained in the model (fol-

lowing entry of transfusion) only if they were entered by the forced-entry method, which does not calculate a p value for any confounder.2 In the culminating paragraph of the Results section (p 1039), Blumberg et al. offer a step-by-step descrip- tion of how the confounders age and duration of operation became “nonsignificant” following the entry of transfusion into the model,) so that they were dropped from the model and re- mained unadjusted in the published results.2 Similarly, in the final model of Foster et al.? hemoglobin became “nonsignifi- cant” (p = 0.53) after the entry of transfusion. The prevalence of the stepwise method was acknowledged by Triulzi et al. in a 1990 review of 12 studies (including one of their own) that had reported an association of transfusion with postoperative infections: “Stepwise multivariate regression analysis was performed in all but one ... of these 12 studies.”5(P’00)

Blumberg et aL3 and Foster et al.“ discussed their findings as adjusted for all confounders that were eligible for entry into the final model, as opposed to physically present in it. Accord- ingly, and despite what Blumberg and Heal stated in their let- ter,6 the confounders listed there could not be and were not controlled in those analy~es.) .~ Our article did not state that the entire transfusion effect calculated by Blumberg et aL3 and Foster et al.“ should be ascribed to these unadjusted confound- ing factors, which allows Blumberg and Heal to claim that these confounders “do not explain the findings in [their] own stud- ies to date.”6 Nobody has said otherwise. Discrepancies be- tween stepwise and forced-entry analyses vary with data sets; the extent to which a proper analysis (of the latter type) would diminish or eliminate a previously calculated transfusion ef- f e ~ t ~ , ~ is unpredictable.

In the cohort of 132 patients not experiencing a harmful effect of transfusion (p = 0.24) in the report by Marsh et al., the only subgroup to demonstrate an adverse effect consisted of 13 patients receiving plasma protein fraction and 1 patient transfused with fresh-frozen p l a ~ m a . ~ These investigators hoped that “other authors might analyze their data to see if they can confirm an independent harmful effect of plasma protein fra~tion.”’(fi~~) “Transfusion with whole blood had a relative risk of 2.47 ... but did not reach statistical significance (p = 0.073)”7(fi25) in the study of Marsh et al., a finding that Blumberg and Heal6 now consider “almost identical” to their original findings.8 Apparently, what constitutes a “factual er- ror” depends on the definition of a “fact.”

Our Table 1 I listed one study in which the adverse transfu- sion effect had withstood control for all known confounders? Information on all known confounders had been collected by those authors and possibly by the authors of one other study.3 Blumberg and Heal claim that six of the listed studies achieved such control. This is most puzzling to us, as we are aware of no method whereby statistical control for the effect of a vari- able would be possible even if information on that factor is not available for analysis. We also find absurd the present omis- sions of duration of operation and rectal location of the tumor from the list of factors associated with a poorer outcome,6 given what was demonstrated and discussed earlier by Blumberg et

We reported a 37-percent increase in the risk of cancer re- currence associated with transfusion, basing that finding on calculations that incorporated the confounding effects of all known and unknown confounders other than tumor location in the rectum.’ In an April 1993 meta-analysis (that followed receipt of the revised version of our manuscript’), Chung et al.” calculated an 80-percent increase in risk that was also confounded by tumor location in the rectum. While discussing rectal location of the tumor as a ubiquitous confounder, those authors pointed out that “it is not possible to say whether trans-

a1.3

45 2 LETTERS TO THE EDITOR TRANSFUSION Vol. 34. No 5-1994

fusion is responsible for the worse outcome among transfused patients or whether it is due to one or more confounding vari- ables.”1° (H3’) The other meta-analysis cited by Blumberg and Heal6 has appeared only in abstract form, which precludes appropriate evaluation of the methods used. Those authors stated, however, that 12 of 25 reviewed studies (48%. or “a handful of exceptions,” according to Blumberg and Heal6) had observed no adverse transfusion effect.

We emphasized that the calculated 37-percent increase in risk “should not be interpreted as supporting a true deleterious transfusion effect,”’ (P760) given the extent of residual uncon- trolled confounding. We endorsed’ a generally accepted opin- ion” that experimental research on humans (intended to prove a causal relationship) is timely only after a valid statistical association has been established by observational studies. This has been misrepresented6 as opposition to further study of this issue in humans. Nothing could be further from the truth, as evidenced from the closing statement of our article: “If a small adverse transfusion effect is established in the future, a radical alteration in transfusion practice will likely be justified in light of the prevalence and public health implications of the prob- lem”’ (P7M) We fail to comprehend how Blumberg and Heal manage consistently to confuse what has already been proved to be causal with what might be established in the future.

Along these same lines, we are not impressed by the hu- man data that “suggest that white cell reduction might be use- ful in reducing the immunosuppressive effects of transfusion.”6 Certainly, this was not the conclusion of the single random- ized clinical trial reported to date.”

ELEFTHERIOS C. VAMVAKAS, MD, MPHIL, PHD, MPH, MPA

S. BREANNDAN MOORE, MB, BCH, FRCPI Division of Transfusion Medicine

Mayo Clinic and Mayo Foundation Rochester, MN 55905

References I . Vamvakas E, Moore SB. Perioperative blood transfusion and

colorectal cancer recurrence: a qualitative statistical overview and meta-analysis. Transfusion I993;33:754-65. Dixon WJ, ed. Survival analysis with co-variates (2L). BMDP Statistical Software Manual, vol2, 1992 revision. Berkeley, CA: University of California Press, 1992:825-64. Blumberg N, Aganval MM, Chuang C. Relation between re- currence of cancer of the colon and blood transfusion. Br Med J (Clin Res Ed) 1985;290:1037-9. Foster RS Jr, Costanza MC, Foster JC, Wanner MC, Foster CB. Adverse relationship between blood transfusions and survival after colectomy for colon cancer. Cancer 1985;55: 1195-201. Triulzi DJ, Blumberg N, Heal JM. Association of transfusion with postoperative bacterial infection. Crit Rev Clin Lab Sci

6. Blumberg N, Heal JM. Perioperative blood transfusion and colorectal cancer recurrence (letter). Transfusion

Marsh J, Donnan IT, Hamer-Hodges DW. Association between transfusion with plasma and the recurrence of colorectal car- cinoma. Br J Surg 1990;77:623-6. Blumberg N, Heal JM, Murphy P, Agarwal MM, Chuang C. Association between transfusion of whole blood and recurrence

2.

3.

4.

5 .

1990;28:95- 107.

1994;34:450- 1. 7.

8.

9.

10.

of cancer. Br Med J (Clin Res Ed) 1986;293:530-3. Parrott NR, Lennard TW, Taylor RM, Proud G, Shenton BK, Johnston JD. Effect of perioperative blood transfusion on re- currence of colorectal cancer. Br J Surg 1986;73:970-3. Chung M, Steinmetz OK, Gordon PH. Perioperative blood transfusion and outcome after resection for colorectal cancer.

11. Hennekens CH, Buring JE. Statistical association and cause- effect relationships. In: Epidemiology in medicine. Boston: Little, Brown, 1987:30-53.

12. Busch ORC, Hop WCJ, Hoynck van Papendrecht MAW, Marquet RL, Jeekel J. Blood transfusions and prognosis in colorectal cancer. N Engl J Med 1993;328: 1372-6.

Editor’s Note: The usual limits on length and number of references cited per letter were waived in this instance to al- low a more complete discussion of this important topic.

Ferritin levels and iron deficiency

To the Editor: The recent article by Ho’ and subsequent correspondence

from Saxena and Shulman2 highlighted the chronic problem of reliable evaluation of iron stores. There is general agree- ment that serum ferritin is the preferred single test. Although bone marrow or liver biopsy may still be regarded as primary standards, these tests are not appropriate or feasible for large clinical or population studies. The initial evaluations of the relationship between serum ferritin levels and body iron sta- tus compared serum ferritin levels and the presence or absence of tissue iron. To estimate the latter, simple stains such as po- tassium ferrocyanide or, less often, more complicated physi- cal measures of iron content were used. The general conclu- sion of these various studies was that it was very unlikely that stainable storage iron would be present at fenitin levels of less than 15 pg per L.3

A problem arises when one introduces the concept of nor- mal ranges or, perhaps more correctly stated, reference ranges. These are mathematical and not clinical concepts. They relate to the statistical theory of the distribution of variables in popu- lations. This simplest concept is the use of 2 standard devia- tions from the population mean of a presumed normal sample distribution. Statisticians have no problem with this concept, but physicians tend to ascribe clinical properties to what is strictly a mathematical model. A relationship exists between means and standard deviations, or confidence intervals. Lower means are usually associated with a lower minimum reference point. In the case of serum ferritin, this means the lower refer- ence point in a sample of women of child-bearing age will be lower than that in a male population of similar age. This has resulted in the biologically nonsensical assumption that a woman has to have a lower femtin level before it is likely that iron will not be demonstrable in the tissue stores. Provided there are no major differences between the equilibrium coefficients of tissue and serum fenitin in men and women, this is obvi- ously a false conclusion. Clearly the level of ferritin that should precipitate investigation and treatment of iron deficiency should be the same for both sexes. In other words, if the question of interest is the percentage of individuals who have absent stor- age iron, then the same minimum level of ferritin should be used for both sexes.

The problem of relating statistical theory to biologic real- ity has been addressed extensively over the last three decades by both Feinstein4 and Murphy5 and makes for interesting as well as informative reading.

CEDRIC J. CARTER, MB, FRCP Department of Pathology

Faculty of Medicine University of British Columbia Hospital

221 I Westbrook Mall Br J Surg I993;80:427-32. Vancouver, BC, Canada V6T 2B5